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Lasso-type sparse regression and high-dimensional Gaussian graphical models 2012
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Title | Lasso-type sparse regression and high-dimensional Gaussian graphical models |
Creator |
Chen, Xiaohui |
Publisher | University of British Columbia |
Date Created | 2012-04-30T21:53:36Z |
Date Issued | 2012-04-30 |
Date | 2012 |
Description | High-dimensional datasets, where the number of measured variables is larger than the sample size, are not uncommon in modern real-world applications such as functional Magnetic Resonance Imaging (fMRI) data. Conventional statistical signal processing tools and mathematical models could fail at handling those datasets. Therefore, developing statistically valid models and computationally efficient algorithms for high-dimensional situations are of great importance in tackling practical and scientific problems. This thesis mainly focuses on the following two issues: (1) recovery of sparse regression coefficients in linear systems; (2) estimation of high-dimensional covariance matrix and its inverse matrix, both subject to additional random noise. In the first part, we focus on the Lasso-type sparse linear regression. We propose two improved versions of the Lasso estimator when the signal-to-noise ratio is low: (i) to leverage adaptive robust loss functions; (ii) to adopt a fully Bayesian modeling framework. In solution (i), we propose a robust Lasso with convex combined loss function and study its asymptotic behaviors. We further extend the asymptotic analysis to the Huberized Lasso, which is shown to be consistent even if the noise distribution is Cauchy. In solution (ii), we propose a fully Bayesian Lasso by unifying discrete prior on model size and continuous prior on regression coefficients in a single modeling framework. Since the proposed Bayesian Lasso has variable model sizes, we propose a reversible-jump MCMC algorithm to obtain its numeric estimates. In the second part, we focus on the estimation of large covariance and precision matrices. In high-dimensional situations, the sample covariance is an inconsistent estimator. To address this concern, regularized estimation is needed. For the covariance matrix estimation, we propose a shrinkage-to-tapering estimator and show that it has attractive theoretic properties for estimating general and large covariance matrices. For the precision matrix estimation, we propose a computationally efficient algorithm that is based on the thresholding operator and Neumann series expansion. We prove that, the proposed estimator is consistent in several senses under the spectral norm. Moreover, we show that the proposed estimator is minimax in a class of precision matrices that are approximately inversely closed. |
Genre |
Thesis/Dissertation |
Type |
Text |
Language | Eng |
Collection |
Electronic Theses and Dissertations (ETDs) 2008+ |
Date Available | 2012-04-30T21:53:36Z |
DOI | 10.14288/1.0072755 |
Degree |
Doctor of Philosophy - PhD |
Program |
Electrical and Computer Engineering |
Affiliation |
Applied Science, Faculty of |
Degree Grantor | University of British Columbia |
Graduation Date | 2012-11 |
Campus |
UBCV |
Scholarly Level | Graduate |
URI | http://hdl.handle.net/2429/42271 |
Aggregated Source Repository | DSpace |
Digital Resource Original Record | https://open.library.ubc.ca/collections/24/items/1.0072755/source |
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Lasso-Type Sparse Regression and High-Dimensional Gaussian Graphical Models by Xiaohui Chen M.Sc., The University of British Columbia, 2008 B.Sc., Zhejiang University, 2006 A THESIS SUBMITTED IN PARTIAL FULFILLMENT OF THE REQUIREMENTS FOR THE DEGREE OF DOCTOR OF PHILOSOPHY in The Faculty of Graduate Studies (Electrical and Computer Engineering) THE UNIVERSITY OF BRITISH COLUMBIA (Vancouver) April 2012 c© Xiaohui Chen 2012 Abstract High-dimensional datasets, where the number of measured variables is larger than the sample size, are not uncommon in modern real-world applications such as functional Magnetic Resonance Imaging (fMRI) data. Conventional statistical signal processing tools and mathematical models could fail at handling those datasets. Therefore, developing statistically valid models and computationally efficient algorithms for high- dimensional situations are of great importance in tackling practical and scientific problems. This thesis mainly focuses on the following two issues: (1) recovery of sparse regression coefficients in linear systems; (2) estimation of high-dimensional covariance matrix and its inverse matrix, both subject to additional random noise. In the first part, we focus on the Lasso-type sparse linear regression. We propose two improved versions of the Lasso estimator when the signal-to-noise ratio is low: (i) to leverage adaptive robust loss functions; (ii) to adopt a fully Bayesian mod- eling framework. In solution (i), we propose a robust Lasso with convex combined loss function and study its asymptotic behaviors. We further extend the asymptotic analysis to the Huberized Lasso, which is shown to be consistent even if the noise distribution is Cauchy. In solution (ii), we propose a fully Bayesian Lasso by unifying discrete prior on model size and continuous prior on regression coefficients in a single modeling framework. Since the proposed Bayesian Lasso has variable model sizes, we propose a reversible-jump MCMC algorithm to obtain its numeric estimates. In the second part, we focus on the estimation of large covariance and precision matrices. In high-dimensional situations, the sample covariance is an inconsistent es- timator. To address this concern, regularized estimation is needed. For the covariance matrix estimation, we propose a shrinkage-to-tapering estimator and show that it has ii Abstract attractive theoretic properties for estimating general and large covariance matrices. For the precision matrix estimation, we propose a computationally efficient algorithm that is based on the thresholding operator and Neumann series expansion. We prove that, the proposed estimator is consistent in several senses under the spectral norm. Moreover, we show that the proposed estimator is minimax in a class of precision matrices that are approximately inversely closed. iii Preface This thesis is written based on a collection of manuscripts, resulting from collaboration between several researchers. Chapter 2 is based on a journal paper published in IEEE Transactions on Information Theory [31], co-authored with Prof. Z. Jane Wang and Prof. Martin J. McKeown. Chapter 3 is based on a journal paper published in Signal Processing [32], co-authored with Prof. Z. Jane Wang and Prof. Martin J. McKeown. Chapter 4 is based on a submitted journal paper co-authored with Prof. Z. Jane Wang and Prof. Martin J. McKeown. Chapter 5 is based on a journal paper to appear in IEEE Transactions on Signal Processing [30], co-authored with Prof. Young-Heon, Kim and Prof. Z. Jane Wang. Chapter 6 is based on a conference paper appears in 2010 International Conference in Image Processing [35], co-authored with Prof. Z. Jane Wang and Prof. Martin J. McKeown. The research outline was designed jointly by the author, Prof. Z. Jane Wang and Prof. Martin J. McKeown. The majority of the research, including literature survey, model design, theorem proofs, numerical simulation, statistical data analysis and results report, was conducted by the author, with suggestions from Prof. Z. Jane Wang and Prof. Martin J. McKeown. The manuscripts were primarily drafted by the author, with helpful comments from Prof. Z. Jane Wang and Prof. Martin J. McKeown. The biomedical application description in Section 5.6, Chapter 5, is written based on a grant proposal by Prof. Martin J. McKeown. iv Table of Contents Abstract . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . ii Preface . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . iv Table of Contents . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . v List of Tables . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . xi List of Figures . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . xii List of Acronyms . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . xiv Acknowledgements . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . xvi 1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 1 1.1 Challenges of High-Dimensional Modeling: An Overview . . . . . . . 1 1.2 Lasso-Type Sparse Linear Regression . . . . . . . . . . . . . . . . . . 3 1.2.1 Prior Arts . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 3 1.2.2 Our Contributions . . . . . . . . . . . . . . . . . . . . . . . . 8 1.3 High-Dimensional Covariance and Precision Matrix Estimation . . . 11 1.3.1 Challenges . . . . . . . . . . . . . . . . . . . . . . . . . . . . 11 1.3.2 Estimating Covariance Matrix . . . . . . . . . . . . . . . . . 13 1.3.3 Estimating Precision Matrix . . . . . . . . . . . . . . . . . . 15 1.4 Thesis Outline . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 17 2 Robust Lassos and Their Asymptotic Properties . . . . . . . . . . 20 v Table of Contents 2.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 20 2.1.1 Sparsity-Promoting Linear Models . . . . . . . . . . . . . . . 20 2.1.2 Summary of Present Contributions . . . . . . . . . . . . . . . 24 2.1.3 Organization of the Chapter . . . . . . . . . . . . . . . . . . 25 2.2 A Convex Combined Robust Lasso . . . . . . . . . . . . . . . . . . . 26 2.2.1 A Robust Lasso with the Convex Combined Loss . . . . . . . 26 2.2.2 Asymptotic Normality and Estimation Consistency . . . . . . 26 2.2.3 A Bound on MSE for the Noiseless Case . . . . . . . . . . . . 30 2.3 The Adaptive Robust Lasso and Its Model Selection Consistency . . 32 2.4 The Huberized Lasso . . . . . . . . . . . . . . . . . . . . . . . . . . . 33 2.5 Random Designs . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 36 2.6 Numeric Examples . . . . . . . . . . . . . . . . . . . . . . . . . . . . 38 2.7 Conclusion and Discussion . . . . . . . . . . . . . . . . . . . . . . . . 44 3 A Bayesian Lasso via Reversible-Jump MCMC . . . . . . . . . . . 46 3.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 46 3.1.1 Sparse Linear Models . . . . . . . . . . . . . . . . . . . . . . 46 3.1.2 Related Work and Our Contributions . . . . . . . . . . . . . 47 3.1.3 Notations . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 49 3.2 A Fully Bayesian Lasso Model . . . . . . . . . . . . . . . . . . . . . 50 3.2.1 Prior Specification . . . . . . . . . . . . . . . . . . . . . . . . 50 3.3 Bayesian Computation . . . . . . . . . . . . . . . . . . . . . . . . . . 53 3.3.1 Design of Model Transition . . . . . . . . . . . . . . . . . . . 54 3.3.2 A Usual Metropolis-Hastings Update for Unchanged Model Di- mension . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 55 3.3.3 A Birth-and-Death Strategy for Changed Model Dimension . 56 3.4 Simulations . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 58 3.4.1 Setup . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 58 3.4.2 Empirical Performance Comparisons . . . . . . . . . . . . . . 59 vi Table of Contents 3.4.3 Convergence Analysis . . . . . . . . . . . . . . . . . . . . . . 62 3.5 A Diabetes Data Example . . . . . . . . . . . . . . . . . . . . . . . . 65 3.6 A Real fMRI Application . . . . . . . . . . . . . . . . . . . . . . . . 66 3.6.1 Application Description . . . . . . . . . . . . . . . . . . . . . 66 3.6.2 Results . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 68 3.7 Discussion and Conclusion . . . . . . . . . . . . . . . . . . . . . . . . 69 4 Shrinkage-To-Tapering Estimation of Large Covariance Matrices 72 4.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 72 4.2 Comparison Between Tapering and Shrinkage Estimators . . . . . . 73 4.2.1 Tapering Estimator . . . . . . . . . . . . . . . . . . . . . . . 73 4.2.2 Shrinkage Estimator . . . . . . . . . . . . . . . . . . . . . . . 76 4.2.3 Comparison of Risk Bounds Between Tapering and Shrinkage Estimators . . . . . . . . . . . . . . . . . . . . . . . . . . . . 79 4.3 A Shrinkage-to-Tapering Estimator . . . . . . . . . . . . . . . . . . . 84 4.3.1 Problem Formulation . . . . . . . . . . . . . . . . . . . . . . 84 4.3.2 Approximating the Oracle . . . . . . . . . . . . . . . . . . . . 85 4.4 Simulation . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 87 4.4.1 Model 1: AR(1) Model . . . . . . . . . . . . . . . . . . . . . 88 4.4.2 Model 2: Σ ∈ G(α−1, C, C0) . . . . . . . . . . . . . . . . . . . 91 4.4.3 Model 3: Fractional Brownian Motion . . . . . . . . . . . . . 92 4.5 Conclusion . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 96 5 Efficient Minimax Estimation of High-Dimensional Sparse Precision Matrices . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 100 5.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 100 5.1.1 Innovation and Main Results . . . . . . . . . . . . . . . . . . 102 5.1.2 Comparison with Existing Work . . . . . . . . . . . . . . . . 102 5.2 Approximately Inversely Closed Sparse Matrices . . . . . . . . . . . 104 vii Table of Contents 5.2.1 The Neumann Series Representation of Ω . . . . . . . . . . . 104 5.2.2 A Class of Sparse Matrices with Approximately Inverse Close- ness . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 105 5.3 Proposed Estimator . . . . . . . . . . . . . . . . . . . . . . . . . . . 108 5.3.1 Algorithm . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 108 5.3.2 Consistency . . . . . . . . . . . . . . . . . . . . . . . . . . . . 109 5.3.3 Sharpness: Optimal Under Minimax Risk . . . . . . . . . . . 113 5.3.4 Model Selection Consistency . . . . . . . . . . . . . . . . . . 115 5.3.5 Extensions . . . . . . . . . . . . . . . . . . . . . . . . . . . . 116 5.4 Practical Choices of η, r and τ . . . . . . . . . . . . . . . . . . . . . 117 5.5 Numerical Experiments . . . . . . . . . . . . . . . . . . . . . . . . . 118 5.6 Application to An Real fMRI data . . . . . . . . . . . . . . . . . . . 122 5.6.1 Modeling F→STN . . . . . . . . . . . . . . . . . . . . . . . . 122 5.6.2 Learned F→STN Connectivities . . . . . . . . . . . . . . . . 123 5.7 Conclusion and Discussion . . . . . . . . . . . . . . . . . . . . . . . . 125 6 fMRI Group Analysis of Brain Connectivity . . . . . . . . . . . . . 128 6.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 128 6.2 Methods . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 130 6.2.1 A Group Robust Lasso Model . . . . . . . . . . . . . . . . . . 130 6.2.2 A Group Sparse SEM+mAR(1) Model . . . . . . . . . . . . . 133 6.3 A Simulation Example . . . . . . . . . . . . . . . . . . . . . . . . . . 135 6.4 fMRI Group Analysis in Parkinson’s Disease . . . . . . . . . . . . . 136 6.4.1 Data Description . . . . . . . . . . . . . . . . . . . . . . . . . 136 6.4.2 Learned Brain Connections . . . . . . . . . . . . . . . . . . . 138 7 Conclusions and Discussions . . . . . . . . . . . . . . . . . . . . . . . 140 7.1 Contribution Summary . . . . . . . . . . . . . . . . . . . . . . . . . 140 7.2 Directions for Future Research . . . . . . . . . . . . . . . . . . . . . 143 viii Table of Contents 7.2.1 Estimation of Conditional Independence for High-Dimensional Non-Stationary Time Series . . . . . . . . . . . . . . . . . . . 143 7.2.2 Estimation of Eigen-Structures of High-Dimensional Covari- ance Matrices . . . . . . . . . . . . . . . . . . . . . . . . . . . 145 7.2.3 Multi-Task Lasso for Group Analysis . . . . . . . . . . . . . . 146 Bibliography . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 147 Appendix . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 161 A.1 Notations . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 161 A.2 Proofs for Chapter 2 . . . . . . . . . . . . . . . . . . . . . . . . . . . 163 A.2.1 Proof of Theorem 2.2.1 . . . . . . . . . . . . . . . . . . . . . 163 A.2.2 Proof of Corollary 2.2.2 . . . . . . . . . . . . . . . . . . . . . 165 A.2.3 Proof of Corollary 2.2.3 . . . . . . . . . . . . . . . . . . . . . 166 A.2.4 Proof of Theorem 2.2.4 . . . . . . . . . . . . . . . . . . . . . 166 A.2.5 Proof of Proposition 2.2.5 . . . . . . . . . . . . . . . . . . . . 168 A.2.6 Proof of Theorem 2.3.1 . . . . . . . . . . . . . . . . . . . . . 171 A.2.7 Proof of Theorem 2.4.1 . . . . . . . . . . . . . . . . . . . . . 174 A.2.8 Proof of Theorem 2.5.1 . . . . . . . . . . . . . . . . . . . . . 177 A.2.9 Proof of Auxiliary Lemmas . . . . . . . . . . . . . . . . . . . 179 A.3 Appendix for Chapter 3 . . . . . . . . . . . . . . . . . . . . . . . . . 184 A.3.1 Derivation of the Joint Posterior Distribution in (3.3) . . . . 184 A.3.2 MCMC Algorithm for the Binomial-Gaussian Model . . . . . 185 A.4 Proofs for Chapter 4 . . . . . . . . . . . . . . . . . . . . . . . . . . . 188 A.4.1 Proof of Theorem 4.2.2 . . . . . . . . . . . . . . . . . . . . . 188 A.4.2 Proof of Theorem 4.2.3 . . . . . . . . . . . . . . . . . . . . . 189 A.4.3 Proof of Theorem 4.3.1 . . . . . . . . . . . . . . . . . . . . . 189 A.5 Proofs for Chapter 5 . . . . . . . . . . . . . . . . . . . . . . . . . . . 191 A.5.1 Proof of Lemma 5.2.1 . . . . . . . . . . . . . . . . . . . . . . 191 ix Table of Contents A.5.2 Proof of Theorem 5.2.2 . . . . . . . . . . . . . . . . . . . . . 194 A.5.3 Proof of Proposition 5.3.1 . . . . . . . . . . . . . . . . . . . . 195 A.5.4 Proof of Theorem 5.3.3 . . . . . . . . . . . . . . . . . . . . . 196 A.5.5 Proof of Proposition 5.3.5 . . . . . . . . . . . . . . . . . . . . 198 A.5.6 Proof of Theorem 5.3.6 . . . . . . . . . . . . . . . . . . . . . 199 A.5.7 Proof of Theorem 5.3.9 . . . . . . . . . . . . . . . . . . . . . 202 A.6 Proofs for Chapter 6 . . . . . . . . . . . . . . . . . . . . . . . . . . . 203 A.6.1 Proof of Theorem 6.2.1 . . . . . . . . . . . . . . . . . . . . . 203 Appendix x List of Tables 2.1 Theoretic and empirical asymptotic variances of √ n(β̂n − β). . . . . 39 3.1 Some special functions and probability density functions. . . . . . . . 50 3.2 RMSEs of the Lasso [115], Gauss-Lasso, Lar [45], Gauss-Lar, Gibbs sampler based Bayesian Lasso [102], Binomial-Gaussian (BG) [59], pro- posed BG-MCMC and proposed Bayesian Lasso (BLasso) with RJ- MCMC algorithm. . . . . . . . . . . . . . . . . . . . . . . . . . . . . 59 3.3 F -scores of estimated sparsity patterns of the Lasso [115], Lar [45], Gibbs sampler based Bayesian Lasso [102], Binomial-Gaussian (BG) [59], the proposed BG-MCMC and the proposed Bayesian Lasso (BLasso) with RJ-MCMC algorithm. . . . . . . . . . . . . . . . . . . . . . . . 63 3.4 Estimated coefficients for the diabetes data set. . . . . . . . . . . . . 66 3.5 Correlations between the coefficient estimates of MLE, Lar [45], Lasso [115] and the proposed BLasso on two fMRI sub-datasets. . . . . . . . . . . 69 3.6 CPU time for for Lar [45], Lasso [115], the Bayesian Lasso based on the Gibbs sampler (GS) [102], BG [59], BG-MCMC, and proposed Bayesian Lasso. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 71 5.1 Estimation error under the spectral norm, specificity, and sensitivity of Ω̂, Ω̂taper, CLIME, graphical Lasso (GLasso), and SCAD. . . . . . . 119 6.1 MSEs for the estimated coefficients of grpLasso, RLasso, grpRLasso, and the oracle. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 135 xi List of Figures 2.1 Histograms of √ n(β̂n − β) for the Gaussian mixture with µ = 5. . . . 42 2.2 Histograms of √ n(β̂n − β) for the Gaussian mixture with µ = 5 with adaptation. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 43 3.1 An illustration of model jumping from γ → γ ′ with |γ| = 5 and |γ ′| = 6. 54 3.2 The correlation between the estimated coefficients when using different algorithms. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 68 4.1 Normalized MSE curves of the shrinkage MMSE estimator for the large covariances discussed in Example 4.2.1 and 4.2.2. . . . . . . . . . . . 81 4.2 Model 1: The normalized MSE curves. . . . . . . . . . . . . . . . . . 89 4.3 Model 1: The spectral risk curves. . . . . . . . . . . . . . . . . . . . . 90 4.4 Model 1: The estimated shrinkage coefficients. . . . . . . . . . . . . . 91 4.5 Model 2: The normalized MSE curves. . . . . . . . . . . . . . . . . . 92 4.6 Model 2: The spectral risk curves. . . . . . . . . . . . . . . . . . . . . 93 4.7 Model 2: The estimated shrinkage coefficients. . . . . . . . . . . . . . 94 4.8 Model 3: The normalized MSE curves. . . . . . . . . . . . . . . . . . 95 4.9 Model 3: The spectral risk curves. . . . . . . . . . . . . . . . . . . . . 97 4.10 Model 3: The estimated shrinkage coefficients for different estimators. 98 5.1 A diagram illustrating the proposed Algorithm 3. . . . . . . . . . . . 108 5.2 MCC plots of the estimated Ω by various algorithms. . . . . . . . . . 121 5.3 True sparse precision matrix Ω with p = 200. . . . . . . . . . . . . . . 121 5.4 Estimated Ω for two normal subjects. . . . . . . . . . . . . . . . . . . 124 xii List of Figures 5.5 Identified PC patterns for two normal subjects. . . . . . . . . . . . . 125 5.6 Loadings of the identified PCs for two normal subjects. . . . . . . . . 126 6.1 Weighted network learned for the normal group. . . . . . . . . . . . . 137 6.2 Different connections between normal and “off-medication” networks. 137 7.1 This overview summarizes the challenges raised in Chapter 1, the meth- ods proposed in this thesis, and the relationship between proposed methods and the challenges being addressed. . . . . . . . . . . . . . . 142 xiii List of Acronyms AIC Akaike Information Criterion BIC Bayesian Information Criterion BG Binomial-Gaussian model CLIME Constrained `1-Minimization for Inverse Matrix Estimation CMT Covariance Matrix Taper CV Cross-Validation DBN Dynamic Bayesian Networks FBM Fractional Brownian Motion fMRI functional Magnetic Resonance Imaging iid independent and identically distributed LAR(S) Least Angle Regression Lasso Least Absolute Shrinkage and Selection Operator LS Least Squares MAP Maximum A Posterior estimator mAR(r) multivariate Autoregressive model with order r MCC Mathews Correlation Coefficient MH Metropolis-Hastings algorithm/ratio xiv List of Acronyms MLE Maximum Likelihood Estimator MM Minorizatoin-Maximization algorithm (M/R)MSE (Minimum/Root) Mean-Squared Error OAS Oracle Approximating Shrinkage PCA Principle Component Analysis PD Parkinson’s Disease RIC Risk Inflation Criterion RJ-MCMC Reversible-Jump Markov chain Monte Carlo algorithm RLAD Regularized Least Absolute Deviation ROI Region of Interest RW Random Walk SCAD Smoothly Clipped Absolute Deviation SEM Structural Equation Modeling SNR Signal-to-Noise Ratio SPICE Sparse Permutation Invariant Covariance Estimator STO Shrinkage-to-Tapering Oracle estimator STOA Shrinkage-to-Tapering Oracle Approximating algorithm TNR True Negative Rate TPR True Positive Rate xv Acknowledgements First and foremost, I owe innumerable thanks to my PhD advisers, Prof. Z.Jane Wang and Prof. Martin J. McKeown, for being great mentors, both professionally and personally. This thesis would never be possible without their continuous support over the years. Many of their valuable and insightful suggestions not only encouraged me to constantly learn new things, but also taught me how to be an independent researcher. I am in particular indebted to them for generously allowing me with enough freedom for exploring new research topics of my own interests. I would like also to express thanks to my wonderful collaborators, co-authors, and fellow graduate students at UBC. In particular, I would like to thank Prof. Young- Heon, Kim (Dept. Mathematics) for many stimulating discussions and personal en- couragement. The research work in this thesis has been supported by a Pacific Alzheimer’s Research Foundation (PARF) Centre Grant Award and a Graduate Fellowship from UBC. xvi Chapter 1 Introduction 1.1 Challenges of High-Dimensional Modeling: An Overview Statistical estimation in high-dimensional situations, where the number of measured variables p is substantially larger than the sample size n (a.k.a. large-p-small-n), is fundamentally different from the estimation problems in the classical settings where we have small-p-large-n. Since high-dimensional datasets are not uncommon in mod- ern real-world applications, such as gene expression microarray data and functional Magnetic Resonance Imaging (fMRI) data, precise estimation of high-dimensional models is of great importance in tackling such practical and scientific problems. Gen- erally speaking, learning salient information from relatively a few samples when many more variables are present is not possible without knowing special structures in the data. To alleviate the ill-posed problem, it is natural to restrict our attention to sub- sets of all solutions with certain special structures or properties and meanwhile to incorporate the regularization ideas into estimation. Sparsity is one commonly hy- pothesized condition and it seems to be realistic for many real-world applications. There has been a surge in the literature, termed compressed sensing in signal pro- cessing literature and Lasso in statistical literature, on the recovery of sparse signals in under-determined linear systems [23, 24, 26–28, 42, 43, 115]. Many beautiful re- sults on sparse representation, recovery conditions and algorithms have been reported in the literature. We remark that the literature on this topic is too extensive for us 1 1.1. Challenges of High-Dimensional Modeling: An Overview to give an exhaustive list. My PhD thesis mainly focuses on the following two issues: (1) the recovery of sparse regression coefficients in linear systems; (2) estimation of high-dimensional covariance matrix and its inverse matrix (a.k.a precision matrix, or the Gaussian graphical model in machine learning language), both subject to random noise. It is emphasized that, in my PhD work, these two problems are studied from both theoretic and algorithmic perspectives. Although significant progress has been made on sparsity in high-dimensionality during the last decade, there are still a number of challenges attracting intensive research activities in the statistics, machine learning, and signal processing commu- nities. These include: C1 Minimal sampling: What are the fundamental information-theoretic limits of the sample size in order to obtain theoretically guaranteed correct and stable estimates? C2 Computational tractability: Can we develop concrete algorithms that are computationally feasible or even efficient for the seemingly daunting large-scale combinatorial problems in terms of computational cost? C3 Robustness: How can we make the feature selection tools adaptive to data and protective against non-Gaussian noise? C4 Consistency (e.g. estimation and model selection consistency under random noise): Can we guarantee the proposed algorithms and models work appropriately in theory, at least asymptotically? C5 Optimality: Is it possible to improve the proposed models in terms of conver- gence rate? C1 has been relatively well-studied in the literature and it has close connections to the approximation theory. As mentioned earlier, a large volume of compressed sensing papers have made beautiful solutions to this [19, 23–26, 28, 41–43, 117]. 2 1.2. Lasso-Type Sparse Linear Regression For C2, different high-dimensional estimation problems may have different problem features and we will see that convex relaxation and certain simple matrix operations often achieve computational efficiency. C3 is a practical concern since essentially all literature, with only a few exceptions [75, 107, 123], considers robust estimation procedures under the assumption of error distributions with heavy-tails. C4 and C5 together offer us an assurance to use the models from a theoretic perspective. While there are many potential practical applications for this method, a moti- vating application of this work is brain effective connectivity modeling using fMRI data, where the goal is to infer the connectivity network between a large number of (spatially-fixed) brain regions-of-interests (ROIs). Studying brain connectivity is crucial in understanding brain functioning and can provide significant insight into the pathophysiology of neurological disorders such as Parkinson’s disease. Based on prior neuroscience knowledge, the connections between brain regions generally can be considered a priori to form a sparse network. Several linear regression based for- malisms have been popular for inferring brain connectivity using fMRI and we have recently employed the unified structural equation modeling (SEM) and multivariate autoregressive (mAR) models to capture both spatial and temporal brain connec- tions. Moreover, it is well-known that fMRI data is typically very noisy. Therefore we formulate brain connectivity modeling as a problem of sparse linear regression under large variance noise [34]. 1.2 Lasso-Type Sparse Linear Regression 1.2.1 Prior Arts Recovering sparse signals from high-dimensional data sets has been attracting inten- sive research attention during the last decade. By sparse signals we mean that the underlying model generating all measured quantities can be approximated by a few numbers of the true signals and the approximation errors are due to (random) noise. 3 1.2. Lasso-Type Sparse Linear Regression In this part of the thesis, we consider the problem of estimating the coefficient vector in a linear regression model, defined as y = Xβ + e, (1.1) where the random measurement error vector e = (e1, · · · , en)∗ is assumed to be independent and identically distributed (iid) with zero mean and a constant finite second moment σ2 for each component. We regard e as a column vector, and use e∗ to denote its conjugate transpose. Here, X is the n × p design matrix which can either be non-stochastic or random. As usual, rows of X represent the p-dimensional observations and columns of X represent the predictors. yn×1 is the response vector and βp×1 is the coefficient vector to be estimated. We are interested in the setup where p is independent of n and fixed, but can be a large positive integer. There are many variable selection models proposed in the literature from both frequentist and Bayesian perspectives. Nevertheless, we are interested in sparsity- promoting linear regression models based on the least absolute shrinkage and selection operator, i.e. the Lasso [115], because of its popularity and its attractive computa- tional and theoretical properties [13, 16, 45, 75, 77, 86, 90, 94–96, 100, 101, 121, 128, 130, 132]. So far in the statistics community, the Lasso is probably the most popular variable selection tool used to estimate a sparse coefficient vector. In signal processing literature, the minimization of the `1 norm regularized linear model is often termed the basis pursuit [28]. Specifically, the Lasso estimator of Tibshirani [115] is defined as the following: Definition 1.2.1. The Lasso estimator is defined as β̂n = arg min u∈Rp ( 1 n n∑ i=1 (yi − x∗iu)2 + λn n p∑ j=1 |uj|γ ) , (1.2) where xi means the i th-row of X and γ = 1. Here, λn ≥ 0 is a shrinkage tuning parameter. A larger λn yields a sparser linear 4 1.2. Lasso-Type Sparse Linear Regression sub-model whereas a smaller λn corresponds to a less-sparse one. In extreme cases, λn = 0 gives the unregularized model and λn =∞ produces the null model consisting of no predictor. More generally, for γ > 0, (1.2) is called the bridge regression estimator by [55], and γ = 2 yields the ridge regression [68]. It is clear that (1.2) is convex for γ ≥ 1, and that it can produce sparsity when 0 < γ ≤ 1, since the penalized objective function has a non-trivial mass at zero. Therefore, the Lasso can be viewed as a sparsity-promoting convexification of the `2 loss plus the `0 penalty so that standard convex optimization technologies can be applied to efficiently solve (1.2) for the Lasso [42, 117]. The popularity of the Lasso partially relies on the existence of fast and efficient implementation algorithms. For example, using the piecewise linearity of the Lasso estimator, a modification of the Least Angle Regression (LARS) algorithm can compute the whole optimal path (corresponding to all λn ∈ [0,∞]) of the Lasso estimator on the same order as computational complexity of the least squares with size n×min(n, p) [45]. A similar homotopy algorithm was proposed in [101]. These attractive algorithms allow the scalability of the Lasso to high-dimensional situations. On the other hand, asymptotic properties of the Lasso estimator have also been extensively studied and analyzed. In a seminal work [77], Knight and Fu first derived the asymptotic distribution of the Lasso estimator (more generally the bridge estima- tor) and proved its estimation consistency under the shrinkage rate λn = o( √ n) and λn = o(n). More specifically, as long as errors are iid and possess a common finite second moment σ2, the √ n scaled Lasso estimator with a sequence of properly tuned shrinkage parameters {λn}n∈N has an asymptotic normal distribution with variance σ2C−1, where n−1 ∑n i=1 xix ∗ i → C and C is a positive definite matrix. Later, [86] showed that there is a non-vanishing probability of the Lasso selecting wrong models with the optimal prediction criterion such as cross-validation (CV). [95] also discov- ered the conflict between model selection consistency and optimal prediction in the Gaussian graphical model setup. [130] found a sufficient and necessary condition required on the design matrix for the Lasso estimator to be model selection consis- 5 1.2. Lasso-Type Sparse Linear Regression tent, i.e. the irrepresentable condition. This condition was also observed by [132]. In graphical models, [95] obtained a similar set of conditions for the variable selection consistency of the Lasso estimator, namely the neighborhood stability assumptions. These conditions are in general not easy to verify. Therefore, instead of requiring conditions on the design matrix for model selection consistency, there are also several variants of the original Lasso. For examples, the relaxed Lasso [94] uses two parame- ters to separately control the model shrinkage and selection; the adaptive Lasso [132] leverages a simple adaptation procedure to shrink the irrelevant predictors to 0 while keeping the relevant ones properly estimated; [96] suggested employ a two-stage hard thresholding rule, in the spirit of the Gauss-Dantzig selector [27], to set very small coefficients to 0. Since the groundbreaking work of [27] which provided non-asymptotic upper bounds on the `2 estimation loss of the Dantzig selector with large probability, par- allel `2 error bounds were found for the Lasso estimator by [96] under the incoherent design condition and by [13] under the restricted eigenvalue condition. In a previous work of [27], they showed that minimizing the `1 norm of the coefficient vector sub- ject to the linear system constraint can exactly recover the sparse patterns, provided the restricted isometry condition holds and the support of the noise vector is not too large [26]. [19] tightened all previous error bounds for noiseless, bounded error and Gaussian noise cases. These bounds are nearly optimal in the sense that they achieve within a logarithmic factor the LS errors as if the true model were known (oracle property). [121] derived a set of sharp constraints on the dimensionality, sparsity of the model and the number of observations for the Lasso to correctly recover the true sparsity pattern. The `∞ convergence rate of the Lasso estimator was obtained by [90]. Other bounds for the sparsity oracle inequalities of the Lasso can be found in [16]. As we have mentioned earlier, there is a second view of variable selection ap- proaches built on the Bayesian paradigm. Recent work has been proposed in the 6 1.2. Lasso-Type Sparse Linear Regression direction of Bayesian Lasso [102]. In [102], with a conditional Gaussian prior on β and the non-informative scale-invariant prior on the noise variance being assumed, a Bayesian Lasso model is proposed and a simple Gibbs sampler is implemented. It is shown that the Bayesian Lasso estimates in [102] are strikingly similar to those of the ordinary Lasso. Since this Bayesian Lasso in [102] involves the inversion of the co- variance matrix of block coefficients at each iteration, the computational complexity prevents its practical application with, say, hundreds of variables. Moreover, similar to the regular Lasso, the Bayesian Lasso in [102] uses only one shrinkage parameter t to control model size and shrink estimates. Nonetheless, it is arguable whether the two effects can be simultaneously well-handled by a single tuning parameter [94]. To mitigate this non-separability problem, [99] proposed an extended Bayesian Lasso model by assigning a more flexible, covariate-adaptive penalization on top of the Bayesian Lasso in the context of Quantitative Trait Loci (QTL) mapping. Alterna- tively, introducing different sources of sparsity-promoting priors on both coefficients and their indicator variables have been studied, e.g. in [104], where a normal-Jeffrey scaled-mixture prior on coefficients and an independent Bernoulli prior with small success probability on the binary index vector are combined. Despite those appealing properties of the Lasso estimator and the advocacy of using the Lasso, the Lasso estimate is not guaranteed to provide a satisfactory estima- tion and detection performance, at least in some application scenarios. For instance, when the data are corrupted by some outliers or the noise is extremely heavy-tailed, the variance of the Lasso estimator can be quite large, usually become unacceptably large, even when the sample size approaches infinity [77]. Asymptotic analysis [77] and non-asymptotic error bounds on the estimation loss [13] both suggest that the performance of the Lasso linearly deteriorates with the increment of the noise power. A similar observation can sometimes be noted when the dimensionality of the linear model is very high while the data size is much smaller. 7 1.2. Lasso-Type Sparse Linear Regression 1.2.2 Our Contributions In the above discussion, the distributions of measurement errors have not been spec- ified. Prior literature has mainly been concerned with either exact recovery of sparse signals [23, 24, 26, 28, 42, 117] or stable recovery under iid noise with moderate vari- ance, usually assumed to bounded or Gaussian (bounded in moments) [19, 25, 41, 43, 86]. In general, the error vector e is assumed to be iid Gaussian random variables with variance σ2, i.e. e follows the distribution N(0, σ2In×n). As we have seen in the previous section, e.g. [77], it is clear that the accuracy of the Lasso estimator critically depends on σ2 and the estimator is well suited for errors with a small or moderate variance σ2. It is also noted that when σ2 becomes larger, the Lasso estimator has a variance that is unbounded in the limiting case. This implies an undesirable property in real applications: instability. The reason for this poor performance with large σ2 lies in the sensitivity of the `2 loss function to a few large errors which may arise from heavy-tailed distribution or outliers. This explains why empirical examples show that the Lasso estimator can behave poorly if data are contaminated [75]. The standard assumption of random errors with small variance σ2, however, is unlikely to hold in many real applications, since outliers and/or heavy-tailed error distributions are commonly encountered in real situations. Therefore, the practical usage and efficiency of the Lasso can be limited. For example, it is typical for DNA microarray data to have a very low signal-to-noise ratio (SNR), meaning that σ2 is large. Furthermore, in practice the number of observations we are able to afford can be less than the dimensionality of the assumed model. Therefore, a robust variable selection model is necessary to obtain a good estimator in terms of accuracy, at least asymptotically. By robustness, here we mean two things: 1. The estimate is asymptotically stable in the presence of large noise. More specif- ically, we hope that, with more and more data being collected, the variability of the estimate is acceptable even if the measurement errors (the errors in the responses) get larger and larger. 8 1.2. Lasso-Type Sparse Linear Regression 2. The estimate is robust against contamination of the assumed errors. More specifically, even when outliers are found in the responses, the estimation per- formance is comparable to the situations of having no outliers. These two issues can be partially reflected in σ2. The first scenario can be viewed as errors following a distribution with heavy tails (e.g. Student-t distribution, Cauchy distribution), while the second one can be modeled as errors and outliers together contributing to form a mixture model of distributions. In either of the two scenarios, the corresponding σ2 can be very large or even infinity. In the first part of my thesis (Chapter 2 and 3), robust Lasso-type regression mod- els are considered when the noise has heavy-tails. More specifically, two solutions are proposed: (i) to leverage adaptive robust loss functions [31], as apposed to the Eu- clidean loss in standard Lasso; (ii) to adopt a fully Bayesian modeling framework [32]. Both solutions are aiming to obtain stabilized estimates. In solution (i), we propose a robust version of the Lasso by adopting a convex combined loss function and derive the asymptotic theory of the proposed robust Lasso estimator. We show that the ordinary Lasso and the regularized least absolute deviation (RLAD) [123] are two special cases of the proposed robust Lasso model. Although the RLAD is a robust model selection tool, it has limitations in terms of uniqueness, smoothness, and efficiency of estimation. Specifically, since the objective function of the RLAD is purely piecewise linear (thus may not be strictly convex) in β, its solution may not necessarily be unique in general [14]. Moreover, since the optimal path for the RLAD is discontinuous in λn, its estimator may have jumps with respect to (w.r.t.) a small amount of perturbation of the observed data even when the solution is unique. Finally, if the error distribution does not have many extreme values, then the RLAD is not an efficient estimator: the asymptotic efficiency of the RLAD estimator is just 63.7% compared with the Lasso estimator under the Gaussian error distribution. In contrast, the proposed robust Lasso model has advantages in terms of generality and flexibility. Combining `1 and `2 losses yields a robust solution 9 1.2. Lasso-Type Sparse Linear Regression and the combination weight can be tuned, either analytically or numerically estimated from data, to achieve the minimum asymptotic variance. Our asymptotic analysis also shows that under certain adaptation procedures and shrinkage conditions, the proposed approach is indeed model selection consistent. Meanwhile, for variables with non-zero coefficients, it will be shown that the proposed robust model has unbiased estimates and the variability of the estimates is stabilized compared with the ordinary Lasso estimator. Therefore, the oracle property in the sense of [51] is achieved for the proposed method. We further derive a parallel asymptotic analysis of an alternative robust version of the Lasso with the Huber loss function, a.k.a. the Huberized Lasso. To the best of our knowledge, currently there is no asymptotic theory for the Huberized Lasso, although [107] empirically studied its performance. For the Huberized Lasso, asymptotic normality and model selection consistency are established under much weaker conditions on the error distribution, i.e. no finite moment assumption is required for preserving similar asymptotic results as in the convex combined case. Thus, the Huberized Lasso estimator is well-behaved in the limiting situation when the error follows a Cauchy distribution, which has infinite first and second moments. The analysis result obtained for the non-stochastic design is extended to the random design case with additional mild regularity assumptions. These assumptions are typically satisfied for auto-regressive models. In solution (ii), we introduce two parameters in the proposed Bayesian Lasso model to separately control the model selection and estimation shrinkage in the spirit of [94] and [127]. In particular, we propose a Poisson prior on the model size and the Laplace prior on β to identify the sparsity pattern. Since the proposed joint posterior distribution is highly nonstandard and a standard MCMC is not applicable, we employ a reversible-jump MCMC (RJ-MCMC) algorithm to obtain the proposed Bayesian Lasso estimates by simultaneously performing model averaging and parame- ter estimation. It is worth emphasizing that, though RJ-MCMC algorithms have been 10 1.3. High-Dimensional Covariance and Precision Matrix Estimation developed in the literature before model selection and estimation purposes (e.g. [4] proposed a hierarchical Bayesian model and developed an RJ-MCMC algorithm for joint Bayesian model selection and estimation of noisy sinusoids; similarly [111] pro- posed an accelerated truncated Poisson process model for Bayesian QTL mapping), these methods are not intended for promoting sparse models whereas our model uti- lizes sparsity promoting priors in conjunction with the discrete prior on the model size. One advantage of the proposed model is that it requires no cross-validation for parameter tuning, which is computationally intensive and inevitable in the Lasso to determine the optimal parameters. 1.3 High-Dimensional Covariance and Precision Matrix Estimation 1.3.1 Challenges In the second part of my thesis, I focus on the estimation of large covariance Σ and precision matrices Ω = Σ−1, when the number of observations is far fewer than the number of parameters in the matrix (Chapter 4 and 5). Estimation of the covariance and precision matrices for high-dimensional datasets is attracting increasing recent attention [10, 11, 20–22, 36, 47, 80]. It is challenging because: (i) there are p(p+1)/2 unknown parameters to estimate from n observations when p n; (ii) Σ (hence Ω) is intrinsically positive definite. The estimation problem was partially motivated by many modern high-throughput devices that make huge scientific data available to us. While there are many potential practical applications for this method, a motivating application of this work is brain effective connectivity modeling using fMRI data, where the goal is to infer the connectivity networks, represented by non-zero entries in Ω, of a large number of (spatially-fixed) brain regions-of-interests (ROIs). In particular, some of the proposed models and algorithms have been further applied to learn the brain connectivity networks for Parkinson’s disease (PD) [35], the second 11 1.3. High-Dimensional Covariance and Precision Matrix Estimation most common neuro-degenerative disorder in Canada. Studying brain connectivity is crucial in understanding brain functioning and can provide significant insight into the pathophysiology of neurological disorders such as PD. Based on prior neuroscience knowledge, the connections between brain regions generally can be considered a priori to form a sparse network. Conventional statistical signal processing tools and mathematical models could fail at handling those huge datasets, due to either theoretical or algorithmic reasons. Example applications of covariance estimation for a large number of variables include, and of course are not limited to, array signal processing [1, 64], hyperspectral image classification [9], construction of genetic regulatory networks from microarray pro- files [39] and brain functional MRI networks [91]. Suppose we have n data points Xn×p = {x1, · · · ,xn}T that are iid from a zero-mean, p-dimensional multivariate Gaussian N(0,Σ). The most natural estimator of the covariance matrix Σ is the unstructured sample covariance matrix 1 Σ?n = n −1 n∑ i=1 xix T i ∈ Rp×p2. (1.3) It has been well-known from the classical normal distribution theory that Σ?n is a “good” estimator of Σ when p is fixed and n→∞. Please see [3] for a thorough and recent discussion on this subject. Unfortunately, the tools and results from the classical theory fail to work when the dimensionality p grows as the data size increases, a well-known fact called the curse-of-dimensionality. For instance, from the eigen-structure perspective, random matrix theory predicts that a recentred and rescaled version of the largest eigenvalue 1We assume in this definition that the sample mean vector x̄ = 0. Statistical literature often uses Σ?n = (n − 1)−1 ∑ i(xi − x̄)(xi − x̄)T = (n − 1)−1(X − X̄)T (X − X̄) where X̄ is the matrix stacking x̄ n-times. These two definitions, however, are asymptotically equivalent by noting that they have the same limiting spectral law and X−X̄ = (I−n−111T )X which implies that ‖X−X̄‖ ≤ ‖I − n−111T ‖‖X‖ = ‖X‖ since the largest singular value of (I − n−111T ) is 1. 2In Chapter 4 and 5, we assume the samples take values in Rp and thus use xTi instead of the conjugate transpose x∗i . Nonetheless, we shall see from the concluding remarks of Chapter 5 that nothing prevents the obtained results extending to Cp. 12 1.3. High-Dimensional Covariance and Precision Matrix Estimation of Σ?n for a certain class of Σ has a Tracy-Widom limiting law, when p/n ≤ 1 as n and p both go to infinity [46]. Therefore in particular, it is suggested that the vanilla principle component analysis (PCA) is not suitable when a large number of variables are projected to lower dimensional orthogonal subspaces based on a limited number of observations [72, 73]. Consider, for example, the identity covariance matrix Σ with all eigenvalues being equal to 1. Asymptotic random matrix theory roughly states that the largest eigenvalue is λmax(Σ ? n) ∼= (1 + √ p/n)2 and the smallest eigenvalue is λmin(Σ ? n) ∼= (1 − √ p/n)2, for n/p approaching to some positive ratio with n and p both going to infinity. In this case, the curse-of-dimensionality is phenomenal in the sense that the spectrum of the sample covariance matrix is more spread than the spectrum of Σ, the Dirac δ mass at 1. A natural solution to mitigate this difficulty is to restrict our attention to subsets of covariance matrices with certain special structures or properties and meanwhile incorporate the regularization ideas into estimation. Sparsity is one commonly hy- pothesized condition and it seems to be realistic for many real-world applications. Considering certain sparse covariance matrices, simple banding [10], tapering [20], and thresholding [11] on Σ?n are shown to be consistent estimators for Σ. Surpris- ingly, some of these conceptually and computationally simple estimators are even shown to be optimal in the sense of minimax risk [20–22]. 1.3.2 Estimating Covariance Matrix Prior Arts Significant recent progress has been made in both theory and methodology devel- opment for estimating large covariance matrices. Regularization has been widely employed. Broadly speaking, regularized estimation of large covariance matrices can be classified into two major categories. The first category includes Steinian shrinkage- type estimators that shrink the covariance matrix to some well-conditioned matrices under different performance measures. For instances, Lediot and Wolf (LW) [82] 13 1.3. High-Dimensional Covariance and Precision Matrix Estimation proposed a shrinkage estimator by using a convex combination between Σ?n and p−1Tr(Σ?n)I and provided a procedure for estimating the optimal combination weight that minimizes the mean-squared errors and that is distribution-free. Chen et. al. [36] further extended the idea and improved the LW estimator through two strategies: one is based on the Rao-Blackwellization idea to condition on the sufficient statistics Σ?n and the other is to approximate the oracle by an iterative algorithm. Closed- form expressions of both estimators were given in [36]. More recently, Fisher and Sun [53] proposed using diag(Σ?n) as the shrinkage target with possibly unequal vari- ances. These shrinkage estimators are amenable for general covariance matrices with “moderate-dimensionality”. Here, by moderate-dimensionality, we mean that p grows nicely as n increases, e.g. p→∞ and p = O(nk) for some 0 < k ≤ 1. Estimators in the second category directly operate on the covariance matrix through operators such as thresholding [10], banding [11], and tapering [20]. Band- ing and tapering estimators are suitable for estimating covariance matrices where a natural ordering exists in the variables such as covariance structures in time-series. Banding simply sets the entries far away from the main diagonal to be zeros and keeps the entries within the band unchanged. Tapering is similar to banding, with the difference in that it gradually shrinks the off-diagonal entries within the band to 0. We can view banding as a hard-thresholding rule while tapering is a soft-thresholding rule, up to a certain unknown permutation [12]. In contrast, thresholding can deal with general permutation-invariant covariance matrices and introduce sparsity with- out requiring additional structures. These estimators are statistically consistent if certain sparsity is assumed and the dimensionality p grows at any sub-exponential rate of n, which allows much larger covariance matrices be estimable. In fact, it is further known that tapering and thresholding estimators are minimax [20–22]. The rate-optimality under the operator norm is not true for the banding estimator in [11] and it was shown that tapering is generally preferred to banding [20]. However, it is worth mentioning that, when the assumed sparsity condition is invalid, all the above 14 1.3. High-Dimensional Covariance and Precision Matrix Estimation estimators in the second category become sub-optimal. Our Contributions Despite recent progress on large covariance matrix estimation, there has been rela- tively little fundamental theoretical study on comparing the shrinkage-category and tapering-category estimators. To fill this gap, we first study the risks of shrinkage estimators and provide a comparison of risk bounds between shrinkage and taper- ing estimators. Further, motivated by the observed advantages and disadvantages of shrinkage and tapering estimators under different situations, to properly estimate gen- eral and high-dimensional covariance matrices, we propose a shrinkage-to-tapering es- timator that combines the strengths of both shrinkage and tapering approaches. The proposed estimator has the form of a general shrinkage estimator with the crucial difference that the shrinkage target matrix is a tapered version of Σ?n. By adaptively combining Σ?n and a tapered Σ ? n, the proposed shrinkage-to-tapering oracle (STO) estimator inherits the optimality in the minimax sense when sparsity is present (e.g. AR(1)) and in the minimum mean-squared error (MMSE) sense when sparsity is ab- sent (e.g. fractional Brownian motion). Therefore, the proposed estimator improves upon both shrinkage and tapering estimators. A closed-form of the optimal combi- nation weight is given and a STO approximating (STOA) algorithm is proposed to determine the oracle estimator. 1.3.3 Estimating Precision Matrix Prior Arts Estimation of Ω is a more difficult task than estimating Σ because of the lack of natural and pivotal estimators as Σ?n when p > n. Nonetheless, accurately estimating Ω has important statistical meanings. For example, in Gaussian graphical models, a zero entry in the precision matrix implies the conditional independence between the corresponding two variables. Further, there are additional concerns in estimating Ω 15 1.3. High-Dimensional Covariance and Precision Matrix Estimation beyond those we have already seen in estimating large covariance matrices. First, since Σ?n is a natural estimator of Σ, so is (Σ ? n) −1 of Ω. However it is obvious that Σ?n is invertible only if p < n. Even worse, assuming Σ ? n is invertible, (Σ ? n) −1 still does not converge to Ω in the sense of eigen-structure when p/n→ c > 0 [46, 72]. Secondly, it is known that, under mild hypotheses and for a certain class of sparse matrices, applying simple hard thresholding to Σ?n yields a consistent [11] and optimal estimator of Σ in the sense of minimax risk ([21, 22]). Therefore, [Tt(Σ ? n)] −1, where Tt is the thresholding operator with cutoff t, is a natural estimator of Ω. Indeed, [21] has showed that this estimator is rate-optimal under the matrix L1 norm when minimaxity is considered. Nonetheless, it is possible that [Tt(Σ ? n)] −1 fails to preserve sparsity (including the sparsity measure in terms of strong `q-balls, see definition in (5.2)), because a sparse matrix does not necessarily have a sparse inverse which plays a central role in Gaussian graphical models. Hence, the natural estimator [Tt(Σ ? n)] −1 of Ω proposed in [21] can be unsatisfactory. Thirdly, state-of-the-art precision matrix estimation procedures are essentially based on penalized likelihood maximization or constrained error minimization ap- proaches, e.g. CLIME [18], SPICE [108], graphical Lasso [56] and variants [7], adap- tive graphical Lasso [50], SCAD [80], and neighborhood selection [95, 126]. They are optimization algorithms with different objective functions. They have, however, a common structural feature in the objective functions: one term is the goodness-of-fit and the other term measures the model size which is often formulated by sparsity promoting penalties such as matrix 1-norm, SCAD, etc. The interior point method is standard for solving the optimization problems; but it is computationally infeasi- ble when the dimensionality is large. Moreover, its high computational cost can be magnified by the parameter tuning procedure such as cross-validation. It is worth mentioning that the graphical Lasso [56] can be solved in a looped LAR fashion [45] and thereby its computational cost to estimate Ω ∈ Sk (see Eqn. (5.1) for defini- tion) is equivalent to solving a sequence of p least squares problems, each of which 16 1.4. Thesis Outline has the complexity of O(p3) in terms of basic algebraic operations over some rings, e.g. the real or complex numbers. Therefore, the computational complexity of the graphical Lasso is O(p4). Moreover, since the graphical Lasso has an outer loop for sweeping over the p columns, it is empirically observed that the graphical Lasso is problem-dependent and its computational cost can be prohibitively high when the true precision matrix is not extremely sparse. Our Contributions In light of these challenges in estimating the precision matrix when p n, we propose a new easy-to-implement estimator with attractive theoretic properties and computa- tional efficiency. The proposed estimator is constructed on the idea of the finite Neu- mann series approximation and constitutes merely matrix multiplication and addition operations. The proposed estimator has a computational complexity of O(log(n)p3) for problems with p variables and n observations, representing a significant improve- ment upon the aforementioned optimization methods. The proposed estimator is promising for ultra high-dimensional real-world applications such as gene microarray network modeling. We prove that, for the class of approximately inversely closed precision matri- ces, the proposed estimator is consistent in probability and in L2 under the spectral norm. Moreover, its convergence is shown to be rate-optimal in the sense of minimax risk. We further prove that the proposed estimator is model selection consistent by establishing a convergence result under the entry-wise ∞-norm. 1.4 Thesis Outline Now, we outline the structure of the rest of this thesis. Chapter 2 [31] presents our research on the robust Lasso models and their asymp- totic properties. More specifically, we propose a robust version of the Lasso and derive the limiting distribution of its estimator, from which the estimation consistency can 17 1.4. Thesis Outline be immediately established. We further prove the model selection consistency of the proposed robust Lasso under an adaptation procedure for the penalty weight. Meanwhile, a parallel asymptotic analysis is performed for the Huberized Lasso, a previously proposed robust Lasso [107]. We show that the Huberized Lasso estima- tor preserves similar asymptotics even with a Cauchy error distribution. Therefore, our analysis shows that the asymptotic variances of the two robust Lasso estimators are stabilized in the presence of large variance noise, compared with the unbounded asymptotic variance of the ordinary Lasso estimator. Finally, the asymptotic analysis from the non-stochastic design is extended to the case of random design. Chapter 3 [32] presents our research on the Bayesian Lasso model. In this work, we first utilize the Bayesian interpretation of the Lasso model and propose several hyper-priors to extend the Lasso to the fully Bayesian paradigm. Since the proposed Bayesian Lasso contains discrete and continuous (hyper-)parameters that simulta- neously control the model size and parameter shrinkage, we construct a provably convergent reversible-jump MCMC algorithm to obtain its numeric estimates. We use simulations to show the improved performance of the proposed Bayesian Lasso model in terms of estimation error and pattern recovery. Chapter 4 [33] presents our research on the estimation of high-dimensional covari- ance matrices based on a small number of iid Gaussian samples. In this chapter, we first study the asymptotic risk of the MMSE shrinkage estimator proposed by [36] and show that this estimator is statistically inconsistent for a typical class of sparse matrices that often appear as the covariance of auto-regressions. We then propose a shrinkage-to-tapering oracle estimator that improves upon both shrinkage and taper- ing estimators. We further develop an implementable approximating algorithm for the proposed estimator. Chapter 5 [31] presents our research on the estimation of high-dimensional preci- sion matrices. In this research, we propose an efficient algorithm involving only matrix multiplication and addition operations based on the truncated Neumann series repre- 18 1.4. Thesis Outline sentation. The proposed algorithm has a computational complexity of O(log(n)p3) in terms of basic algebraic operations over real or complex numbers. We prove that, for the class of approximately inversely closed precision matrices, the proposed estimator is consistent in probability and in L2 under the spectral norm. Moreover, its con- vergence is shown to be rate-optimal in the sense of minimax risk. We further prove that the proposed estimator is model selection consistent by establishing a conver- gence result under the entry-wise ∞-norm. Finally, we apply the proposed method to learn functional brain connectivity from frontal cortex directly to the subthalamic nucleus based on fMRI data. Chapter 6 [35] presents an application of the group robust Lasso model to an fMRI group analysis. In this work, we consider incorporating sparsity into brain con- nectivity modeling to make models more biologically realistic and performing group analysis to deal with inter-subject variability. To this end, we propose a group robust Lasso model by combining advantages of the group Lasso and robust Lasso model developed in Chapter 2. The group robust Lasso is applied to a real fMRI data set for brain connectivity study in Parkinson’s disease, resulting in biologically plausible networks. Chapter 7 briefly summarizes the major contributions of the thesis and provides concluding remarks. A number of future topics are discussed. Appendices include the notations used in the thesis, proofs of lemmas and theo- rems stated in the thesis body. 19 Chapter 2 Robust Lassos and Their Asymptotic Properties 2.1 Introduction 2.1.1 Sparsity-Promoting Linear Models In this chapter, we consider the problem of estimating the coefficient vector in a linear regression model, defined as y = Xβ + e. (2.1) Here X is the n× p design matrix which can either be non-stochastic or random. As per convention, rows of X represent the p-dimensional observations and columns of X represent the predictors. y is the response vector and β is the coefficient vector to be estimated. We regard e as a column vector, and use e∗ to denote its conjugate transpose. The random measurement error vector e = (e1, · · · , en)∗ is assumed to be iid with zero mean. Here we do not have to generally assume that the error possesses a finite second moment σ2 for each component. Recovering the true sparse coefficients in the presence of random errors with large variability is of primary interest of this chapter. For a sparse regression problem where p is large but the number of true predictors with non-zero coefficients is small, the traditional least squares (LS) estimator for the full model in (2.1) may not be feasible. Even if the LS estimator exists and is unique, it can have an unacceptably high variance since including unnecessary predictors can 20 2.1. Introduction significantly degrade estimation accuracy. More importantly, the LS estimator cannot be naturally interpreted as extracting (sparse) signals from the ambient linear system, which can make subsequent inferences difficult. In fact, for the case of p being large, overfitting is a serious problem for all statistical inferences based on the full model since the data always prefer models with a greater number of coefficients. Therefore, sparse representation for a large linear system is crucial for extracting true signals and improving prediction performance. Robust model selection for recovering sparse representations is a problem of interest for both theory and applications. Depending on the particular research interest and application, the problem of recovering sparse representations can be formulated accordingly to different scenar- ios depending upon the relative magnitudes of p and n. One scenario is the over- determined case, i.e. p < n, such as p being fixed and n → ∞. Another scenario of great interest is the under-determined case, i.e. p > n. For instance, the typical setup in compressed sensing is p n with n being fixed for a deterministic X [26]. In this chapter, following [51, 77, 122, 123], we assume the classical over-determined case, such that p < n and p is fixed, and formulate the sparse linear regression model accordingly. In particular, to study the asymptotic performance, we use the setting as in [77, 122] that n→∞ and p is presumably large. Though differently formulated, these two scenarios are in fact related, and theoret- ical results from one scenario have implications for the other. Wainwright showed that asymptotic results of the Lasso estimator (defined in (2.2)) in the classical scenario [77] continue to be true in the double-asymptotic scenario (both n and p approach to infinity) [121]. Connections between these two scenarios can also be found in [130], and both scenarios are areas of active research. There has been extensive development in the theory and application for under-determined systems [23, 24, 26–28, 42, 43]. For instance, sparse recovery for under-determined linear systems is vital to the area of compressed sensing [24–26, 42]. Additionally, identifying non-zero coefficients in a large over-determined linear system subject to random errors is also pertinent to 21 2.1. Introduction the signal and image processing and machine learning communities. For examples, face recognition using the sparse PCA, where contaminated face images are recovered by only a few principal components representing different facial features [123]. In [69], face images can be represented by learnt features that are basis vectors with sparse coefficients in the matrix factorized domain. In [119], sparse mAR model with a fixed number of brain ROIs whose intensity signals are measured by magnetic res- onance scans over time is used to model brain connectivity. The fused Lasso [57], which is closely related to the total variation denoising in signal processing, is used to reconstruct images with sparse coefficients and gradients. We note that our approach differs from the typical scenario in compressed sensing since they serve different purposes. Compressed sensing research mainly focuses on the theory and algorithms required to (exactly or approximately) discover a sparse representation [24–26, 42], e.g. determining the value of n to exactly recover sparse signals in the absence of noise. Since error bounds on the estimates suggest that the approximation quality deteriorates linearly with the dispersion of errors [41], any sparse approximation, β̂ ≈ β, will be inaccurate when the underlying error distribution has heavy tails. In this paper, we concentrate on developing robust estimators to recover the true sparse coefficients in the presence of large noise. We also study the properties of the proposed estimators such as estimation and model selection consistency. Therefore, our analysis is put in an asymptotic framework. For the purpose of variable/model selection in linear regression problems, a variety of methodologies have been proposed. Though different methods employ different selection criteria, all approaches share a common feature: penalizing models with a larger number of coefficients. There is no general agreement on which cost function and penalization criterion type is optimal. However, prior approaches can be classified according to the choice of loss/cost and penalization functions: 1. `2 loss with various functional forms of the penalty: `2 loss (also referred as LS cost function, the squared error loss) is widely employed. Early approaches 22 2.1. Introduction used the `2 loss coupled with penalties proportional to the model size, i.e. the `0 norm of the coefficient vector. For example, the Akaike Information Crite- rion (AIC) [2], Bayesian Information Criterion (BIC) [109], and Risk Inflation Criterion (RIC) [54] are popular model selection criteria of this type. Since these penalized LS cost functions are combinatorial in nature, estimating and comparing all possible models may become computationally impractical, par- ticularly with high dimensional models. Therefore, more efficient and practical model-shrinkage tools have been proposed, such as ridge regression which com- bines the LS estimator with the `2 penalty for the coefficients [68]. Although ridge regression can reduce the model size in terms of the numeric magnitudes of estimates, it shrinks all coefficients and thus cannot perform model selec- tion and parameter estimation at the same time. The least absolute shrinkage and selection operator (Lasso) proposed by [115] is a popular and useful tech- nique for simultaneously performing variable/model selection and estimation. Specifically, the Lasso estimator of Tibshirani is defined as β̂n = arg min u∈Rp ( 1 n n∑ i=1 (yi − x∗iu)2 + λn n p∑ j=1 |uj|γ ) , (2.2) where xi is the i th-row of X and γ = 1. This is the same definition (1.2) used in Chapter 1 and the properties of the Lasso estimator has been thoroughly discussed therein. In addition, we remark that [51] proposed a non-convex penalized LS cost function, the Smoothly Clipped Absolute Deviation (SCAD) model, which can avoid the over-penalization problem of the Lasso. 2. `∞ loss with the `1 penalty: As an alternative to the `2 cost function, the Dantzig selector [27] combines the `∞ error measurement criterion and the `1 penalty: min u∈Rp ‖u‖`1 subject to ‖X∗(y −Xu)‖`∞ ≤ λn. (2.3) A small λn ensures a good approximation and the minimization yields a maxi- 23 2.1. Introduction mized sparsity. The Dantzig selector can be efficiently solved by recasting as a linear programming problem. 3. Robust losses with the `1 penalty: The lack of robustness of the `2 loss is well- known. Since the `1 loss function is more robust to outliers, the corresponding regression model leads to a robustified version of the LS regression. This re- gression model is called the Least Absolute Deviation regression (LAD) in the literature. Unfortunately, in the context of linear model selection, robustness has not received much attention compared to say, the Lasso. This is largely due to the difficulty of handling a non-differentiable `1 loss function. To our best knowledge, there are only a few studies considering robust losses. The regular- ized LAD (RLAD) model for robust variable selection has been proposed which can be recasted as a linear program [123]. The RLAD adopts the `1 loss coupled with the `1 penalty. An alternative robust version of Lasso can be formed by using the Huber loss with the `1 penalty to create a Huberized Lasso which is robust to contamination [107]. 2.1.2 Summary of Present Contributions We shall propose a robust Lasso and mainly focus on the asymptotic theory of its es- timator since there is no general theory that guarantees the consistency of a selection criterion. Our asymptotic analysis put forth here shows that under certain adapta- tion procedure and shrinkage conditions, the proposed estimator is model selection consistent. Meanwhile, for variables with non-zero coefficients, it will be shown that the proposed robust model has unbiased estimates and the variability of the estimates is stabilized compared with the ordinary Lasso estimator. Therefore, the oracle prop- erty [51] is achieved for the proposed method. Now, we summarize our contribution as follows: 1. We propose using a convex combined loss of `1 (LAD) and `2 (LS), rather than the pure LS cost function, coupled with the `1 penalty to produce a robust 24 2.1. Introduction version of the Lasso. Asymptotic normality is established, and we show that the variance of the asymptotic normal distribution is stabilized. Estimation consistency is proved at different shrinkage rates for {λn} and further proved by a non-asymptotic analysis for the noiseless case. 2. Under a simple adaptation procedure, we show that the proposed robust Lasso is model selection consistent (defined in (2.14)), i.e. the probability of the selected model to be the true model approaches to 1. 3. As an extension of the asymptotic analysis of our proposed robust Lasso, we study an alternative robust version of the Lasso with the Huber loss func- tion, the Huberized Lasso. To the best of our knowledge, currently there is no asymptotic theory for the Huberized Lasso, although [107] empirically studied its performance. For the Huberized Lasso, asymptotic normality and model selection consistency are established under much weaker conditions on the error distribution, i.e. no finite moment assumption is required for preserving sim- ilar asymptotic results as in the convex combined case. Thus, the Huberized Lasso estimator is well-behaved in the limiting situation when the error follows a Cauchy distribution, which has infinite first and second moments. 4. The analysis result obtained for the non-stochastic design is extended to the random design case with additional mild regularity assumptions. These as- sumptions are typically satisfied by auto-regressive models. 2.1.3 Organization of the Chapter The rest of the chapter is organized as follows. We introduce the proposed robust version of the Lasso with convex combined loss in Section 2.2. Its asymptotic behavior is then studied and compared with the Lasso. Section 2.3 defines an adaptive robust Lasso and its model selection consistency is proved. Section 2.4 concerns the Lasso with the Huber loss function and its asymptotic behavior is analyzed. Section 2.5 25 2.2. A Convex Combined Robust Lasso extends the analysis results from the non-stochastic design to the random design under additional mild regularity conditions. In Section 2.6, a simulation study is used to support the theoretical results found in previous sections. 2.2 A Convex Combined Robust Lasso 2.2.1 A Robust Lasso with the Convex Combined Loss As discussed earlier, the `2 loss in the Lasso model is not robust to heavy-tailed error distributions and/or outliers. This indicates that the Lasso is not an ideal goodness- of-fit measure criterion in the presence of noise with large variance. In order to build a robust, sparsity-promoting model, we propose a flexible robust version of the Lasso, where the estimator β̂n is defined as β̂n = arg min u∈Rp ( 1 n n∑ i=1 L(u; yi,xi) + λn n ‖u‖`1 ) (2.4) where the cost function, L(u; yi,xi) = δ (yi − u∗xi)2 + (1− δ) |yi − u∗xi| , is a convex combination of the `2 and `1 losses of yi − u∗xi, and δ ∈ [0, 1]. Note that this reduces to the traditional Lasso if δ = 1, and it reduces to the RLAD if δ = 0. 2.2.2 Asymptotic Normality and Estimation Consistency In order to ensure that there is no strong dependency among the predictors (columns of X), namely model identifiability, we need a regularity assumption on the design matrix. Here, we assume the following classical conditions: Non-stochastic design assumptions 1. (Design matrix assumption) The Gram matrix Cn = n −1∑n i=1 xix ∗ i converges 26 2.2. A Convex Combined Robust Lasso to a positive definite matrix C as n→∞. 2. (Error assumption) (a) The error has a symmetric common distribution w.r.t. the origin. So Eei = 0 and the median of ei is 0. (b) ei has a continuous, positive probability density function (p.d.f.) f w.r.t. the Lebesgue measure in a neighborhood of 0. (c) ei possesses a finite second moment σ 2. Remark 1. [77] and [130] made an additional assumption on the design matrix for the fixed p case: n−1 max 1≤i≤n x∗ixi → 0 (2.5) as n → ∞, i.e. max1≤i≤n |xi| = o( √ n). However, we shall show that this regularity condition is unnecessary and it is actually a direct consequence of assumption 1. Please refer to Lemma A.2.3 for a proof. Note that (2.5) has already been observed by [105]. Our first main theorem described below is to establish the asymptotic normality of the robust Lasso estimator. Since the loss function is non-differentiable, the usual Taylor expansion argument fails and a more subtle argument is required. Theorem 2.2.1. Under the above assumptions 1 and 2, if λn/ √ n → λ0 ≥ 0, then √ n(β̂n − β)⇒ argmin(V ) where V (u) = (δ + (1− δ)f(0))u∗Cu+ u∗W + λ0 p∑ j=1 [ujsgn (βj) 1(βj 6= 0) + |uj|1(βj = 0)] and W ∼ N(0, ((1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10)C). An immediate consequence of Theorem 2.2.1 is: 27 2.2. A Convex Combined Robust Lasso Corollary 2.2.2. If λn = o( √ n), then β̂n is √ n-consistent. A couple of observations can be made from Corollary 2.2.2. If δ = 0, then we have an asymptotic variance of √ n(β̂n−β) equal to 14f(0)2C−1, which is the reduced case of using a pure `1 loss without penalization [105]. When compared with the asymptotic variance of the ordinary Lasso σ2C−1 (c.f. Theorem 2 in [77]) which is unbounded when σ2 goes to infinity, our estimator has a finite asymptotic variance when δ is chosen carefully. As long as the value of δ2σ2 is well controlled, the corresponding estimator can be stabilized asymptotically. Hence, it is desirable to seek a δ ∈ [0, 1] which yields the minimum of the asymptotic variance. Assume for now that we know the error distribution and let us consider the asymptotic variance in (10). Let x = δ−1 − 1 ≥ 0 for 0 < δ ≤ 1, and define v(x) = x2 + 4σ2 + 4M10x 4(1 + f(0)x)2 = 1 4 × x+ 4σ 2x−1 + 4M10 f(0)2x+ x−1 + 2f(0) . (2.6) Ignoring the terms 4M10 and 2f(0) for a moment, it is easy to observe from the arithmetic-geometric mean inequality that the numerator of v(x) is minimized at 2σ and the denominator at f(0)−1. If 2σ > 1/f(0), then the numerator of v(x) will dominate its denominator. Hence, v(x) is minimized when x → ∞, i.e. δ → 0. In another word, the convex combined robust Lasso is reduced to the RLAD for the case of having noise with large variance, to achieve the optimal asymptotic variance. Similarly, if 2σ < 1/f(0), the denominator dominates the numerator as x → 0, i.e. δ → 1. The optimal weight of the robust Lasso corresponds to the special case of the ordinary Lasso when the noise has a moderate variance. Nevertheless, taking also the terms 4M10 and 2f(0) into account, the optimal δ may lie in the interval of (0, 1). Hence, our robust Lasso provides better flexibility. In practice, the error distribution is usually unknown and thus the analytical form of the optimal weight is unavailable. Fortunately, we can still estimate the convex combined weight from the data by allowing the weight to be data dependent 28 2.2. A Convex Combined Robust Lasso δn = δ({xi, yi}i∈{1,··· ,n}). For example, an intuitive choice of measuring the spreadness of empirical errors is to use a renormalized quantity such as δn = (σ̂ 2 + 1)−1/2, (2.7) where σ̂2 = n− p−1∑ni=1 (yi − x∗i β̂LS)2 and β̂LS is the LS estimator for the linear model. If the noise variance is large, then δn is likely to concentrate within a small neighborhood of zero, and thus the robust Lasso behaves more like the RLAD. On the contrary, the `2 component can dominate the `1 if the error distribution has a small variance. When σ2 is large, the robust Lasso estimator with the data-driven weight in (2.7) has an asymptotic variance which is not larger than 9 4f(0)2 C−1, as shown in the following Corollary 2.2.3. Corollary 2.2.3. Suppose λn/ √ n → λ0 ≥ 0 and choose δn as in (2.7). Then √ n(β̂n − β)⇒ argmin(V ) where V (u) = (δ + (1− δ)f(0))u∗Cu+ u∗W + λ0 p∑ j=1 [ujsgn (βj) 1(βj 6= 0) + |uj|1(βj = 0)] and W ∼ N (0, vδC) , with vδ = ( 1− √ 1 σ2 + 1 )2 + 4σ2 σ2 + 1 + 4 √ 1 σ2 + 1 ( 1− √ 1 σ2 + 1 ) M10. Remark 2. By Jensen’s inequality, we have M210 = (E|ei|)2 ≤ Ee2i = σ2. So vδ ≤ [( 1− √ 1 σ2 + 1 ) + 2σ√ σ2 + 1 ]2 ≤ 9. 29 2.2. A Convex Combined Robust Lasso In light of the √ n-rate convergence, we actually allow the sequence of {λn} to grow faster while meantime preserving the estimation consistency, as demonstrated in the following Theorem 2.2.4. Theorem 2.2.4. Under the assumptions 1 and 2, if λn/n → λ0 ≥ 0, then β̂n P→ argmin(Z) where Z(u) = δ(u− β)∗C(u− β) + δσ2 + (1− δ)r + λ0 ‖u‖`1 , (2.8) and r = limn→∞ n−1 ∑n i=1E|yi − u∗xi| < ∞. In particular, if λn = o(n), then argmin(Z) = β so that β̂n is a consistent estimator of β. 2.2.3 A Bound on MSE for the Noiseless Case We have seen the asymptotic variance of the robust Lasso estimator, which does not necessarily hold for the case of having a finite sample size n. A more interesting question is that, given a fixed design matrix Xn×p and the assumed linear model, how accurately we can recover the true β using the robust Lasso? In this section, we would like to answer this question under a simpler scenario, i.e. in the noiseless case. This explicit estimation error bound due to the bias provides an implication on the asymptotic behavior of the robust Lasso estimator in the presence of noise. Indeed, statistical common sense tells us that the variance of the robust Lasso estimator can be smaller than the unpenalized case because of the bias-variance trade-off. Therefore, the mean squared estimation loss is expected to be controlled at the order of the LS+LAD, provided the bias term is small. More specifically, our observation is that, under the shrinkage rate λn = o(n) and certain assumptions on X, the proposed robust Lasso can accurately estimate β in terms of `2 loss in the absence of noise. Assume that β is S-sparse, i.e. |supp (β)| = 30 2.2. A Convex Combined Robust Lasso |A| = S. Let φmin(S) = min T≤S inf |u|≤T,u6=0 u∗Cnu u∗u (2.9) φmax(S) = max T≤S sup |u|≤T,u6=0 u∗Cnu u∗u (2.10) be the restricted extreme eigenvalues of the submatrices of Cnwith the number of columns being less than or equal to S. We assume a similar incoherent design condi- tion as in [96]. That is, we assume there is a positive integer S0 such that S0φmin(S0) Sφmax(p− S0) > 16. (2.11) This condition measures the linear independency among restricted sets of the columns of X. A large value of the LHS in (2.11) prevents degeneracy of the restricted columns of X. With this incoherent design hypothesis, we can show that the `2 estimation loss decays to 0 if {λn} grow at a proper rate. Proposition 2.2.5. Assume σ2 = 0 and β is S-sparse. Suppose that the incoherent design condition (2.11) holds. Then the robust Lasso estimator β̂n for δ ∈ (0, 1] defined in (2.4) satisfies ∥∥∥β̂n − β∥∥∥ `2 ≤ λn n √ S δD0 − 1− δ δ √ nD0 , (2.12) where D0 = 1− 4 √ Sφmax(p− S0) S0φmin(S0) . (2.13) Remark 3. In the noiseless case, Proposition 2.2.5 suggests that if limn→∞ λn/n = 0, then ∥∥∥β̂n − β∥∥∥ `2 → 0 as n → ∞. This condition on the shrinkage rate is exactly the one assumed in Theorem 2.2.4. Hence, both the asymptotic and non-asymptotic analysis show that λn = o(n) is sufficient for the conclusion that β̂n is consistent. 31 2.3. The Adaptive Robust Lasso and Its Model Selection Consistency 2.3 The Adaptive Robust Lasso and Its Model Selection Consistency We have established the estimation consistency of the robust Lasso so far. However, in many scenarios, it is also desirable to have the model selection consistency, defined as P ( supp ( β̂n ) = supp (β) ) → 1 (2.14) as n → ∞. Note that neither estimation consistency nor consistency in the `2 norm necessarily implies the model selection consistency. Consider, for a counterexample, that β̂n[j] = n −1 for β[j] = 0. In terms of choosing a sequence of shrinkage tuning parameters {λn}n∈N, [86] and [95] showed that the ordinary Lasso has a conflict between the consistency for model selection and optimal prediction. As a solution to achieve both estimation and model selection consistency, the adaptive Lasso [132] was proposed and its model selection consistency and asymptotic normality under certain rate of shrinkage were proved. To extend the idea of the adaptive Lasso [132] to our proposed robust Lasso, we define the adaptive robust Lasso as β̂n = arg min u∈Rp ( 1 n n∑ i=1 L(u; yi,xi) + λn n p∑ j=1 ŵj |uj| ) , (2.15) where ŵ = (ŵ1, · · · , ŵp)∗ is a vector of adaptive weights, which allow unequal penal- ties for the coefficients. For example, we can take ŵ = 1/ ∣∣∣β̂LS∣∣∣γ for some γ > 0. Let A = {j : βj 6= 0} and An = {j : β̂n[j] 6= 0}. By definition, the estimator {β̂n} is said to be consistent for model selection if and only if P (An = A) → 1 as n → ∞. Now following the similar argument as in [132], we have the following theorem showing the model selection consistency of the adaptive robust Lasso. Theorem 2.3.1. Suppose assumption 1 and 2 are satisfied. Let λn = o( √ n) and λnn (γ−1)/2 → ∞ for some γ > 0. Then the adaptive robust Lasso defined in (2.15) 32 2.4. The Huberized Lasso with ŵ = 1/ ∣∣∣β̂LS∣∣∣γ has the following properties: 1. Asymptotic normality for the true non-zero coefficients, i.e. √ n ( β̂n[A]− β[A] ) ⇒ N ( 0, (1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10 4[δ + (1− δ)f(0)]2 C −1 11 ) . 2. Model selection consistency if ‖xj‖`1 = O( √ n) for all j /∈ A. Remark 4. The additional condition, ‖xj‖`1 ≤ K √ n, for the model selection consis- tency is not trivial. It can be implied, for example, by ∑n i=1 ∣∣∣ xi√n∣∣∣ 1(an ∣∣∣ xi√n∣∣∣ > τ)→ 0 for every τ > 0 and an = (∑n i=1 ∣∣∣ xi√n∣∣∣3)−1/2. 2.4 The Huberized Lasso For the robustness purpose, as an alternative for using a convex combination of `1 and `2 losses, we can use the Huber loss function and thus the corresponding `1-penalized model is called the Huberized Lasso [107]. The Huberized Lasso is defined as β̂ H n = argmin u∈R ( 1 n p∑ i=1 L(u; yi,xi) + λn n ‖u‖`1 ) , (2.16) where L(u; yi,xi) = (yi − u ∗xi)2 if |yi − u∗xi| ≤ δ, 2δ|yi − u∗xi| − δ2 if |yi − u∗xi| > δ. The Huberized Lasso enjoys the everywhere differentiability which is not true for the convex combined loss. Although the Huberized Lasso has already been used as a robustified version of the Lasso, currently there is no asymptotic theory for it. Here, we expect the Huberized Lasso to have similar asymptotic properties to the case of the convex combination loss. We first establish the asymptotic normality of the Huberized Lasso. It is worth mentioning that the proof details are considerably more complicate than the convex combined loss case. 33 2.4. The Huberized Lasso Remarkably, as shown in Theorem 2.4.1 below, we note that no condition is re- quired on the finiteness of the variance or even the first moment for the error distribu- tion in order to achieve the asymptotic normality (and model selection consistency) for the Huberzied Lasso estimator (and its adaptive version). In other words, assumption 2(c) is not required, and the minimal set of assumptions only include the symmetry of error distribution and the continuity of its p.d.f. around the transition points ±δ. Therefore, the asymptotic normality and model selection consistency results are still valid for the Cauchy errors, whose first and second moments are infinite. Theorem 2.4.1. Under the assumptions 1, 2(a), and 2(b), if λn/ √ n→ λ0 ≥ 0 and f is continuous at ±δ, then √n(β̂Hn − β)⇒ argmin(V ) where V (u) = K0δu ∗Cu+ 2u∗W + λ0 p∑ j=1 [ujsgn (βj) 1(βj 6= 0) + |uj|1(βj = 0)] (2.17) and W ∼ N (0, (δ2M0δ +K2δ)C) . Here, the assumption 2(b) is understood as the continuity of f around ±δ. Corollary 2.4.2. If λn = o( √ n), then β̂ H n is √ n-consistent. Proof. For λ0 = 0, V (u) = K0δu ∗Cu+ 2u∗W is minimized at argmin(V ) = −C −1W K0δ ∼ N ( 0, δ2M0δ +K2δ K20δ C−1 ) . We can give a numerical example of the stabilized asymptotic variance of the Huberized estimator when the error is Cauchy distributed with zero mean and scale 34 2.4. The Huberized Lasso parameter s: f(ei) = s pi(s2 + e2i ) . Take δ = 1, then s = 3 gives the asymptotic variance 20.53 while s = 1 stabilizes the variance to 2.55! Similarly, the adaptive Huberized Lasso is defined as in (2.15) with the change of the loss function. The following Theorem 2.4.3 shows that the adaptive Huberized Lasso is model selection consistent. Since the proof almost follows the same line of Theorem 2.3.1, we omit the details here. Theorem 2.4.3. Suppose assumptions 1, 2(a), and 2(b) are satisfied. Let λn = o( √ n) and λnn (γ−1)/2 → ∞ for some γ > 0. Then the adaptive Huberized Lasso defined in (2.15) with ŵ = 1/ ∣∣∣β̂LS∣∣∣γ has the following properties: 1. Asymptotic normality for the true non-zero coefficients, i.e. √ n ( β̂ H n [A]− β[A] ) ⇒ N ( 0, δ2M0δ +K2δ K20δ C−1 ) . (2.18) 2. Model selection consistency. Remark 5. The adaptive weight used in the Huberized Lasso needs to be adjusted to β̂LAD in the case that the least squares estimator is not guaranteed to be a consistent estimator, for instance, when the error is Cauchy. Then the theorem continues to be true due to the fact that β̂LAD is also a √ n-consistent estimator of β under assumptions 1), 2a), and 2b) [105]. Remark 6. The additional assumption for the model selection consistency of the robust Lasso with the convex combined loss, i.e. ‖xj‖`1 = O( √ n) for all j /∈ A, is not required for that of the Huberized Lasso. The difference lies in the fact that the Huberized Lasso objective function is differentiable everywhere. Its derivative agrees with the derivative of the Lasso on [−δ, δ] and is less than that of the Lasso otherwise. 35 2.5. Random Designs 2.5 Random Designs Until now, we have discussed the limiting behaviors of the non-stochastic design ma- trix case. In practice, since no infinitely precise measurement device exists, there are also measurement errors in the predictors. This is also the situation for autoregression models. It would be interesting to ask the question at what extent and under what assumptions the previous results of the non-stochastic design case still hold for the random design case. Let (Ω,F , {Fn}n∈N, P ) be a filtered stochastic process, that is Fn is an increasing sequence of sub-σ-fields of F and F0 = {∅,Ω}. Let σ(ei) be the σ-field generated by r.v. ei. Random design assumptions 1. (Random design matrix assumption) Cn = 1 n ∑n i=1 xix ∗ i P→ C where C is a positive definite matrix. 2. (Measurability assumption) xi is Fi−1-measurable for all i ∈ N. 3. (Error assumption) (a) The error has a symmetric common distribution w.r.t. the origin. So Eei = 0 and median of ei is 0. (b) ei has a continuous, positive p.d.f. f w.r.t. the Lebesgue measure in a neighborhood of 0. (c) ei possesses a finite second moment σ 2. (d) σ(ei) is independent of Fi−1 for all i ∈ N. Example 2.5.1. Consider the auto-regression model xi = β ∗xi−1 + ei−1, (2.19) 36 2.5. Random Designs where {ei} are assumed to be i.i.d. Let Fn = σ({ei}i∈{0,··· ,n}). Then the measurability assumption and part d) of the error assumption of the random design are satisfied. Theorem 2.5.1. If X and {en}n∈N obey the set of random design assumptions and λn/ √ n → λ0 ≥ 0, then the robust Lasso estimator β̂n defined in (2.4) satisfies √ n(β̂n − β)⇒ argmin(V ) where V (u) = (δ + (1− δ)f(0))u∗Cu+ u∗W + λ0 p∑ j=1 [ujsgn (βj) 1(βj 6= 0) + |uj|1(βj = 0)] and W ∼ N(0, ((1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10)C). As a special case of random design, we now consider the Gaussian random matrix. Corollary 2.5.2. Suppose X is an n× p Gaussian random matrix obeying the ran- dom design matrix and measurability assumptions and {en}n∈N obeys the error as- sumptions. If λn/ √ n→ λ0 ≥ 0, then the robust Lasso estimator β̂n defined in (2.4) satisfies √ n(β̂n − β)⇒ argmin(V ) where V (u) = (δ + (1− δ)f(0))u∗Cu+ u∗W + λ0 p∑ j=1 [ujsgn (βj) 1(βj 6= 0) + |uj|1(βj = 0)] and W ∼ N(0, ((1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10)C). Proof. The corollary follows easily from Theorem 2.5.1 and the fact that the smallest singular value of X, σmin(X)→ 1 P -a.s. by the strong law of large numbers or [5]. Remark 7. Using the same conditioning argument, we can show the asymptotic nor- mality of the Huberized Lasso for the random design case as well. 37 2.6. Numeric Examples 2.6 Numeric Examples Since the Huberized loss is differentiable, piecewise quadratic and the penalty is piecewise linear in β, it follows that β̂ H n (λ) is piecewise linear in λ and hence the whole path of shrinkage can be efficiently computed with the LARS-Lasso algorithm [107]. In contrast, since the convex combined loss is not differentiable at yi−u∗xi = 0, it is not guaranteed that the solution path is piecewise linear in λ. Nonetheless, since the objective function in this case is convex, we can still solve it with an unconstrained convex optimization procedure. Here for a fair comparison, we used CVX, a package for specifying and solving convex programs [62]. The underlying model we assume is as follows: yi = x ∗ iβ + ei, (2.20) where β = (3, 1.5, 0, 0, 2, 0, 0, 0)∗. X is realized from a Gaussian random matrix with zero mean and unit variance. So we have C = I8×8. The errors are generated based on two different mechanisms, with more details given shortly. The intercept term is not considered since it can always be estimated by the mean of y. Therefore, the response y is centered before applying any shrinkage model. The shrinkage tuning parameter λn is chosen to be n 1/3 for all Lasso and robust Lasso models such that they have both parameter estimation and model selection consistency for adaptive weight ŵ = 1/|β̂LSn |γ with γ = 1. Note that the shrinkage sequence chosen here is not universal and optimal in terms of prediction. It is used merely to demonstrate the validity of the derived theory. Practical determination of {λn} is usually by the BIC, cross-validation procedure, etc. The theoretic variances of asymptotic normality can be numerically computed. We set δ = 0.1 for the convex combined loss, and δ = 1 for the Huber loss. All simulations are averaged and reported over 100 simulated date sets, each of which contains n = 1, 000 data points. The following two error distributions are considered: 38 2.6. N u m eric E x am p les Error distribution Model β1 β2 β3 β4 β5 β6 β7 β8 Gaussian Mixture µ = 5, σ = 1 Lasso 13.50 13.50 13.50 13.50 13.50 13.50 13.50 13.50 (13.84) (8.74) (11.43) (13.94) (12.22) (14.12) (11.87) (16.00) RLASSO 7.66 7.66 7.66 7.66 7.66 7.66 7.66 7.66 (8.05) (6.44) (5.71) (5.84) (6.94) (6.71) (4.80) (7.41) HLASSO 5.65 5.65 5.65 5.65 5.65 5.65 5.65 5.65 (7.03) (5.61) (4.99) (5.56) (5.87) (6.39) (4.10) (5.79) adaLASSO 13.5 13.5 0 0 13.5 0 0 0 (14.52) (13.81) (5.33) (8.32) (13.71) (5.50) (9.63) (11.15) adaRLASSO 7.66 7.66 0 0 7.66 0 0 0 (8.02) (7.94) (0.51) (1.28) (9.19) (0.54) (1.75) (1.76) adaHLASSO 5.65 5.65 0 0 5.65 0 0 0 (6.57) (6.57) (0.58) (1.13) (7.95) (0.54) (1.35) (1.94) Student-t ν = 3, σ = 3 Lasso 27 27 27 27 27 27 27 27 (28.77) (28.13) (26.50) (26.87) (30.53) (25.52) (23.58) (23.96) RLASSO 12.87 12.87 12.87 12.87 12.87 12.87 12.87 12.87 (17.27) (17.54) (15.27) (14.64) (20.33) (13.25) (13.15) (11.85) HLASSO 13.45 13.45 13.45 13.45 13.45 13.45 13.45 13.45 (14.92) (15.08) (12.42) (11.06) (16.43) (11.88) (13.08) (10.00) adaLASSO 27 27 0 0 27 0 0 0 (28.46) (27.22) (19.96) (23.74) (25.89) (18.38) (14.90) (15.48) adaRLASSO 12.87 12.87 0 0 12.87 0 0 0 (19.59) (16.19) (4.59) (4.98) (16.25) (3.05) (2.36) (2.61) adaHLASSO 13.45 13.45 0 0 13.45 0 0 0 (16.08) (12.95) (3.31) (2.94) (13.89) (1.89) (1.35) (1.98) Table 2.1: Theoretic and empirical (shown in the parentheses) asymptotic variances of √ n(β̂n − β) for the mixture of Gaussian and Student-t error distributions, respectively. RLASSO is the robust Lasso with the convex combined loss function and HLASSO is the Huberized Lasso. The adaLASSO, adaRLASSO, and adaHLASSO are the corresponding adaptive versions. β1 to β8 are the coefficients of eight predictors. 39 2.6. Numeric Examples 1. Symmetric Gaussian mixture with three components. The errors are simulated from a Gaussian mixture distribution with symmetric two-side outliers, i.e. the error is assumed to have the following p.d.f. f(ei) = 1 4 N(ei;−µ, σ2) + 1 2 N(ei; 0, σ 2) + 1 4 N(ei;µ, σ 2), (2.21) where N(ei;µ, σ 2) denotes the p.d.f. of a normal random variable with mean µ and variance σ2. It is clear that f satisfies the error assumption. The results are reported for µ = 5 and σ = 1. Figure 2.1 shows the histograms of √ n(β̂n − β) for the non-adaptive models and Figure 2.2 shows those of adaptive models. The theoretic and empirical variances of the limiting distribution of √ n(β̂n−β) are shown in Table 2.1. Several conclusions can be drawn here: (a) The variances of √ n(β̂n − β) based on simulations are quite close to the theoretic asymptotic variances (see Figure 2.1). (b) The variances of the scaled convex combined robust Lasso and the Hu- berized Lasso estimators are smaller than that of the Lasso estimator, as expected (see Table 2.1). (c) Although the adaptive Lasso has been proved to be model selection consis- tent, the simulation study shows that the adaptive Lasso performs poorly when the noise variance σ2 is large, at least for a relatively large sample size n = 1, 000. In contrast, the two adaptive robust Lassos show signif- icant performance improvements over the ordinary Lasso (see Figure 2.2 for the zero coefficients). (d) Based on the simulations results, it is observed that the non-adaptive (ro- bust) Lassos do not seem to be model selection consistent even though the irrepresentable condition of [130] is met in our simulation setup. Closer examination reveals that the particular shrinkage sequence we chose is not a model selection consistency one for the non-adaptive cases because 40 2.6. Numeric Examples n−1/6−c 9∞ for any 1 > c ≥ 0 as given by Theorem 1 in [130]. 2. Student-t errors with heavy tails. The setup is the same as the Gaussian mixture case, except that the errors are generated from a Student-t distribution with the degree of freedom ν = 3 and σ = 3. The theoretic and empirical values of the variances of the asymptotic distribution of √ n(β̂n − β) are given in Table 2.1. Based on the histogram results, same observations can be noted as in the Gaussian mixture case. Therefore, due to the space concern, we do not report the figures here. 41 2.6. N u m eric E x am p les Figure 2.1: Histograms of argmin(Vn) = √ n(β̂n − β) for the Gaussian mixture with µ = 5. Green curve is the fitted normal distribution to the estimated values of argmin(Vn) from data over 100 simulations. Red curve is its theoretic asymptotic normal distribution of argmin(V ). Three rows represent the Lasso, convex combined robust Lasso, and Huberized Lasso models in the order of top-down. Columns represent the eight predictors in the order of (3, 1.5, 0, 0, 2, 0, 0, 0). 42 2.6. N u m eric E x am p les Figure 2.2: Histograms of argmin(Vn) = √ n(β̂n−β) for the Gaussian mixture with µ = 5 with adaptation. Green curve is the fitted normal distribution to the estimated values of argmin(Vn) from data over 100 simulations. Red curve is its theoretic asymptotic normal distribution of argmin(V ). Three rows represent the Lasso, convex combined robust Lasso, and Huberized Lasso models in the order of top-down. Columns represent the eight predictors in the order of (3, 1.5, 0, 0, 2, 0, 0, 0). 43 2.7. Conclusion and Discussion 2.7 Conclusion and Discussion In the presence of noise with large variance, the standard Lasso may behave poorly in estimating the true regression coefficients. We propose a flexible, robust version of the Lasso, which combines the advantages of both the `1 and `2 losses. The asymp- totic normality and model selection consistency are established at certain shrinkage rates. The limiting behavior of the Huberized Lasso, another robust Lasso, is also studied. Analysis results derived from the non-stochastic design case are extended to the random design, for which auto-regression models can be suitably handled. The asymptotic analysis framework presented in this chapter provides an appro- priate starting point for future, more general analysis on such robust models, and we hope that the current finite-dimensional asymptotic results can provide certain implication and insight into more challenging settings. For instance, the asymptotic analysis could shed light into non-asymptotic analysis since finite sample size results can be closely related to the asymptotic ones. We have derived a finite sample size `2 estimation error bound for the noiseless case (Proposition 2.2.5) where in fact p is allowed to be greater than n as long as the “incoherent design” condition is valid, and its conclusion agrees with that of the finite-dimensional asymptotic analysis. A future direction is to derive non-asymptotic estimation error bounds in the presence of noise for the situations where p > n and p → ∞. This is challenging, since analyzing the penalized robust losses is much more complicated than the regular Lasso. However, those error bounds can provide great insights into the finite sample size behavior of the robust estimator and could be useful to many research areas such as compressed sensing. As mentioned in the beginning of this chapter, the motivating biomedical appli- cation of this work is brain effective connectivity modeling using fMRI data. Since biomedical research is usually conducted at a group level, means to address within- group, inter-subject variability are required. The proposed robust Lassos can be easily extended to group versions by minimizing the summation of block Euclidean norms 44 2.7. Conclusion and Discussion objective/regularized function [48, 128]. Therefore, sparse features can be learned at the group level and we have already shown some promising preliminary group anal- ysis results regarding brain connectivity in fMRI [35]. The results are presented in Chapter 6. 45 Chapter 3 A Bayesian Lasso via Reversible-Jump MCMC 3.1 Introduction 3.1.1 Sparse Linear Models This chapter considers the same multivariate linear regression model (2.1) in Chap- ter 2. Here, we give further motivations of the present chapter, which aims to obtain more stable estimates in a fully Bayesian paradigm. As we have seen in Chapter 1, there have been many variable/model selection methods proposed in the literature from both frequentist and Bayesian perspectives, with a good overview given in [74, 106]. With such a wealth of methods, it is difficult to argue which model is universally preferable. Among different methods, Bayesian approaches using Markov chain Monte Carlo (MCMC) have recently become popu- lar [40]. For instance, the “spike and slab” priors on the regression coefficients were proposed in [98]. Using the Laplace or Bernoulli-Gaussian mixture prior on the re- gression coefficients and introducing latent variables to identify subsets have been the popular choice. Assuming a hierarchical Bayes Gaussian mixture model with latent variables to identify subsets, a Gibbs sampling approach was presented in [58]. The work in [58] was further explored in [79] by embedding the priors jointly. Several MCMC methods have been compared in [40] for selecting the regression coefficients. In this paper, we are interested in the general category of MCMC-based Bayesian approaches; however, as motivated by the success of Lasso model which is to be dis- 46 3.1. Introduction cussed shortly, we will employ a Bayesian model different from the above mixture models and propose a fully Bayesian Lasso framework with the RJ-MCMC approach (not the regular MCMC approaches as in the above work). The penalized likelihood approach in (1.2) has an alternative Bayesian interpreta- tion. As noted in [115], Lasso estimates can be interpreted as Maximum A Posteri- ori (MAP) estimates with the regression coefficients possessing independent Laplace (a.k.a. double-exponential) priors. More recently, [70] proposed a more general op- timization approach, the Minorize-Maximize (MM) algorithm [81], to transfer the problem of maximizing the posterior function w.r.t. the Laplace prior to sequen- tially maximizing its quadratic surrogate functions. Motivated by this connection between Lasso estimates and the Bayesian interpretation for the Laplace prior, sev- eral Laplace-like priors have been recently proposed for promoting sparsity, e.g. a mixture of delta-mass at 0 and the Laplace prior was studied in [127] and Jeffrey’s non-informative mixing distribution on the prior of β in [52]. The popularity and the good performance of the Lasso model motivates us to employ a Laplace prior for the regression coefficients in our proposed RJ-MCMC based Bayesian approach, a Bayesian Lasso estimator. The observations in [102] actually suggested potential advantages of the Laplace prior over a Gaussian (or a Student-t) prior. 3.1.2 Related Work and Our Contributions It must be emphasized that the non-Bayesian Lasso and Lasso-like approaches have one aspect in common: they are optimization methods with the goal of determin- ing the model parameters that maximize some objective function. Meanwhile, the number of variables set to be zero in these methods critically depends on the tun- ing shrinkage parameter, where its value can generally be selected to minimize the generalized cross-validation errors. In this paper, apart from those Lasso-like opti- mization methods, we propose a new fully Bayesian framework to deal with the Lasso objective function. Such a fully Bayesian approach with the Laplace prior on β, 47 3.1. Introduction referred as a Bayesian Lasso, does not require cross-validation (CV) type methods to determine the optimum shrinkage parameter as in Lasso, since a non-informative prior is given for the shrinkage controlling parameter and its posterior distribution is completely derived from the observed data. By integrating parameters w.r.t. their posterior distributions, the proposed Bayesian Lasso estimator has a different poste- rior distribution from the ordinary Lasso (Laplace prior), and can yield more robust estimates. Very recently, work has been proposed in the direction of Bayesian Lasso [102]. In [102], with a conditional Gaussian prior on β and the non-informative scale-invariant prior on the noise variance being assumed, a Bayesian Lasso model is proposed and a simple Gibbs sampler is implemented. It is shown that the Bayesian Lasso estimates in [102] are strikingly similar to those from the ordinary Lasso. Since this Bayesian Lasso in [102] involves the inversion of the covariance matrix of a block coefficients at each iteration, the computationally complexity prevents its practical application with, say, hundreds of variables. Moreover, similar to the regular Lasso, the Bayesian Lasso in [102] uses only one shrinkage parameter t to both control model size and shrink estimates. Nonetheless, it is arguable whether the two simultaneous effects can be well-handled by a single tuning parameter [94]. To mitigate this non-separability problem, [99] proposed an extended Bayesian Lasso model by assigning a more flexible, covariate-adaptive penalization on top of the Bayesian Lasso in the context of Quantitative Trait Loci (QTL) mapping. Alternatively, introducing different sources of sparsity-promoting priors on both coefficients and their indicator variables have been studied, e.g. in [104], where a normal-Jeffery scaled-mixture prior on coefficients and an independent Bernoulli prior with small success probability on the binary index vector are combined. Motivated by this observation, we introduce two parameters in the proposed Bayesian Lasso model to separately control the model selection and estimation shrinkage issues in the spirit of [94] and [127], and propose a Poisson prior on the model size together with the Laplace prior on β to identify the sparsity 48 3.1. Introduction pattern. Since the proposed joint posterior distribution is highly nonstandard and a standard MCMC is not applicable, we employ a reversible-jump MCMC (RJ-MCMC) to obtain the proposed Bayesian Lasso estimates by simultaneously performing model averaging and parameter estimation. It is worth emphasizing that, though RJ-MCMC algorithms have been developed in the literature before model selection and estimation purposes (e.g. [4] proposed a hierarchical Bayesian model and developed an RJ- MCMC algorithm for joint Bayesian model selection and estimation of noisy sinusoids; similarly [111] proposed an accelerated truncated Poisson process model for Bayesian QTL mapping), these methods are not intended for promoting sparse models whereas our model utilizes sparsity promoting priors in conjunction with the discrete prior on the model size. As we show later, the proposed fully Bayesian Lasso framework provides estima- tion performance improvements when compared with Lasso, the Gibbs sampler-based Bayesian Lasso in [102] and the Binomial-Gaussian model in [59]. When handling the nearly singular case (p ≈ n), the performance improvement from the proposed Bayesian Lasso estimate is even more significant. As a side benefit, we also extend the proposed RJ-MCMC estimation framework to the Binomial-Gaussian model in [59], and the developed BG-MCMC approach yields significant performance improvements over the original non-Bayesian approach in [59]. 3.1.3 Notations We now define some notations to be used throughout the chapter. Let γ be a p- length binary vector where ones denote non-zero coefficients and zeros denote zero coefficients. Equivalently, position of ones in γ can be thought as the active set or support of a linear regression model while position of zeros is called the inactive set. |γ| is used to denote the number of non-zeros in γ, meaning the cardinality of the support of γ. Some special functions and probability density functions are listed in Table 3.1. 49 3.2. A Fully Bayesian Lasso Model Table 3.1: Special functions and probability density functions used in the chapter. Name Functional form Gamma function Γ(a) = ∫∞ 0 ta−1 exp(−t) dt Beta function B(a, b) = Γ(a)Γ(b) Γ(a+b) Gamma distribution Ga(x; a, b) = b a Γ(a) xa−1 exp(−bx) Inverse Gamma distribution IG(x; a, b) = b a Γ(a) x−(a+1) exp(− b x ) Beta distribution Beta(x; a, b) = x a−1(1−x)b−1 B(a,b) The rest of the chapter is organized as follows. In Section 3.2, we first describe the new hierarchical, fully Bayesian Lasso model, and then propose an RJ-MCMC algorithm to simulate this posterior distribution for computing the unbiased minimum variance estimator of the regression coefficient vector. Simulations are carried out in Section 3.4 to evaluate the performance of the proposed approach. Section 3.5 presents the results on a diabetes data set. In Section 3.6, we apply the proposed method to real fMRI data to demonstrate the applicability of the proposed Bayesian Lasso method. 3.2 A Fully Bayesian Lasso Model 3.2.1 Prior Specification The proposed fully Bayesian model has the basic structure of a standard linear regres- sion model in (2.1), with the addition of priors over the parameters to be estimated. Objective hyper-priors are chosen on the sparsity promoting priors. Firstly, to achieve the parsimonious estimation goal, we assign sparsity promoting priors on each of the non-zero coefficients βj’s. Here, independent Laplace priors are assumed for β, i.e. each component βj in the active set γ with |γ| = k follows the distribution pi(βj|τ,γ) = 1 2τ exp ( −|βj| τ ) , (3.1) where τ is a shrinkage tuning parameter for βj with j ∈ γ. Otherwise, βj ≡ 0 for j /∈ γ. 50 3.2. A Fully Bayesian Lasso Model Since in practice we may not have prior knowledge regarding how much shrinkage amount should be put on the coefficients before performing any experiment, it is reasonable to also assign a non-informative prior on the hyper-parameter τ . A non- informative prior for the noise variance σ2 is given based on the same rationale. Together, traditional non-informative priors are put on the higher level as: p(τ, σ2) ∝ (τσ)−1. (3.2) Due to Lindley’s paradox [89], one needs to be very careful to assign improper priors for τ and σ2 when performing model selection/averaging. Since we do not allow the null model with no predictor, τ and σ2 are the common parameters for all possible sub-models. Hence, the underdetermined proportional constants are the same for all sub-models and therefore do not affect the model comparisons based on the posteriors. Further, prior probabilities are also assigned to each possible sub-model. Specifi- cally, a sub-model containing k predictors follows a right-truncated Poisson distribu- tion at p, p(k|λ) = e −λλk Ck! , where C is a normalization constant, and k = 1, ..., p. Within the set of sub-models having k predictors, each sub-model is assumed to be with equal prior probability. To complete the prior specification, the parameter λ is also assumed to follow a non-informative prior as p(λ) ∝ λ−1. Therefore, the posterior distribution of the parameter vector (β, τ, σ2,γ, λ) can be expressed as (up to a multiplicative constant) p(β, τ, σ2,γ, λ|y, X) ∝ e −λλk−1( p k ) k! τ−(k+1)σ−(n+1) exp ( −||β||1 τ − ||y −Xβ|| 2 2 2σ2 ) . (3.3) The derivation detail of (3.3) can be found in the Appendix A.3.1. It is further 51 3.3. Bayesian Computation noted that, to reduce the computational cost, the parameters τ, σ2, and, λ can be analytically integrated out. More specifically, by noting the normalization constants of the respective Gamma and inverse Gamma p.d.f.’s (Table 3.1), we can show that pi(β,γ|y, X) ∝ Γ(k)B(k, p− k + 1) ‖β‖−k1 ‖y −Xβ‖−(n−1)2 . (3.4) Comparing (3.4) with the posterior distribution of the standard Lasso reveals inter- esting observations. (3.4) comprises two parts: ‖β‖−k1 is the prior information and ‖y −Xβ‖−(n−1)2 is the likelihood. These two parts are linked through polynomial terms, representing a different weighting scheme from Lasso. By integrating out the nuisance parameters analytically, a more stable estimator is possible, in addition to the advantage of requiring fewer computations. It is obvious that this joint posterior distribution is in a non-standard form and there is no closed-form analytic expression of E(β|y,X) w.r.t. its posterior distribu- tion. Hence, we must resort to simulation-based approaches to compute the numeric estimates. In this paper, we will develop a Markov chain Monte Carlo (MCMC) type of estimation method. Remark 8. For the standard Lasso approach, there is one shrinkage parameter t in (1.2) to both control model size and shrink estimates, but for some applications, it is desirable to separate those two effects. Here, in the proposed framework, we have two parameters (i.e. τ and λ) to separately control the model selection and estimation shrinkage issues. Roughly speaking, a Poisson prior on the size of active set k controls the size of the expected number of selected predictors and a Laplace prior on β recovers the non-zero coefficients which can best represent the full model conditioned on k. 52 3.3. Bayesian Computation Algorithm 1: The proposed RJ-MCMC based Bayesian Lasso Input: The number of iterations T . Random walk step size . Data: X and y. Output: { θ(t) = (β(t),γ(t)) ∣∣t ∈ {0, · · · , T}}. begin1 Initialization: set θ(0) = (β(0),γ(0)) and t = 1.2 repeat3 if k(t−1) = 1 then4 k(t) ← k(t−1) + U({0, 1})3.5 else if k(t−1) = p then6 k(t) ← k(t−1) − U({0, 1}).7 else8 k(t) ← k(t−1) + U({−1, 0, 1}).9 end10 Sample s ∼ N(0, 2).11 K ← γ(t−1) and Kc ← {1, · · · , p} \K.12 if k(t−1) = k(t) then13 Sample j ∼ U(K).14 Update β (t) j ← β(t−1)j + s with an MH step, details in Section 3.3.2.15 else if k(t) = k(t−1) + 1 then16 Sample j ∼ U(Kc).17 Perform a “birth” move and update β (t−1) j , details in Section 3.3.3.18 else19 Sample j ∼ U(K).20 Perform a “death” move and update β (t−1) j , details in Section 3.3.3.21 end22 t← t+ 1.23 until t = T.24 end25 3.3 Bayesian Computation Since the joint posterior distribution in (3.4) contains both discrete and continuous parameters, a closed-form solution of the unbiased minimum variance estimator is infeasible: the posterior expectation of coefficients E(β|X,y) and the posterior prob- ability of the inclusion of coefficients E(γ|X,y). Moreover, standard MCMC is not applicable in this case since the model dimension is not fixed. To address this diffi- culty, we propose a hybridized MCMC sampler to simultaneously perform model av- 53 3.3. Bayesian Computation Figure 3.1: An illustration of model jumping from γ → γ ′ with |γ| = 5 and |γ ′| = 6. New predictors (in red) are created from current model. A position with 1 indicates a non-zero coefficient, 0 denotes current model excludes this coefficient. eraging and parameter estimation. Our proposed algorithm falls into the RJ-MCMC umbrella [63]. RJ-MCMC is a powerful prototype that creates MCMC algorithms for variable dimensional models and may be better than separate within-model MCMC runs if we aim at making joint inference about the models and their parameters. Moreover, running separate MCMC for each model is computationally prohibitive for large scale problems. The proposed algorithm is summarized in Algorithm 1, with more details given in the following paragraphs. 3.3.1 Design of Model Transition It is clear that the distribution of β depends on the model dimensionality. For ex- ample, deleting predictors will force their corresponding index set to zeros. Here, we propose three types of model moves as following: 1. γ → γ; 2. γ → γ ′; 3. γ ′ → γ. To smoothly move between models and allow fast mixing, we design local jumps. At each sampling step, the current model is only allowed to stay in the same dimension or move to its neighboring models. As illustrated in Figure 3.1, the proposed model has either the same dimension as the previous one, or one predictor added or deleted from the current model. Moreover, we assign each possibility with equal probabilities, 54 3.3. Bayesian Computation as p(γ → γ) = 1 3 , p(γ → γ ′) = 1 3(p− k) , p(γ ′ → γ) = 1 3(k + 1) , for k = 2, · · · , p− 2 and |γ| = k and |γ ′| = k+1. The boundary models need slightly different probabilities. For |γ| = 1, we do not allow the null model with no predictor at all. Hence for a model with just one predictor, it can only stay in one-dimensional or move to two-dimensional, each with probability 1 2 . Namely, for |γ| = 1 and |γ ′| = 2 p(γ → γ) = 1 2 , p(γ → γ ′) = 1 2(p− 1) . Similarly, for |γ| = p and |γ ′| = p− 1, we have p(γ → γ) = 1 2 , p(γ → γ ′) = 1 2p . 3.3.2 A Usual Metropolis-Hastings Update for Unchanged Model Dimension For the models with unchanged dimension, the standard Metropolis-Hastings (MH) algorithm is used to update β [67]. Specifically at iteration t, a predictor at position j ∈ γ to be updated is randomly selected from current non-zero coefficients. Then, a proposal distribution q(θ(t),θ′) is chosen to update this predictor, where θ(t) is current parameter estimate and θ′ is the proposed parameter. Here, a Gaussian random walk (RW) is used as our proposal, N(0, 2), with some fixed small step size . Set β′ = β+uej where ej is the jth standard Euclidean basis. Then the acceptance 55 3.3. Bayesian Computation probability becomes min {(‖β′‖1 ‖β‖1 )−k × (‖y −Xβ′‖2 ‖y −Xβ‖2 )−(n−1) , 1 } . (3.5) 3.3.3 A Birth-and-Death Strategy for Changed Model Dimension Since there is no concept of metric structure and compactness as in the Euclidean space for trans-dimensional jumps inΘ, designing an optimal or even a valid proposal is not an easy task. The standard MCMC optimal scaling proposal has no analogue for reversible jump moves [15]. For the model jumping moves, there are two commonly used proposals: i) birth-and-death and ii) split-and-merge. The birth-and-death is a simple form of model transformation: In the birth step, a new predictor is added to the current model, by generating parameters of a new predictor from a prior distribution; in the death step, a predictor is removed from current model, and the reversibility constraint must be satisfied according to the detailed balance equation. [63] showed that if there exists a σ-finite symmetric measure µ, with respect to which pi(dθ)q(θ, dθ′) is absolutely continuous with pi is our target posterior distribution in (3.4), then the detailed balance condition holds for all Borel subsets B,B′ ⊂ B(Θ), ∫ B×B′ α(θ,θ′)ρ(θ,θ′)µ(dθ, dθ′) = ∫ B′×B α(θ′,θ)ρ(θ′,θ)µ(dθ′, dθ), (3.6) if the acceptance ratio α(θ,θ′) is chosen to be α(θ,θ′) = ρ(θ′,θ) ρ(θ,θ′) ∧ 1, (3.7) where the extended µ-integrable function ρ is the Radon-Nikodym derivative of pi× q with respect to µ. It is easy to see that the unchanged model dimension update γ → γ is just a special case by taking µ as the Lebesgue measure on B(Rk)⊗B(Rk), the product Borel σ-algebra on Rk. 56 3.3. Bayesian Computation For a birth move, a new predictor is created by random generation from the inactive set. The proposed predictors are accepted with the probability given by the generalized Metropolis-Hasting ratio. More specifically, a coefficient is randomly generated outside the current support set (i.e. generating a j ∈ γc and setting the value of this coefficient with a zero mean Gaussian realization u). Since the model dimension is augmented by generating an additional variable u, there is a Jacobian term for the acceptance probability of the birth move which is 1 in this case. By putting the posterior ratio computed from (3.4) and model transition proba- bilities together into (3.7), the acceptance probability is given by min { k2 p− k × ‖β′‖−(k+1)1 ‖y −Xβ′‖−(n−1)2 ‖β‖−k1 ‖y −Xβ‖−(n−1)2 × p(γ ′ → γ) p(γ → γ ′) ×N(u; 0, 2)−1, 1 } (3.8) where N(u;µ, 2) means the Gaussian density N(µ, 2) evaluated at u. Similarly, the death move is simply the reverse of the birth move. The acceptance probability for γ → γ ′ with |γ| = k and |γ ′| = k − 1 is given by min { p− (k − 1) (k − 1)2 × ‖β′‖−(k−1)1 ‖y −Xβ′‖−(n−1)2 ‖β‖−k1 ‖y −Xβ‖−(n−1)2 × p(γ ′ → γ) p(γ → γ ′) ×N(u; 0, 2), 1 } . (3.9) (3.8) and (3.9) ensure the reversibility of the constructed Markov chain by (3.6). Remark 9. If one is also interested in the nuisance parameters τ , σ2 and λ, it is easy to extend the current algorithm to include embedded Gibbs samplers. Since the full conditional distributions of τ , σ2, and λ can be given in closed forms, the Gibbs sampler [61] is used to simulate these parameters from their posterior distributions, τ |· ∼ IG k,∑ j∈γ |βj| (3.10) σ2|· ∼ IG ( n− 1 2 , ‖y −Xβ‖22 2 ) (3.11) λ|· ∼ Ga(k, 1). (3.12) 57 3.4. Simulations Here |· means the conditional distribution given everything else, including data and other parameters. 3.4 Simulations 3.4.1 Setup Data sets of 100 data points are simulated, with a series of different model dimensions. The value of p is set to be 15, 45, and 90. Each setting has a specific sparse structure of the coefficients, and we denote a particular setting by a format such as 30/90, meaning that 30 out of the total 90 predictor coefficients are non-zeros. For instance, for the 90 dimensional case, we set the coefficients to be (3, · · · , 3︸ ︷︷ ︸ 10 times , 0, · · · , 0︸ ︷︷ ︸ 20 times , 1.5, · · · , 1.5︸ ︷︷ ︸ 10 times , 0, · · · , 0︸ ︷︷ ︸ 20 times , 2, · · · , 2︸ ︷︷ ︸ 10 times , 0, · · · , 0︸ ︷︷ ︸ 20 times ). Although independent sparsity promoting priors are used (3.1), correlations between predictors are also introduced and compared with uncorrelated data and their influ- ence on the performance of various algorithms is explored. We set the correlation level to be 0.5 in our simulations. Several standard linear model selection approaches are compared, including the Lasso [115] (and its variant Gauss-Lasso for extending Lasso to accommodate both model selection and regression objectives, where Lasso is used as model selection followed by a least squares estimate based on the selected model), Lar [45] (and its variant Gauss-Lar, where Lar is used as model selection followed by a least squares estimate based on the selected model), a Gibbs sampler based Bayesian Lasso [102] and the Binomial-Gaussian (BG) model [59]. To examine the effects on the overall performance of the proposed Poisson-Laplace model and the proposed MCMC estimation approach, we also extend the BG model in [59] to the fully Bayesian framework, and develop a corresponding MCMC algorithm, referred as BG-MCMC, with details given in Appendix A.3.2. We shall see that the proposed 58 3.4. Simulations Table 3.2: RMSEs averaged over 100 simulations for n = 100. The number in the bracket is the standard deviations of the estimated RMSEs. Methods under com- parison are the Lasso [115], Gauss-Lasso, Lar [45], Gauss-Lar, Gibbs sampler based Bayesian Lasso [102], Binomial-Gaussian (BG) [59], proposed BG-MCMC and pro- posed Bayesian Lasso (BLasso) with RJ-MCMC algorithm. No correlation Correlation=0.5 3/15 15/45 30/90 3/15 15/45 30/90 0.728 1.023 7.968 0.561 0.859 9.895 Lasso [115] (0.146) (0.158) (2.441) (0.116) (0.163) (2.360) Gauss-Lasso 0.153 0.559 5.751 0.280 0.765 9.848 (0.070) (0.125) (3.059) (0.137) (0.145) (5.073) 0.728 1.021 6.250 0.560 0.868 8.251 Lar [45] (0.146) (0.159) (2.349) (0.116) (0.172) (2.668) Gauss-Lar 0.153 0.561 3.602 0.280 0.815 6.732 (0.070) (0.123) (2.700) (0.137) (0.155) (3.503) 0.376 0.832 1.837 0.499 1.126 2.422 Gibbs sampler [102] (0.087) (0.121) (0.294) (0.112) (0.170) (0.375) 0.165 0.429 10.362 0.199 0.604 11.537 BG [59] (0.077) (0.103) (3.254) (0.103) (0.130) (4.437) BG-MCMC 0.180 0.435 0.704 0.220 0.577 0.909 (0.072) (0.106) (0.129) (0.107) (0.147) (0.149) Proposed BLasso 0.157 0.417 0.708 0.199 0.577 1.497 (0.068) (0.080) (0.235) (0.092) (0.111) (0.959) BG-MCMC method provides significant performance improvements over the original non-Bayesian method in [59]. The shrinkage tuning parameter t for the Lasso and Lar is determined by 10-fold CV with the minimal prediction errors. The proposed BG-MCMC and the Bayesian Lasso estimators are initialized at the LS estimate of the full model and run for 100,000 iterations with the first half runs being discarded as warm-up. The step size of RW is 0.05. The performances of the coefficient esti- mates are measured by the Root Mean Squared Errors (RMSEs) and all results are averaged over 100 simulations. 3.4.2 Empirical Performance Comparisons The RMSE performances of the eight models are summarized in Table 3.2. Several observations can be summarized as follows: First, performances of the Lasso and Lar are similar to each other; this is in 59 3.4. Simulations particular pronounced in lower dimension cases. This is because the Lasso (imple- mented in a modified Lar algorithm [45]) rarely drops variables from the active set and hence is very similar to the Lar. Further, compared with the Lasso and Lar, the proposed Bayesian Lasso with RJ-MCMC algorithm consistently yields much smaller RMSE and smaller estimation variability. This observation is even more significant for the p ≈ n cases where the MLE can be highly unstable. The reduced estimation variability is likely due to the fact that the parameters are estimated based on aver- aged models, rather than conditioning on a single best plausible model given by the penalized MLE principle. Moreover, since the Lasso uses one tuning parameter to simultaneously select variables and shrink estimates, the estimation may be shrunk along with the decreased model size. However, for our proposed approach, since two parameters (τ and λ) incorporate together to control these two effects, it is more flexible and likely to obtain an unbiased estimate. To support this claim, we reported the RMSE results of Gauss-Lasso and Gauss-Lar in Table 3.2 and also reported the estimated sparsity patterns in Lasso and Lar in Table 3.3. These two tables together show that the major source of RMSEs of Lasso and Lar comes from the errors made in the model selection stage. The proposed RJ-MCMC based Bayesian Lasso also consistently yields better es- timation accuracy than the Gibbs sampler-based Bayesian Lasso. The Gibbs sampler method yields the MSEs between the proposed RJ-MCMC method and Lasso/Lar. The BG method in [59] has comparable, slightly worse performance over the pro- posed Bayesian Lasso in lower dimension regimes. However, when the model size increases comparable to the data size (e.g. p = 90 and n = 100), the performance of the BG method substantially degrades. The major advantage of the BG model is that the marginal likelihood of the model can be given in closed-form, conditioned on the known active set for a model. However, since comparing marginal likelihood of all possible models requires the enumeration of the 2p possible models, an exhaustive search is not computationally feasible so a common means to approximate the exact 60 3.4. Simulations solution is to adopt a stepwise searching strategy. As used in [59], forward selection is used to traverse between models, however false predictors introduced in earlier stages of the algorithm cannot be eliminated at a later stage. Degraded performance of BG is especially pronounced when predictors are highly correlated or the sample size is not large enough. Hence, for medium/large scale problems with no special structure (e.g. orthogonality among predictors), the BG method is not a good choice for model selec- tion in practice. In contrast, the stochastic search algorithm proposed in this paper successfully avoids being trapped by sub-optima with the price of increased computa- tional cost. Empirically, we observe that the proposed fully Bayesian algorithms can accurately estimate the model and associated parameters with a reasonable sampling size. By extending the proposed fully Bayesian framework to the BG model [59], we further confirm the strength of the MCMC approach. We derive an MCMC algorithm for the BG model, as in Appendix A.3.2. Many parameters can be integrated out because of the Gaussianality. The main interesting quantity is the model index pa- rameter, the posterior of which can be viewed as the posterior probability of including the corresponding coefficient. It is seen from Table 3.2 that the derived BG-MCMC achieves similar performances as the proposed RJ-MCMC based Bayesian Lasso. We note empirically that there is not much room for improvement for the two fully Bayesian approaches, the proposed BLasso and BG-MCMC. With the assumption that we are given an oracle revealing the location of the true non-zero coefficients, it is easy to see that the optimal least squares estimator has an RMSE converging to √ k n σ for large n. In our setup, with σ2 = 1, the fundamental information-theoretic limits of RMSEs in Table 3.2 are 0.173, 0.387, 0.548, respectively. In our simulations, we empirically observe that the performances of the proposed RJ-MCMC based Bayesian Lasso and BG-MCMC approximate the lower estimation error bound. Therefore, we suggest that there is a substantial advantage over greedy search and optimization based methods by using the proposed fully Bayesian framework coupled with the 61 3.4. Simulations stochastic search. We also examine the sparsity patterns recovered by different algorithms under consideration. For BG, Lasso and Lar, sparsity patterns are characterized by the locations of non-zero coefficients. For BG-MCMC and the proposed Bayesian Lasso with RJ-MCMC, a non-zero coefficient is declared when its posterior probability passes the threshold 0.5. For the Bayesian Lasso based on the Gibbs sampler, since the distribution of β is assumed to have a conditional zero-mean normal distribution, we calculate its two-side tail probabilities that exceed the posterior estimate ∣∣∣β̂j∣∣∣ and then compare the tail probabilities with a significance level 0.1. To compare the algorithms, we evaluate the performances of support recovery in terms of their F -scores which combine the precision and recall (true positive rate) measures. More specifically, the precision P is the fraction of detected true positive among all identified positives whereas the recall R is the identified true positive ratio to the total number of true positives. The F -score is defined as the harmonic mean of the precision and recall, i.e. F = 2PR P +R . (3.13) The closer F -score of an algorithm is to 1, the better performance it has. It is clear from Table 3.3 that the proposed BLasso and BG-MCMC yield the best and stablest performances, with the proposed BLasso slightly outperms BG- MCMC. Lasso, Lar, Gibbs sampler, BG have comparable support detecting capability in lower dimensional problems; however, their performances significantly degrade as the problem size gets larger. 3.4.3 Convergence Analysis We now prove that the proposed RJ-MCMC framework in Algorithm 1 converges to the posterior distribution of (γ,β) given in (3.4). The proof is based on the standard argument, e.g. see [97]. Let M = {( γ(i),β(i) )} i∈N be the Markov chain constructed by Algorithm 1 such that the detailed balance condition (3.6) implies pi(γ,β|y) is an 62 3.4. Simulations Table 3.3: F -scores of estimated sparsity patterns averaged with standard deviations in brackets over 100 simulations for n = 100. Methods under comparison are the Lasso [115], Lar [45], Gibbs sampler based Bayesian Lasso [102], Binomial-Gaussian (BG) [59], the proposed BG-MCMC and the proposed Bayesian Lasso (BLasso) with RJ-MCMC algorithm. No correlation Correlation=0.5 3/15 15/45 30/90 3/15 15/45 30/90 0.996 0.888 0.713 0.927 0.802 0.599 Lasso [115] (0.025) (0.062) (0.179) (0.099) (0.049) (0.133) 0.996 0.887 0.798 0.927 0.763 0.673 Lar [45] (0.025) (0.063) (0.119) (0.099) (0.046) (0.099) 0.964 0.612 0.528 0.929 0.572 0.517 Gibbs sampler [102] (0.070) (0.066) (0.037) (0.109) (0.062) (0.041) 0.850 0.505 0.230 0.791 0.393 0.141 BG [59] (0.097) (0.066) (0.017) (0.048) (0.086) (0.045) BG-MCMC 0.993 0.993 0.991 0.994 0.998 0.999 (0.031) (0.014) (0.012) (0.028) (0.008) (0.004) Proposed BLasso 1.000 1.000 1.000 1.000 1.000 0.979 (0.000) (0.003) (0.004) (0.000) (0.000) (0.070) invariant distribution for M . With this target distribution, it suffices to show that M is ergodic with respect to pi(γ,β|y) [116]. This is equivalent to show that M is pi(γ,β|y)-irreducible and aperiodic. Aperiodicity is obvious and the only part we need to argue is the pi(γ,β|y)-irreducibility. The idea of showing the irreducibility of M is to find a particular path that monotonically shrinks the model size to one with certain positive probability. We can assume without loss of generality that the only predictor in the destination model is the first one. Let K(γ,β;γ ′, dβ′) be the transition kernel of M : P (γ ′,β′ ∈ B|γ,β) = ∫ B K(γ,β;γ ′, dβ′) (3.14) for all B ∈ B(R|γ ′|). In order to prove that M is p(γ,β|y)-irreducible, it suffices to establish a µ-irreducibility of M for some σ-finite measure µ defined on the mea- surable space ( [p]× Rp, 2[p] ⊗ B(Rp)), where 2[p] denotes the power set of [p] and [p] = {1, · · · , p}, see [116]. Taking µ(γ,β) = p−1I{1}(|γ|)N(0, 1), we want to show that, for any γ ∈ 2[p] and β ∈ R|γ |, there is a non-vanishing probability for the state 63 3.4. Simulations (γ,β) to commute ({1, 0, 0, · · · }, B) for every µ({1, 0, 0, · · · } × B) > 0. Considering the one-step transition kernel (3.14) for the event that a death occurs, we have by construction K(γ,β;γ ′, dβ′) = 1 3k min{A, 1}ISγ ,β (β ′)dβ′, (3.15) where A is the first term in the death probability ratio in (3.9) and Sγ ,β = { β′ ∈ Rk−1 |∃j s.t. γ = {j} ∪ supp (β′)} . Note that ‖β‖1 = ‖β′‖1 + |u| ≥ ‖β′‖1. Let Ck = ‖xj‖2 2 u2+2〈y−Xβ,xj〉u ‖y−Xβ‖2 2 < ∞ and αk = (1 + Ck) −1. The Cauchy-Schwartz inequality implies that αk > 0. We have A ≥ C ′kαk > 0, where C ′k = p−(k−1)(k−1)2 p(γ ′→γ) p(γ→γ ′)N(u; 0, 2) > 0. So we deduce that K(γ,β;γ ′, dβ′) ≥ C ′kαkISγ ,β (β ′)dβ′ 3p . Iterating this process k − 1 times, we can obtain P ({1, 0, 0, · · · } ×B|γ,β) ≥ ∫ B k∏ i=2 K(γi,βi;γi−1, dβi−1) dβ1 ≥ (3p)−(k−1)µ({1, 0, 0, · · · } ×B) k∏ i=2 (C ′iαi) > 0. The last step may be complemented by a standard MH-step (3.5) to update the coeffi- cient with the same dimension. This shows that we can reach the state ({1, 0, 0, · · · }, B) with a strictly positive probability. In summary, the above facts lead to the following convergence theorem. Theorem 3.4.1. Let ( γ(i),β(i) ) be the Markov chain with transition kernel given by the proposed RJ-MCMC algorithm in Algorithm 1. This Markov chain converges to the posterior probability distribution pi(γ,β|y) in (3.4), regardless of the initialization 64 3.5. A Diabetes Data Example of the algorithm, i.e. ∥∥pi(i)(γ,β)− pi(γ,β|y)∥∥ TV → 0, (3.16) where pi(i)(γ,β) means the empirical distribution of ( γ(i),β(i) ) and ‖·‖TV means the total variation norm on bounded signed measures, ‖pi‖TV = sup B∈B pi(B)− inf B∈B pi(B). (3.17) Remark 10. In addition to the birth-and-death proposal used in Algorithm 1, there is another main proposal studied in the literature, namely the split-and-merge strategy. It is worth noting that the convergence of the split-and-merge trans-dimensional strat- egy can also be established in a similar way. Theoretically, both proposals guarantee that the correspondingly designed algorithms converge to the right target distribu- tion. The two proposals may result in different empirical convergence rates which are problem-dependent. As shown earlier, the adopted birth-and-death proposal yields satisfactory estimation performance in our simulations. 3.5 A Diabetes Data Example This is a benchmark data set used in [45]. It contains n = 442 measurements from diabetes patients. Each measurement has ten baseline predictors: age, sex, body mass index (BMI), average blood pressure (BP), and six blood serum measurements (S1-S6). The response variable is a quantity that measures progression of the diabetes one year after baseline. The response is centered and the predictors are normalized to have zero means and unit variances, before applying any model selection methods. We set the random walk step size of the proposed RJ-MCMC to 7 in order to control the acceptance ratio of proposed models to be around 30%. In our experiment, the acceptance ratio is 30.54%. We empirically observe that a smaller step size would 65 3.6. A Real fMRI Application Table 3.4: Estimated coefficients for the Lasso [115], Lar [45], the Bayesian Lasso based on the Gibbs sampler (GS) [102], BG [59], BG-MCMC, and proposed Bayesian Lasso for diabetes data. Predictor Lasso Lar GS BG BG-MCMC Proposed BLasso Age -0.081 -0.325 -0.337 0 0 0 Sex -10.920 -11.228 -11.043 -10.646 -10.923 -10.371 BMI 25.013 24.854 24.877 24.905 25.197 24.861 BP 15.074 15.303 15.183 15.380 15.633 14.655 S1 -15.803 -29.293 -20.997 -35.624 -29.872 -7.022 S2 5.196 15.970 9.551 25.314 20.377 0 S3 -4.498 1.232 -2.550 0 -2.778 -7.785 S4 5.839 7.434 6.174 0 0 3.168 S5 27.645 32.648 29.568 37.798 35.675 24.767 S6 3.101 3.174 3.202 0 0 2.227 cause the algorithm to explore the model space more slowly; while a larger one would have a higher rejection rate. It is clear from Table 3.4 that models under comparison provide results with certain similarities, except for the predictors S1, S2, and perhaps S3. S1 selected by the proposed RJ-MCMC based Bayesian Lasso has a negative coefficient with smaller magnitude than others. For S2, the estimated coefficients are less consistent across different models. For instance, Lasso and Gibbs sampler have positive coefficients with smaller magnitudes than those of the Lar and BG. For the proposed Bayesian Lasso, this predictor is essentially interpreted as insignificant. Therefore, it is not clear whether or not S1-S3 covariates should be selected from a solely computational point of view, and further medical or physical interpretation is needed to justify the choice of different models in a real-world problem. 3.6 A Real fMRI Application 3.6.1 Application Description In this section we demonstrate an application to real fMRI data derived from subjects with Parkinson’s Disease (PD). The problem of interest here is to employ the sparse 66 3.6. A Real fMRI Application linear regression modeling for learning brain functional connectivity using fMRI data. The fMRI data we have are from ten normal people and eight PD patients. During the fMRI experiment, subjects continually squeezed a bulb in their right hand to control an inflatable ring so that the ring moved through an undulating tunnel without touching the sides. A trial of the task was five minutes. Normal subjects performed only one trial; the PD subjects performed the same task twice, once before medication, the other after the medication. fMRI data were collected with a Philips Achieva 3.0 T scanner with a TR interval of 2s. Hence, we collected 150 data points for each subject. After motion correction, the fMRI time courses of the voxels within each ROI were averaged to represent the summary activity of each ROI. The averaged time courses were then linearly detrended and normalized to unit variance. In this study, based on previous neuroscience knowledge, eighteen brain regions were selected as the ROIs based on each person’s individual anatomy. The model that we assume here is a linear regression one which incorporates both spatial and temporal effects of brain connectivities. To do so, we combine the structural equation modeling (SEM) [92] and multivariate autoregressive model (mAR) [65]. Specially, let y(t) = (y1(t), · · · , yp(t))T be a p × 1 dimensional vector, which contains the intensity measurements for the p brain ROIs at time t, for t = 1, · · · , T . For mAR component of the unified model, we consider only up to an mth order process. Hence, by combining the SEM part, the joint SEM and mAR model can be expressed as: y(t) = Ay(t)︸ ︷︷ ︸ SEM + m∑ j=1 Φ(j)y(t− j)︸ ︷︷ ︸ mAR + e(t)︸︷︷︸ noise (3.18) where A and Φ(j) are p× p coefficient matrices to be estimated, for j = 1, · · · ,m; e is a noise vector which is assumed to be Gaussian with zero mean and constant variance σ2. A represents the spatial connection strengths between ROIs, while Φ(j)′s are the time lag effect strengths. We are aiming at applying the proposed 67 3.6. A Real fMRI Application (a) the proposed Bayesian Lasso vs Lar (b) the proposed Bayesian Lasso vs Lasso (c) Lar vs Lasso Figure 3.2: The correlation between the estimated coefficients when using different algorithms. (a) the proposed Bayesian Lasso vs Lar; (b) the proposed Bayesian Lasso vs Lasso; (c) Lar vs Lasso. fully Bayesian method to simultaneously select the model order and estimate the functional connectivities between brain regions represented by the coefficients. 3.6.2 Results We primarily want to check the consistency of the estimates from the aforementioned algorithms. First, we plot and examine the correlations between the estimated co- efficients of these algorithms. The result is shown in Figure 3.2. We can see that the proposed Bayesian Lasso, Lar, and Lasso estimates are highly correlated. In par- ticular, the Lar and Lasso estimates have a higher similarity (with the correlation coefficient being 0.967) than being compared with the proposed Bayesian Lasso (with the correlation coefficient being 0.873 and 0.870 for the Lar and Lasso estimates, respectively). 68 3.7. Discussion and Conclusion Table 3.5: Correlations between the coefficient estimates on two fMRI data subsets when using MLE, Lar [45], Lasso [115] and the proposed BLasso. Method MLE Lar Lasso Proposed BLasso Correlation 0.293 0.599 0.587 0.702 We further investigate estimation stability. We take one subject’s fMRI time-series data and split the 150 time data points into two subsets of size 100, with the middle 50 data points overlapped. We then learn two models from these two subsets and examine the correlation between the estimated coefficients from subsets. The MLE approach for the full (non-sparse) model is also included for comparison. It is noted that the proposed Bayesian Lasso reveals greater estimation stability between models derived from subsets of the fMRI data compared to the other three methods, since it yields the highest similarity between the model coefficients derived from the two subsets (represented by a correlation coefficient being 0.702, see Table 3.5). The Lar and Lasso estimates yield a lower correlation (with the correlation coefficient being 0.599 and 0.587 respectively). The MLE approach provides the lowest consistency (with the correlation coefficient being 0.293), which is not a surprising fact since the variability of the MLE estimate is usually larger than the estimates via sparse regression. Moreover, limiting our attention only to the predictor coefficients which are estimated as non-zero from both subsets, we note that the correlations between the estimates derived from two subsets are even higher for the proposed Bayesian Lasso, Lar, and Lasso approaches. 3.7 Discussion and Conclusion In this chapter, we proposed a hierarchical, fully Bayesian version of the Lasso model for inferring sparse linear regression from high-dimensional data sets. Since the joint posterior distribution of the parameters involves both discrete and continuous pa- rameters, we developed a reversible jump MCMC algorithm to compute the unbiased minimum variance estimates. Simulations demonstrated that the proposed Bayesian 69 3.7. Discussion and Conclusion Lasso estimate yields lower estimation errors when compared with popular Lasso type estimates and a Gibbs sampler based Bayesian estimate. One intuitive expla- nation of this observation is that model averaging by the fully Bayesian approach provides better stability than selecting only a single best model. The simulations further demonsrated that the proposed Bayesian Lasso is robust to correlated pre- dictors, even though the hypothesis of independent priors for predictors is assumed in the model design (3.1). We proved the convergence of the proposed RJ-MCMC algorithm for the Bayesian Lasso. Further, we extended the proposed fully Bayesian framework to the Binomial-Gaussian model, and simulations showed that the pro- posed stochastic search could substantially improve the performance of the original BG model-based estimate in [59]. One important direction for future work is to improve the sampling strategy. Cur- rently, we use the Gaussian RW proposal with a fixed variance parameter. However, as shown in [66], the adaptive RJ-MCMC sampler usually facilitates mixing speed, and thus a data-driven adaptive sampler is of particular interest. We also observe that different step sizes can lead to different models and affect the empirical convergence. The proposed RJ-MCMC Bayesian Lasso approach also has limitations. One is its higher computational cost compared with Lasso-type estimates, a limitation common to MCMC-based Bayesian approaches. This limitation prevents using the proposed method from online high-dimensional estimation problems. However for offline/batch estimation problems (e.g. fMRI modeling), the proposed method can usually provide practical accurate estimates with affordable computational complexity. Table 3.6 reports the required CPU times by different methods in the case of p = 15 and uncor- related design. We observed similar empirical complexity results for other settings in our simulations. Basically, among the discussed MCMC-based Bayesian approaches, we note that the Gibbs sampler based Bayesian Lasso requires the highest computa- tional cost, followed by BG-MCMC, while the RJ-MCMC Bayesian Lasso requires the least computational time. In summary, the computational costs of different estimates 70 3.7. Discussion and Conclusion Table 3.6: The required CPU time for Lar [45], Lasso [115], the Bayesian Lasso based on the Gibbs sampler (GS) [102], BG [59], BG-MCMC, and proposed Bayesian Lasso, in the case of p = 15 and uncorrelated design. CPU times are normalized w.r.t. the Lar running time. The last three simulation based methods run 20,000 iterations. Method Lar Lasso BG BG-MCMC RJ-MCMC GS CPU time 1.000 1.296 25.250 346.065 98.749 2848.318 can be ordered as: Lar ≺ Lasso ≺ BG ≺ RJ-MCMC Bayesian Lasso ≺ BG-MCMC ≺ Gibbs sampler based Bayesian Lasso. 71 Chapter 4 Shrinkage-To-Tapering Estimation of Large Covariance Matrices 4.1 Introduction The main goal of this chapter is to consider the estimation problem of high-dimensional covariance matrix from n iid smaples following a zero-mean p-dimensional multivari- ate Gaussian distribution N(0,Σ)4. Importance and motivation of precise estimation of covariance matrices have been discussed in Chapter 1 of this thesis; therefore, we do not repeat them and jump directly to our results. Before proceeding, we remind the audience that the standard and most natural estimator of Σ is the unstructured sample covariance matrix Ŝ = n−1 n∑ i=1 xix T i , where xi means the ith p-dimensional observation sample. This is defined in (1.3) in Chapter 15. Recall that for the classical case where p is fixed and n → ∞, Ŝ is a consistent estimator of Σ. Unfortunately, the covariance estimation problem becomes fundamentally different and more challenging for high-dimensional settings with a small number of samples where p n meaning the concentration p/n → ∞ (i.e. large-p-small-n). From the eigen-structure perspective, random matrix theory 4Without loss of generality (w.l.o.g.), we assume that the diagonal entries of Σ are all normalized to one. 5Note that we have temporally changed the notation Σ?n for sample covariance matrix to Ŝ for simplicity in the current chapter. 72 4.2. Comparison Between Tapering and Shrinkage Estimators predicts that the spectrum of Ŝ is wider than the spectrum of Σ if p/n9 0 and n, p→ ∞ [6]. For example, the Marchenko-Pastur theorem states that the eigenvalues of Ŝ have a deterministic semicircular limiting distribution supported on [(1−√y)2, (1 + √ y)2] where y = limn→∞ p/n > 0, while the spectrum of Σ is the Dirac mass at 1. This chapter considers the high-dimensional settings and focuses on the corresponding problem of estimating large covariance matrices. The rest of the chapter is structured as follows. In Section 4.2, we first introduce the tapering estimator and its minimax risk bounds under Frobenius and spectral norms; we the derive the risk bounds under the same norms for the MMSE shrinkage oracle estimator proposed in [36]. Inconsistency of the shrinkage estimator will be shown by a set of examples. In Section 4.3, we propose a shrinkage-to-tapering oracle (STO) estimator and derive a closed-form expression of the optimal shrinkage weight under MMSE. An approximating algorithm of the STO estimator is further proposed for practical implementation. Section 4.4 compares the numeric performances of the proposed STO estimators, the tapering estimator and other types of shrinkage estimators. The chapter is concluded in Section 4.5. 4.2 Comparison Between Tapering and Shrinkage Estimators The main contribution of this section is to provide a detailed analysis on risk bounds of tapering and shrinkage estimators for large covariance matrices estimation. Before we formally present our analysis, it is necessary to recall some definitions and results from tapering and shrinkage estimators of covariance matrices. 4.2.1 Tapering Estimator We consider a class of tapering estimators. Let S be the set of p×p symmetric matrix and A ◦B be the Schur product of two matrices A and B: A ◦B = (aijbij). 73 4.2. Comparison Between Tapering and Shrinkage Estimators Definition 4.2.1. A covariance matrix taper (CMT) A is an element in S such that∑p j=1 λj(A ◦B) ≤ ∑p j=1 λj(B) for all B ∈ S. In other words, Schur multiplication by any CMT decreases the averaged eigenvalue. Let W be a CMT, a tapering estimator of the covariance matrix is defined as Σ̂taper = W ◦ Ŝ. (4.1) For some C,C0 > 0 and α > 0, we consider the following class of covariance matrices G(α,C,C0) = Σ : maxj ∑|i−j|>k |σij| ≤ Ck−α,∀k, and λmax(Σ) ≤ C0 , (4.2) where α is a smoothing parameter specifying the rate of decay of σij from the main diagonal. We state that the matrices in G(α,C,C0) diagonally dominant. Note that our definition is different from the usual one in the literature and we use this term as a measure of sparsity of covariance matrices when a natural ordering in variables exists, e.g. in time-series models. The following remarkable theorem, proved by Cai, Zhang, and Zhou [20], shows that a covariance tapering estimator based on data generated from i.i.d N(0,Σ) with Σ ∈ G(α,C,C0) is minimax. Theorem 4.2.1. (Cai, Zhang, and Zhou [20]) Suppose log p = o(n) and p ≥ nξ for some ξ > 0; then we have the following minimax convergence rate 1. under the Frobenius risk/normalized MSE: 1 p inf Σ̂ sup Σ∈G(α,C,C0) E ∥∥∥Σ̂− Σ∥∥∥2 F n− 2α+12(α+1) ; (4.3) 2. under the spectral risk: inf Σ̂ sup Σ∈G(α,C,C0) E ∥∥∥Σ̂− Σ∥∥∥2 n− 2α2α+1 + log p n , (4.4) 74 4.2. Comparison Between Tapering and Shrinkage Estimators where the infimum is taken over all possible estimators Σ̂ : Rn×p → Rp×p for Σ based on the data. It is very interesting and important to ask how we can construct CMTs that actually attain the minimax risks. Fortunately, it turns out that there exists such a CMT with different bandwidths (to be defined shortly) that is rate-optimal for each of the two risks in the minimax sense. More specifically, let W = (wij) be defined as wij = 1, for |i− j| ≤ kh 2− |i− j|/kh, for kh < |i− j| < k 0, for |i− j| ≥ k , (4.5) where kh = k/2. First, we can see that such defined W is a valid matrix taper according to Definition 4.2.1 since diag(W ) = 1 and therefore p∑ j=1 λj(W ◦ Σ) = Tr(W ◦ Σ) = Tr(Σ) = p∑ j=1 λj(Σ). Second, it is clear that such defined W and hence W ◦ Σ for every Σ vanish off the stripe that is at most (k− 1) away from the main diagonal. Therefore, k is defined as the bandwidth of the CMT. Third, it should be noted that W ◦Σ does not necessarily preserve the positive definiteness of Σ. However, this concern can be mitigated by first diagonalizing W ◦ Σ and then replacing its negative eigenvalues by zeros. This modification preserves the minimax error bounds (up to a constant of 2) and also the positive definiteness of the resulting estimate. It is noted that the optimal procedures are different under the Frobenius and spectral norms. It has been shown that optimal bandwidths of W under the normalized Frobenius and spectral norms are n1/2(α+1) and n1/(2α+1), respectively. 75 4.2. Comparison Between Tapering and Shrinkage Estimators 4.2.2 Shrinkage Estimator Since the discovery of the Stein’s effect on the inadmissibility of the multivariate normal mean vector by the usual sample mean estimator when p ≥ 3 (see [112] for the original reference), extensive research has been devoted to proposing a broad range of shrinkage estimators such as the James-Stein estimator [71], its truncated version [8], among many others, to improve the performance of the usual estimator in terms of risks induced by a variety of loss functions. Similarly as in the estimation of the mean vector, the sample covariance estimator Ŝ, as we have mentioned, is unsatisfactory for large (high-dimensional) covariance estimation problems. Steinian shrinkage therefore has been an alternatively attractive choice. Estimators of this kind naturally have the form Σ̂(ρ) = (1− ρ)Ŝ + ρT, (4.6) where ρ ∈ [0, 1] is the shrinkage coefficient and T is the shrinkage target matrix. In general, T is supposed to have the properties of being: (i) well-conditioned; (ii) consistent or even optimal in a subspace of p×p symmetric matrices. In other words, the shrinkage estimator is a convex combination between the sample covariance matrix and a “good” target matrix. There are several possible and intuitive choices of T . For instance, we consider T = F̂ := p−1Tr(Ŝ)I, and the shrinkage estimator has the following form Σ̂(ρ) = (1− ρ)Ŝ + ρF̂ . (4.7) Chen et.al [36] defines an MMSE oracle estimator Σ̂o := Σ̂(ρ̂o) where ρ̂o is defined as the solution of the optimization problem ρ̂o = argminρ∈[0,1] E ∥∥∥Σ̂(ρ)− Σ∥∥∥2 F subject to Σ̂(ρ) = (1− ρ)Ŝ + ρF̂ . 76 4.2. Comparison Between Tapering and Shrinkage Estimators The MMSE oracle estimator seeks the best convex combination between the sample covariance matrix and a scaled identity matrix to approximate the true covariance matrix in terms of the mean-squared errors (MSEs). This estimator is said to be an oracle because the optimal solution depends on Σ which is unknown in practice and is the estimation goal. It is shown in [82] that ρ̂o can be given by a distribution-free formula ρo = E[Tr((Σ− Ŝ)(F̂ − Ŝ))] E‖Ŝ − F̂‖2F . (4.8) Under additional Gaussian assumption, the closed-form of ρo is given in [36] ρo = p− 2 + pt p(n+ 1)− 2 + (p− n)t , (4.9) where t = Tr2(Σ)/Tr(Σ2). (4.10) Here t measures the distribution of the off-diagonal entries of Σ. In particular, Tr(Σ2) ≤ Tr2(Σ) ≤ pTr(Σ2), where equalities of the left and right inequalities are attained if and only if Σ = 11T and Σ = I, respectively. So when t = 1, the matrix entries have the most spread support (dense); while when t = p, the energy of Σ concentrates on the diagonal (sparse) . A second shrinkage estimator proposed in [53] combines Ŝ and T = diag(Ŝ) in the same manner as in (4.7). These two estimators share similar properties since the optimal coefficient can be obtained in a single distribution-free framework. Therefore, in the rest of the paper, we focus on the identity target case in (4.7) which is easier and more expressive for our theoretic analysis. 77 4.2. Comparison Between Tapering and Shrinkage Estimators First, we derive the Frobenius risk of the MMSE oracle estimator (4.7), assuming that the data are from i.i.d. N(0,Σ). Theorem 4.2.2. Suppose {xi}ni=1 are i.i.d. Gaussian N(0,Σ) . The Frobenius risk of the MMSE shrinkage oracle estimator (4.7) is given by E‖Σ̂o − Σ‖2F = [ (1− t p )ρo + 2 np ] ‖Σ‖2F . (4.11) From Theorem 4.2.2, we can see that the Frobenius risk of the shrinkage oracle estimator primarily depends on ρo and the property of Σ. The second term in (4.11) contributes negligibly to the total risk when Σ is bounded away from the identity matrix, where t < p. Since it is difficult for us to derive the exact formula, we also derive a lower bound on the risk under the spectral norm. Theorem 4.2.3. Suppose {xi}ni=1 are i.i.d. Gaussian N(0,Σ) . The spectral risk of the MMSE shrinkage oracle estimator (4.7) satisfies E‖Σ̂o − Σ‖2 ≥ ρ2o(1− λmin(Σ))2. (4.12) Theorem 4.2.2 and 4.2.3 are important in the sense that, by giving the pointwise explicit risk bounds of Σ in the parameter space S, it is possible for us to analyze the theoretic properties, such as consistency and admissibility, of the MMSE shrinkage oracle estimator (4.7). Indeed, we will shortly see that this shrinkage estimator is inconsistent for some high-dimensional covariance matrices that may often appear in many real-world applications; therefore, it is inadmissible for a subspace of the parameter set S and this suggests that we shall find alternative solutions. This is the main motivation of the proposed STO estimators which will be introduced shortly in Section 4.3. 78 4.2. Comparison Between Tapering and Shrinkage Estimators 4.2.3 Comparison of Risk Bounds Between Tapering and Shrinkage Estimators We are now ready to compare the risk bounds of the tapering and MMSE shrink- age oracle estimator, thanks to Theorem 4.2.1, Theorem 4.2.2, and Theorem 4.2.3. The comparison is done by studying several specific examples and several interesting conclusions can be drawn. We describe the examples as follows. Example 4.2.1. Consider, for 0 < γ < 1, σij = 1, for i = j,γ|i−j|, for i 6= j. (4.13) In words, the entries of Σ decay exponentially fast when moving away from the main diagonal. This example corresponds to the covariance structure of auto-regression models with order 1, AR(1), and is considered in [11, 36]. We can easily see that, for this Σ, Tr(Σ) = p by using the normalization assumption and Tr(Σ2) = ‖Σ‖2F p(1 + γ2)/(1 − γ2) by summing the squared `2 norm of all diagonals of Σ. More specifically, ‖Σ‖2F = p+ 2(p− 1)γ2 + 2(p− 2)γ4 + · · ·+ 2γ2(p−1) = 2 p−1∑ j=0 (p− j)γ2j − p = 1− γ2p 1− γ2 × 2p− C0 − p→ 1 + γ2 1− γ2p, for p being sufficiently large. Therefore, it follows that t = Tr2(Σ) Tr(Σ2) → 1− γ 2 1 + γ2 p := Cp, as p→∞. (4.14) But this then implies that the Frobenius risk p−1E‖Σ̂o − Σ‖2F in the super-linear 79 4.2. Comparison Between Tapering and Shrinkage Estimators high-dimensional situation is asymptotically, as n→∞, p→∞, and n/p→ 0, C−1 [ (1− C) Cp 2 + p− 2 Cp2 + (1− C)np+ p− 2 + 2 np ] → C−1 − 1 := C(γ) > 0 where C(γ) = 2γ2/(1 − γ2), since C ∈ (0, 1). Therefore we can conclude that the Frobenius risk is 1 p E‖Σ̂o − Σ‖2F = C(γ) + o(1). (4.15) It is clear that the normalized MSE is lower bounded by a positive constant depending on γ and therefore the MMSE shrinkage oracle estimator cannot be a consistency estimator of Σ unless the concentration p/n → 0. Figure 4.1(a) plots the finite sample size behavior of the normalized MSE and its limit (4.15). We can see that the normalized MSE asymptotically approaches to a non-zero value when n/p→ 0 with n, p being large enough. On the contrary, note that Σ in this example satisfies any smoothing parameter α ∈ (0,∞), we hence deduce that, under the Frobenius risk, the convergence rate, n−(2α+1)/(2α+1) in (4.3), of the tapering estimator Σ̂taper (4.1) with the minimax CMT W in (4.5) can be arbitrarily close to n−1. Comparing these two bounds, it is clear that, for this example, the tapering estimator is uniformly superior than the MMSE shrinkage oracle estimator proposed in [36]. Therefore, this oracle estimator is in fact a weak oracle which is overly restrictive in terms of the functional form of the shrinkage estimator. The reason that the tapering estimator outperforms the MMSE shrinkage oracle estimator is that the optimal estimate may not necessarily be decomposed as a simple convex combination between the sample covariance matrix Ŝ and the scaled identity matrix F̂ . Example 4.2.1 is a good evidence of this fact. Furthermore, noticing that (4.9) and (4.14), we have ρo C(γ)p 2 + p− 2 C(γ)p2 + (1− C(γ))np+ p− 2 → 1, when p n and p→∞, (4.16) 80 4.2. Comparison Between Tapering and Shrinkage Estimators Figure 4.1: Normalized MSE curves of the shrinkage MMSE estimator for the large covariances discussed in Example 4.2.1 and 4.2.2. The MMSE estimator fails to be consistent when n/p → 0, because the normalized Frobenius risks converge to the asymptotic values, calculated from (4.15) and (4.18), that are bounded away from 0. (a) Example 4.2.1 with γ = 0.5. (b) Example 4.2.2 with α = 0.3. we see from the eigen-structure perspective that the spectral risk of the MMSE shrink- age oracle estimator (4.7) obeys E‖Σ̂o − Σ‖2 ≥ (1 + o(1))(1− λmin(Σ))2, (4.17) which implies that E‖Σ̂o − Σ‖2 9 0 because λmin(Σ) is a monotone decreasing se- quence as p→∞. As a summary for Example 4.2.1, we conclude that, although the bona fide co- variance matrix in (4.13) has a diagonal-like structure, shrinkage of Ŝ to an identity matrix is in consistent and hence it is not a good choice in high-dimensional situations and this procedure shall be improved by taking into account more refined structural information. On the contrary, tapering is minimax in this example. We now study a second example that has a slower polynomial decay rate than in Example 4.2.1, as considered in [20]. 81 4.2. Comparison Between Tapering and Shrinkage Estimators Example 4.2.2. Consider, for α > 0, σij = 1, for i = j,|i− j|−(α+1), for i 6= j. Based on the definition in (4.2), we can show that Σ ∈ G(α−1, 1, C0). It follows from an analogous argument that ‖Σ‖2F = (1 + C1)p− C2, where C1 = 2 ∑p−1 j=1 j −2(α+1) > 0 and C2 = 2 ∑p−1 j=1 j −2α−1 > 0, both converging as p → ∞. The rest derivation proceeds as in Example 4.2.1 and consequently we can achieve the argument that there exists a constant C(α) > 0 such that 1 p E‖Σ̂o − Σ‖2F = C(α) + o(1), (4.18) as illustrated in Figure 4.1(b). Similar to the analysis in Example 4.2.1, we can see that the tapering estimator outperforms the MMSE shrinkage oracle estimator under both Frobenius and spectral risks. Again, tapering estimator is minimax and MMSE shrinkage-to-identity is inconsistent in this example when the dimensionality is large. Example 4.2.3. In a third example, we consider the covariance structure of a frac- tional Brownian motion (FBM) with the Hurst parameter h ∈ [0.5, 1]: σij = 2 −1[(|i− j|+ 1)2h − 2|i− j|2h + (|i− j| − 1)2h]. The FBM is a model for complex systems that have long-range dependence for h being close to 1, such as modeling the internet traffic [85]. Practical applications usually tune h between 0.5 and 0.9. The covariance matrix in this model does not belong to G(α,C,C0) unless h = 0.5, which is the case of the Brownian motion with 82 4.2. Comparison Between Tapering and Shrinkage Estimators white Gaussian noise process; therefore, tapering estimator does not guarantee to be consistent estimator in this example. To see this, we first observe that σij ≥ 0 since x2h is convex in x for h ∈ [0.5, 1] and then obtain that ‖Σ‖1 = p∑ i=1 p∑ j=1 σij = p+ p−1∑ k=1 (p− k)[(k + 1)2h − 2k2h + (k − 1)2h] = p+ p p−1∑ k=1 { (k + 1)2h − k2h − [k2h − (k − 1)2h]} − p−1∑ k=1 { (k + 1)2h+1 − k2h+1 − [k2h+1 − (k − 1)2h+1]} + p−1∑ k=1 [ (k + 1)2h − (k − 1)2h] = p+ p [ p2h − 1− (p− 1)2h]− [p2h+1 − 1− (p− 1)2h+1]+ [p2h + (p− 1)2h − 1] = p2h, where the second last equality follows from three telescope sums. Consequently, it follows that max j ∑ |i−j|≥1 σij ≥ p−1 ∑ 1≤i6=j≤p σij = p 2h−1 − 1, where the last term is not summable for h > 0.5 as p→∞. But now, we clearly see from definition (4.2) that Σ /∈ G(α,C,C0). On the other hand, since the covariance matrix in Example 4.2.3 is Toeplitz, its Frobenius norm is given by ‖Σ‖2F = p+ 1 2 p−1∑ j=1 (p− j) [(j + 1)2h − 2j2h + (j − 1)2h]2 . For x ≥ 1, let us define f as a function of h: f(h) := fx(h) = (x+ 1) 2h − 2x2h + (x− 1)2h. It is clear that f is continuous, f(1/2) = 0 and f(1) = 2. Consequently we have 83 4.3. A Shrinkage-to-Tapering Estimator ‖Σ‖2F = p when h = 0.5 and ‖Σ‖2F = p2 when h = 1. For h ∈ (0.5, 1), because the function (x2h lnx) is asymptotically convex for any 0.5 < h < 1 as x → ∞, it follows from the Jensen’s inequality that f ′(h) ≥ 0 for sufficiently large x. Therefore, we deduce that f(·) will eventually be an increasing function between [0, 2] for h ∈ (0.5, 1), as x diverges to infinity. So t→ p as h→ 0.5, while t→ 1 as h→ 1. By Theorem 4.2.2, for the MMSE shrinkage oracle estimator, we now have p−1E‖Σ̂o − Σ‖2F = [ (1− t p )ρo + 2 np ] ‖Σ‖2F p = ( p t − 1)ρo + 2 nt = p− t t × p− 2 + pt p(n+ 1)− 2 + (p− n)t + o(1). Therefore, if t = p, i.e. Σ = I, then the Frobenius risk vanishes to zero and the MMSE shrinkage oracle estimator is a consistent estimator; if t = 1, i.e. Σ = 11T , the Frobenius risk is asymptotically 2p/n, meaning that the Frobenius risk for estimating this large covariance matrix depends on the concentration p/n. 4.3 A Shrinkage-to-Tapering Estimator 4.3.1 Problem Formulation Motivated by the above discussions, we propose a Steinian shrinkage type estimator. With the important difference from the shrinkage estimator toward a scaled identity matrix, the proposed estimator shrinks the sample covariance matrix to its tapered version. Basically the proposed estimator subsumes T = diag(Ŝ) in [53] as one special case where W = I. Specifically, the proposed estimator Σ̂STO := Σ̂(ρ̂STO) has the form Σ̂(ρSTO) = (1− ρSTO)Ŝ + ρSTO(W ◦ Ŝ), (4.19) 84 4.3. A Shrinkage-to-Tapering Estimator where ρSTO is determined by the solution to the optimization problem ρSTO = argminρ∈[0,1] E|||Σ̂(ρ)− Σ|||2 subject to Σ̂(ρ) = (1− ρ)Ŝ + ρ(W ◦ Ŝ). Here ||| · ||| can be either Frobenius or spectral norm. By using the tapering estimator as our shrinkage target, we hope this estimator can inherit good properties from both tapering and shrinkage estimators. Throughout the rest of the paper, we shall refer this proposed estimator as the shrinkage-to-tapering oracle (STO) estimator. On one hand, for Σ ∈ G(α,C,C0), we can see from Theorem 4.2.1 that the pro- posed STO estimator reduces to the tapering estimator for large n and p. On the other hand, for Σ /∈ G(α,C,C0), the proposed estimator reduces to an analogy of the MMSE shrinkage oracle estimator. Therefore, we expect that, for an arbitrary large covariance matrix Σ, the proposed estimator could improve upon both tapering and MMSE shrinkage oracle estimators. The optimal coefficient of the STO estimator can be given in a closed-form. Theorem 4.3.1. The coefficient of the proposed STO estimator under the minimum Frobenius risk is ρ̂STO = E(‖Ŝ‖2F − ‖V ◦ Ŝ‖2F )− (‖Σ‖2F − ‖V ◦ Σ‖2F ) E‖Ŝ‖2F + E‖W ◦ Ŝ‖2F − 2E‖V ◦ Ŝ‖2F , (4.20) where V = (vij) with vij = √ wij. Under further Gaussian assumption, we can write (4.20) in a closed-form given as in (48). 4.3.2 Approximating the Oracle The proposed STO estimator is nice in theory for developing the closed-form expres- sion of the optimal coefficient ρ̂STO. Nevertheless, in practice, the true Σ is the target of estimation and thus unknown. So the proposed oracle estimator is not feasible in practice. Therefore, we propose a practical algorithm that approximates the oracle estimator. Following the idea of [36], we define a shrinkage-to-tapering oracle ap- 85 4.3. A Shrinkage-to-Tapering Estimator proximating (STOA) estimator as an iterative procedure between the following two steps: 1. ρ̂STj+1 = [ Tr(Σ̂jŜ)− Tr((V ◦ Σ̂j)(V ◦ Ŝ)) + Tr2(Σ̂j)− Tr(D̂jV 2D̂j) ] /[ (n+ 1)(Tr(Σ̂jŜ) + Tr((W ◦ Σ̂j)(W ◦ Ŝ))− 2Tr((V ◦ Σ̂j)(V ◦ Ŝ))) +Tr2(Σ̂j) + Tr(D̂jW 2D̂j)− 2Tr(D̂jV 2D̂j) ] , (4.21) where D̂j is a diagonal matrix such that D̂j = diag(Σ̂j). 2. Σ̂j+1 = (1− ρ̂STj+1)Ŝ + ρ̂STj+1(W ◦ Ŝ). (4.22) With an appropriate initialization, the two steps are operated iteratively until the sequence {ρ̂STj } converges. Then the STOA estimator is defined as using its limit ρ̂STOA = lim j→∞ ρ̂STj . Currently, for the proposed STOA estimator, we are unable to derive a rigorous theory concerning the convergence as in the oracle approximation shrinkage (OAS) estimator case in [36]. Our empirical experience in the Simulation Section 5.5, however, demon- strates that the STOA algorithm can approximate the STO estimator reasonably well for a broad range of Σ, regardless of its sparsity. For the proposed estimators, another issue is to determine the bandwidth k ofW in the tapering step for calculating the shrinkage target matrixW ◦ Ŝ. We adopt a data- driven approach for estimating k. The procedure is as follows: We randomly split the independent data into two subsets and choose k from a set of candidate values. For each k in the chosen set, ρ̂STOA is estimated based on one data subset which we call the 86 4.4. Simulation training data set and then we calculate the distance, e.g. induced by the Frobenius or spectral norm, between the estimated Σ̂ and the sample covariance matrix computed from the other data set, i.e. the testing data set. Finally the optimal k is determined by the index yielding the smallest distance. Due to the extra step of determining k, the proposed estimator is more computationally expensive than the LW, RBLW, OAS, and MMSE shrinkage oracle estimators. Nonetheless, this validation overload in practice is a minor computational issue because the proposed STO and STOA estimators are quite efficient for any pre-specified k and the computational cost of all shrinkage estimators mentioned in this paper are comparable. We conclude this section by providing the STOA pseudo-code in Algorithm 2. Algorithm 2: The STOA algorithm. Input: Ŝ, kmax Output: Σ̂STOA begin foreach k = 0 : 2 : kmax do Construct the CMT W with bandwidth k as in (4.5) ; Initialize from Ŝ and calculate the optimal shrinkage coefficient by iterating (4.21) and (4.22) until convergence ; Find the best bandwidth of W and corresponding ρ̂STOA by the minimum prediction error on the test data ; end Return Σ̂STOA = (1− ρ̂STOA)Ŝ + ρ̂STOA(W ◦ Ŝ). end 4.4 Simulation We simulate the three examples discussed earlier in this paper to study the finite sample size numeric performances of the proposed estimators. We fix p = 100 for all models and consider different values of n with n = {10, 20, 30, 40, 50}. The STOA 87 4.4. Simulation algorithm is initialized at Σ̂0 = Ŝ and ρ = 0.5. The maximum number of iterations in the STOA algorithm is set to be 10. We compare the proposed STO estimator and its variant STOA with the tapering [20] and several shrinkage estimators including the LW [82], Rao-Blackwellized LW (RBLW) [36], MMSE shrinkage oracle (MMSEO) estimator and its variant oracle approximating shrinkage (OAS) [36] estimator. 4.4.1 Model 1: AR(1) Model The ith-row and jth-column entry of the covariance matrix Σ is σij = γ |i−j|. (4.23) We chose γ = {0.5, 0.7, 0.9}. A smaller γ essentially makes Σ more like the identity matrix. Note that for any γ ∈ (0, 1), Σ ∈ G(α,C,C0) for all α > 0 and some C,C0 > 0. To specify the tapering bandwidth, we need to determine a proper α by data-driven approaches. Here, our tapering estimator is performed on a training data set over a pre-defined grid and then the optimal α is selected by minimizing the Frobenius loss ‖Σ̂taper − Σ‖F on the oracle. (In practical applications, we can use the random splitting scheme discussed above). Estimated normalized MSEs, i.e. the Frobenius risk, and the spectral risk are plotted in Figure 4.2 and Figure 4.3 for various aforementioned estimators. Several interesting observations can be made from Figure 4.2 and Figure 4.3. First, in terms of estimation risks, the STO, STOA, and tapering estimators uniformly improve upon the previous shrinkage-type estimators including LW, RBLW, OAS, and the MMSEO. This validates our Theorem 4.2.2 and Theorem 4.2.3 on finite sample size data. The improvement is visually appreciable even when n is not so large as considered in the asymptotic setup. Second, the proposed STO and STOA also outperform the tapering estimator, although the improvement is smaller than those from the previous shrinkage-type estimators. The improvement on Frobenius 88 4.4. Simulation Figure 4.2: Model 1: The normalized MSE curves as a function of n, averaged over 100 replications. The tapering [20], LW [82], RBLW [36], MMSE shrinkage oracle (MMSEO) [36], and OAS [36] are compared with the proposed STO and STOA estimators. (a) γ = 0.5 (b) γ = 0.7 (c) γ = 0.9 89 4.4. Simulation Figure 4.3: Model 1: The spectral risk curves as a function of n, averaged over 100 replications. Here the legends are the same as in Figure 4.2. (a) γ = 0.5 (b) γ = 0.7 (c) γ = 0.9 risk is slightly more significant than that on spectral risk. Third, it is clear from these two figures that STOA can well approximate the STO estimator. Finally, despite that the STO estimator minimizes the MSE, the results from the spectral risk are similar to that of the Frobenius risk. It suggests that the STO and STOA are robust against the norm under which the risk is minimized. The estimated shrinkage coefficients ρ̂STO and ρ̂STOA are also plotted in Figure 4.4. It is observed that, in general, the coefficients of STO and STOA are closer to 1 than other shrinkage estimators. This means that STO and STOA essentially use W ◦ Ŝ as the estimator, with a slight adjustment by incorporating information directly from Ŝ. This is confirmative to the theory we have seen since the tapering estimator is minimax. Moreover, shrinkage coefficients of LW, RBLW, MMSEO, and OAS estimators tend to decrease as n increases. This makes sense because the more data collected, the larger amount of information should be used from Ŝ, in which case a 90 4.4. Simulation Figure 4.4: Model 1: The estimated shrinkage coefficients for different estimators, averaged over 100 replications. The legends are the same as in Figure 4.2, except that the tapering estimator is excluded and STO/SOTA and STO 2/STOA 2 are the STO/STOA estimates under the Frobenius and spectral risks, respectively. (a) γ = 0.5 (b) γ = 0.7 (c) γ = 0.9 smaller value of ρo shall be adaptively chosen. On the contrary, we do not see this observation for the STO and STOA estimators. In fact, their coefficients seem to converge to 1 in this empirical study, as seen in Figure 4.4. This is due to the fact that the shrinkage target, W ◦ Ŝ, of these two estimators actually contains the data information. When W ◦ Ŝ is truly optimal, then it is sufficient for the STO and STOA estimators to use only the target component and thus the optimal coefficients converges to 1 in this example. 4.4.2 Model 2: Σ ∈ G(α−1, C, C0) We have σij = 1 for i = j0.6|i− j|−(α+1) for i 6= j , (4.24) 91 4.4. Simulation Figure 4.5: Model 2: The normalized MSE curves as a function of n, averaged over 100 replications. (a) α = 0.1 (b) α = 0.3 (c) α = 0.5 (d) α = 0.7 where we choose the smoothing parameter α from α = {0.1, 0.3, 0.5, 0.7}. We sim- ply set k = bn1/2(α+1)c in order to achieve the optimal convergence rate under the Frobineus norm. We remark that the numeric performance can be further improved by cross-validating on a set of bandwidths on the order n1/2(α+1). The risk results of different estimators are shown in Figure 4.5 and Figure 4.6 and the estimated shrink- age coefficients are shown in Figure 4.7. Again, we observe the same pattern on the error curves and essentially same conclusions can be drawn as in Model 1. 4.4.3 Model 3: Fractional Brownian Motion The numeric performance of the STO and STOA estimators when Σ /∈ G(α,C,C0) is studied in the third model, the FBM model. In our setup, we look at the FBM with the Hurst parameter h selected from h = {0.6, 0.7, 0.8, 0.9}. From Figure 4.8, we can see that the normalized MSEs of the MMSE shrinkage 92 4.4. Simulation Figure 4.6: Model 2: The spectral risk curves as a function of n, averaged over 100 replications. (a) α = 0.1 (b) α = 0.3 (c) α = 0.5 (d) α = 0.7 93 4.4. Simulation Figure 4.7: Model 2: The estimated shrinkage coefficients for different estimators, averaged over 100 replications. (a) α = 0.1 (b) α = 0.3 (c) α = 0.5 (d) α = 0.7 94 4.4. Simulation Figure 4.8: Model 3 (FBM): The normalized MSE curves as a function of n, averaged over 100 replications. (a) h = 0.6 (b) h = 0.7 (c) h = 0.8 (d) h = 0.9 95 4.5. Conclusion estimators are smaller than that of the tapering estimator. This is not surprising because: (i) the assumption Σ ∈ G(α,C,C0) is violated and therefore no optimality under the Frobenius risk can be expected in the tapering estimator; (ii) the MMSE es- timators are designed to minimize the Frobenius risk. Notwithstanding, when looking at the spectral risk, Figure 4.9, we observe that the risk of the tapering estimator is smaller than those from the MMSE family. Therefore, the tapering estimator is quite robust in the sense that, although being sub-optimal, it still gives better spectral risk performances than the MMSE shrinkage estimators. In contrast, the MMSE shrink- age estimators are sensitive to norms under which the risk performance is measured; in particularly, they are only optimal in the Frobenius norm. It is observed that the STO and STOA estimators uniformly outperform other shrinkage estimators when h = 0.8 and h = 0.9. In the case of h = 0.6, they are outperformed by LW, RBLW, OAS, and MMSEO estimators but still yield smaller MSEs than the tapering estimator. The case of h = 0.7 appears to be non-uniform; however the curve trends shown in Fig 4.8(b) suggest that STO and STOA may eventually yield a smaller Frobenius risk as n gets larger. 4.5 Conclusion The main contributions of this chapter are summarized as follows: 1. For high-dimensional covariance estimation problems where p/n → ∞, we showed that the MMSE shrinkage oracle estimator is inconsistent under both Frobenius and spectral risks for some typical covariance matrices in G(α,C,C0). Moreover, we showed that the tapering estimator is uniformly superior than the MMSE shrinkage estimator in this case. 2. We proposed a STO estimator that combines the advantages from both the MMSE shrinkage and tapering estimators. In particular, the proposed estimator is suitable for estimating general, high-dimensional covariance matrices. An 96 4.5. Conclusion Figure 4.9: Model 3 (FBM): The spectral risk curves as a function of n, averaged over 100 replications. (a) h = 0.6 (b) h = 0.7 (c) h = 0.8 (d) h = 0.9 97 4.5. Conclusion Figure 4.10: Model 3 (FBM): The estimated shrinkage coefficients for different esti- mators, averaged over 100 replications. (a) h = 0.6 (b) h = 0.7 (c) h = 0.8 (d) h = 0.9 98 4.5. Conclusion oracle estimator in the closed-form was derived and a practical algorithm to approximate the STO estimator was presented. 99 Chapter 5 Efficient Minimax Estimation of High-Dimensional Sparse Precision Matrices 5.1 Introduction In this chapter, we primarily focus on estimating the inverse of the covariance matrix Σ−1, a.k.a. the precision matrix Ω, in high-dimensional situations. Estimation of Ω is a more difficulty task than estimating Σ because of the lack of natural and pivotal estimators as Σ?n when p > n. Nonetheless, accurately estimating Ω has important statistical meanings. For example, in Gaussian graphical models, a zero entry in the precision matrix implies the conditional independence between the corresponding two variables. Further, there are additional concerns in estimating Ω beyond those we have already seen in estimating large covariance matrices; for details see Chap- ter 1. In light of those challenges in estimating the precision matrix when p n, we propose in this paper a new easy-to-implement estimator with attractive theoretic properties and computational efficiency. The proposed estimator is constructed on the idea of the finite Neumann series approximation and constitutes merely matrix multiplication and addition operations. The proposed estimator has a computational complexity of O(log(n)p3) for problems with p variables and n observations, repre- senting a significant improvement upon the aforementioned optimization methods. So our estimator is more promising for ultra high-dimensional real-world applications 100 5.1. Introduction such as gene microarray network modeling. Remark 11. It is possible to further reduce the computational complexity of the pro- posed algorithm by employing more sophisticated matrix multiplication algorithms. For instance, the current fastest matrix multiplication algorithm by Coppersmith and Winograd has an asymptotic complexity of O(p2.376) [38]. Moreover, by explor- ing the sparsity structure in the matrices, the complexity can be further reduced to O(k0.7p1.2 + p2+o(1)) [129], where k is the maximum number of zeros in each of the multipliers. Therefore, we can see that there would be huge computational savings of our algorithm when the covariance matrix is sufficiently sparse. We now state the assumption regarding the sparse matrices studied in this paper. Our sparse matrix class are built on standard sparse matrices. For p k, a p-by-p matrix A with elements aij’s is said to be k-sparse if A ∈ Sk where Sk = { A : sup j ∑ i I(aij 6= 0) ≤ k } . (5.1) Sk is a strict class in the sense that there are matrices containing many small entries while they are dense in support and thus excluded from Sk. Therefore, we choose to consider an alternative sparsity measure in terms of the strong `q-ball introduced in [11]. Define Gq(cn,p) = { A : sup j ∑ i |aij|q ≤ cn,p } , (5.2) for 0 ≤ q < 1, be the collection of matrices with each column belonging to a strong `q-ball with size cn,p. Note that Gq(cn,p) is closed under the matrix L1 norm. Think of matrices under consideration are infinite dimensional for a moment. Subset of sparse matrices can be naturally defined as those matrices with finite strong `q-ball sizes. Therefore, the set Gq of all possible finite strong `q-ball volumes is our main target of study Gq = ⋃ cn,p≥0 Gq(cn,p). (5.3) 101 5.1. Introduction 5.1.1 Innovation and Main Results We summarize the main innovation of this chapter in Theorem 5.1.1, which is an immediate consequence of a series of asymptotic analysis to be reported in Section 5.3. Briefly speaking, we shall describe a computationally efficient algorithm to estimate large precision matrices in a certain approximately inversely closed sparsity class we introduce and show that the resulting estimator is consistent when more and more data are collected. Furthermore, by deriving a lower bound on the estimation error of the precision matrix, the proposed estimator is shown to actually achieve this information-theoretic lower bound and therefore it is rate-optimal. Theorem 5.1.1. Assume p ≥ nξ for some ξ > 0 and cn,p ≤ C(log p/n)(1−q)/2. Then the minimax risk of estimating the precision matrix Ω = Σ−1 on {Σ ∈ Gq(cn,p) ∩ U(m)}, where U(m) is defined in (5.8) and the rows of the data Xn,p follow i.i.d. sub-Gaussian distribution with covariance Σ, obeys inf Ω̂ sup Σ∈Gq(cn,p)∩U(m) E ∥∥∥Ω̂− Ω∥∥∥2 c2n,p( log pn )1−q , (5.4) where the infimum is taken over all possible estimator Ω̂ : Rn×p → Rp×p for Ω based on the data. Furthermore, the proposed estimator based on the Neumann series rep- resentation achieves this minimax risk. 5.1.2 Comparison with Existing Work It is interesting to observe that the same error bound (5.4) applies to the estimation of the covariance matrix for Σ ∈ Gq(cn,p) [22]. Indeed, a closer examination on our proofs (given in the Appendix) reveals that we actually translate the estimation problem of Ω to the estimation of Σ. Since the latter case is well studied in the literature, there are powerful tools and solid theories to be used for our purpose. Therefore, we will show that our proposed estimator of Ω inherits a large portion of nice theoretic properties from estimation of Σ, such as consistency and rate-optimality [20, 22]. 102 5.1. Introduction The CLIME estimator of Ω, Ω̂CLIME proposed in [18], is the solution of the con- strained convex optimization problem minimize ‖Ω‖1 subject to ‖Σ?nΩ− I‖∞ ≤ λn, (5.5) followed by a symmetrization step in order to make Ω̂CLIME a self-adjoint matrix. Under a different set of assumptions which are imposed merely on Ω, the CLIME estimator has similar convergence rate as our proposed estimator, E ∥∥∥Ω̂CLIME − Ω∥∥∥2 . c2n,p( log pn )1−q , (5.6) where cn,p is now the size of the precision matrices in their uniform class. Both estima- tors achieve the optimal convergence rate under the spectral norm; nevertheless, our proposed estimator only consists of thresholding, matrix multiplication and addition operations which require no essential computational overload besides the calculation of Σ?n. Therefore, the proposed estimator is more computationally efficient than the CLIME estimator and thus can be applied to large-scale precision matrix estimation problems. Similar spectral/Frobenius norm convergence results in probability were reported for the graphical Lasso and SCAD models in the special case that q = 0 [80, 108]. Therefore, our results are more general in the sense that we obtain optimal conver- gence results for a broader class of sparse matrices in terms of strong `q balls with small size. The rest of the chapter is organized as following: Section 5.2 introduces the notion of approximately inverse closeness on the set of sparse matrices and identifies a class of such matrices. Section 5.3 proposes a precision matrix estimator based on the Neumann series representation and proves that it is consistent in probability and in L2 under the spectral norm. Moreover, the minimax risk of estimating the precision matrix is studied. By comparing the error bound of our proposed estimator with 103 5.2. Approximately Inversely Closed Sparse Matrices the minimax risk, we show that our estimator is sharp and thus rate-optimal in the sense of minimax risk. In Section 5.4, we discuss the issue of practically determining tuning parameters in the proposed algorithm. Performance comparisons with other optimization based methods using simulations are reported in Section 5.5. A real fMRI application for learning functional brain connectivity of F→STN using the proposed method is presented in Section 5.6. We conclude this paper in Section 5.7 and discuss a few directions for future work. 5.2 Approximately Inversely Closed Sparse Matrices In order to estimate the precision matrix from multivariate Gaussian observations, it is necessary to assume that Σ is non-singular. We consider the following uniform class U(m,M) = {Σ 0 : m ≤ λmin(Σ) ≤ λmax(Σ) ≤M} , (5.7) where Σ 0 means Σ is strictly positive-definite and m,M > 0. Without loss of generality (w.l.o.g.), it is convenient to assume in the sequel that M = 1/m and 0 < m ≤ 1. Therefore, we have the class U(m) = {Σ 0 : m ≤ λmin(Σ) ≤ λmax(Σ) ≤ 1/m} . (5.8) 5.2.1 The Neumann Series Representation of Ω Let Σ ∈ U(m). Set MΣ = λmax(Σ), mΣ = λmin(Σ), and η = 2/(MΣ +mΣ). Since Σ is positive-definite and self-adjoint, it is clear that η minimizes ‖I − tΣ‖ over t > 0 and ‖I − ηΣ‖ = MΣ −mΣ MΣ +mΣ ≤ 1−m 2 1 +m2 < 1, (5.9) 104 5.2. Approximately Inversely Closed Sparse Matrices as shown in [12]. So it follows from the Neumann series expansion that Ω = Σ−1 = η(ηΣ)−1 = η[I − (I − ηΣ)]−1 = η ∞∑ j=0 (I − ηΣ)j def= Br +Rr, where Br = η r∑ j=0 (I − ηΣ)j (5.10) and the residual term is upper bounded by ‖Rr‖ ≤ η ∞∑ j=r+1 ‖I − ηΣ‖j = 1 mΣ ( MΣ −mΣ MΣ +mΣ )r+1 ≤ 1 m ( 1−m2 1 +m2 )r+1 → 0 (5.11) uniformly in Σ ∈ U(m) as r →∞. For our problem of estimating sparse precision matrices, the Neumann represen- tation of Ω motivates us to identify a class of sparse matrices such that the inverse of its any member can be approximated with arbitrary accuracy by elementary linear combinations using only a finite number of members in the class. This idea shall be rigorously formalized in the next section where the notion of approximately inverse closeness is introduced. 5.2.2 A Class of Sparse Matrices with Approximately Inverse Closeness The main contribution of this section is to introduce a class of approximately sparse matrices whose inverses are also approximately sparse. Definition 5.2.1. The set of all k-sparse matrices that are at most ε-distance from A is defined as: Sparseε(A, k) = {B : ‖B − A‖ ≤ ε and B ∈ Sk} . (5.12) Moreover, we generalize the definition to consider the family of all k-sparse ma- 105 5.2. Approximately Inversely Closed Sparse Matrices trices that are within ε-distance from Gq(cn,p) Sε(Gq(cn,p), k) = ⋃ A∈Gq(cn,p) Sparseε(A, k). (5.13) It is clear from the definition that B ∈ Sε(Gq(cn,p), k) if and only if B ∈ Sk and dist (B,Gq(cn,p)) ≤ ε. (5.14) Finally, let Sε(Gq(cn,p)) = ⋃ k≥0 Sε(Gq(cn,p), k) (5.15) be the collection of sparse matrices that can approximate Gq(cn,p) with error ε. Lemma 5.2.1. Given 0 ≤ q < 1, cn,p > 0, and ε > 0, we let A ∈ Gq(cn,p) and kmin = [ q (1− q)εc 1 q n,p ] q 1−q . (5.16) Then the following statements hold: 1. There exists B ∈ Sparseε(A, k) for all k ≥ kmin, i.e. Sparseε(A, k) is non-empty. 2. For each r ∈ N, Ar ∈ Gq(C(q)rcrn,p). In particular, for every k′ ≥ kminC(q) r 1−q c r−1 1−q n,p . (5.17) there exists B′ ∈ Sparseε(Ar, k′). Definition 5.2.2. A collection S of invertible elements (i.e. S−1 exists for all S ∈ S) is said to be inversely closed if S−1 ∈ S for all S ∈ S. It has been shown in [12] that there is a class of band dominated matrices that is inversely closed. In general, for sparse matrices, raising arbitrarily high powers for a 106 5.2. Approximately Inversely Closed Sparse Matrices sparse matrix can cause the elements mixing so well that the sparsity of its inverse gets violated. Therefore, we define below a relaxed version of the inverse closeness. Definition 5.2.3. A collection S of invertible elements is said to be approximately inversely closed if for any ε > 0 and S ∈ S, there is an S ′ ∈ S such that dist ( S−1, S ′ ) ≤ ε. (5.18) Now, consider the uniform class F(q,m) = Gq ∩ U(m). (5.19) It consists of all bounded linear functionals Rp → Rp that are: a) within ε-distance of the set Gq(cn,p); b) uniformly bounded away from singularity and thus invertible; c) permutation-invariant. Beyond these facts, there is one more crucial yet less obvious fact that F(q,m) is actually approximately inversely closed. This is the main theorem of this section: Theorem 5.2.2. F(q,m) is approximately inversely closed. Suppose Σ ∈ Gq(cn,p) ∩ U(m), then for any ε > 0, there exists k′ = kminC(q) r 1−q c r−1 1−q n,p (5.20) for sufficiently large r such that Sparseε(Ω, k ′) is non-empty. We now summarize here the assumptions assumed in this paper: a) both n and p diverge to∞ and n . pξ; b) the observation data X = {xi}ni=1 are sampled from some i.i.d. sub-Gaussian distribution (5.21) with covariance matrix Σ; c) Σ ∈ Gq(cn,p) for some cn,p > 0 such that the precision matrix Ω can be approximated within ε distance by some sparse-k′ matrix. 107 5.3. Proposed Estimator Figure 5.1: The diagram illustrating our proposed Algorithm 3. White spots corre- spond to zero entries, blue ones are the positive entries, and red ones are negative entries in the matrix. For a sparse Σ with approximately inverse closeness, its inverse Ω can be approximated by the finite Neumann series representation. So consistent procedures such as thresholding applied to Σ?n can be used to form a “good” estimator for Ω. Since Ω is not necessarily sparse in the true sense (5.1), an additional trun- cation is applied to estimate Ω because its strong `q volume (5.2) can be controlled, e.g. by Theorem 5.2.2. 5.3 Proposed Estimator 5.3.1 Algorithm The proposed algorithm is based on the Neumann series representation of Ω and the thresholding operator Tt defined through Tt(Σ)(i, j) = σijI(|σij| > t). A diagram illustrating main ideas of the proposed algorithm is shown in Figure 5.1. 108 5.3. Proposed Estimator Now, the proposed estimator Ω̂ is described in Algorithm 3. Algorithm 3: The proposed algorithm Input: Sample covariance matrix Σ?n, thresholding cutoff t, number of truncated Neumann series terms r, approximation tolerance ε. Data: X. Output: Ω̂ begin1 Compute Σ̃n = Tt(Σ ? n).2 Set Ω̃ = η ∑r j=0(I − ηΣ̃n)j.3 Truncate Ω̃ according to (50) such that Ω̂ ∈ Sparseε(Ω̃, k′), where k′ is4 given in (5.20) (i.e. keep the largest k entries of each column of Ω̃ and set others to zeros). end5 The output of the above algorithm is an Sk′ matrix due to the last truncation step. With properly determined ε, we can show that the truncation step does not affect the error bounds when it is removed from the algorithm. The purpose of adding this step is to promote sparsity in the true sense such that the estimator Ω̂ can be clearly interpreted. For instance, a zero ωij is interpreted as the conditional independence between two variables in the Gaussian graphical model (and hence a missing edge in the graph of Ω̂). It is still left open to determine the input parameters t and r and estimate η as well (ε can be chosen as a small value). For practical choices of these parameters, we adopt common data-driven approaches such as cross-validation. Regarding this issue of parameter tuning, we will provide the details in Section 5.4. 5.3.2 Consistency Here following similar ideas as in [11, Theorem 1], we show that the proposed esti- mator Ω̂ is consistent. In fact, we shall show its consistency for data X generated 109 5.3. Proposed Estimator by some i.i.d. sub-Gaussian distribution (a.k.a. rows of X are i.i.d. sub-Gaussian vectors) which is a slightly more general result than xi ∼ N(0,Σ). More specifically, we assume that there exist absolute constants C1, C2 > 0 such that P (∣∣vTxi∣∣ ≥ t) ≤ C1 exp(− t2 2C2 ) (5.21) holds for all t > 0 and ‖v‖ = 1. Assume, w.l.o.g., Exi = 0. With this sub- Gaussian assumption (5.21), standard concentration of measure results (e.g. see [120, Proposition 16 and Corollary 17]) enable us to bound the tail probability of (σ?ij−σij) in a mixture of exponential and Gaussian-like decaying rate. More precisely, we have Proposition 5.3.1. Suppose X is sub-Gaussian obeying (5.21) with covariance Σ ∈ U(m). Then there exist constants C3, C4, C5 > 0, all depending only on C1 and C2 in (5.21) and m, such that P (∣∣σ?ij − σij∣∣ ≥ t) ≤ C3 exp [−C4min( t2C25 , tC5 ) n ] (5.22) for all t ≥ 0. Focusing on a neighborhood of zero, we can see that the large deviation result in [22, Eq.(25)] immediately follows from Proposition 5.3.1. Corollary 5.3.2. Under assumptions in Proposition 5.3.1, P (∣∣σ?ij − σij∣∣ ≥ t) ≤ C3 exp(−8nt2C24 ) (5.23) for |t| ≤ C5. Then applying [11, Theorem 1] we obtain the following theorem. Theorem 5.3.3. Suppose X is sub-Gaussian obeying (5.21) with covariance Σ ∈ Gq(cn,p) ∩ U(m). Assume log p/n → 0 and cn,p(log p/n)(1−q)/2 → 0, as n → ∞. 110 5.3. Proposed Estimator Choose the threshold parameter t = τ √ log p/n for some large τ . Then we have, uniformly in Σ ∈ Gq(cn,p) ∩ U(m), ∥∥∥Ω̂− Ω∥∥∥ ≤ C(q,m, τ)cn,p( log p n )(1−q)/2 + 1 m ( 1−m2 1 +m2 )r+1 , (5.24) with probability greater than 1 − C6p−8τ2/C24+2 approaching to 1 whenever τ > C4/2. Here, r is the number of terms in the Neumann series pre-estimator Ω̃ of Ω̂. An immediate consequence of Theorem 5.3.3 is that Ω̂ is a consistent estimator of Ω whenever cn,p(log p/n) (1−q)/2 = o(1) as n → ∞, and the optimal number of terms in the Neumann power series is chosen such that the two terms on the right hand side of (5.24) match on the same magnitude order. Therefore, we have Corollary 5.3.4. Let r = ⌈ (1− q)(log n− log log p)− 2 log cn,p − 2 logC 2[log(1 +m2)− log(1−m2)] − 1 ⌉ , (5.25) for some C > 0. Under the same assumptions as in Theorem 5.3.3, then for every Σ ∈ Gq(cn,p) ∩ U(m), ∥∥∥Ω̂− Ω∥∥∥ P→ 0, (5.26) as n→∞. Corollary 5.3.4 follows straightforward after a few algebra based on Theorem 5.3.3; thus we omit its proof. We also show below the entry∞-norm consistency of Ω̂ which shall be particularly helpful for developing the model selection consistency shortly. Proposition 5.3.5. Suppose X is sub-Gaussian obeying (5.21) with covariance Σ. Assume log p/n → 0 as n → ∞. Choose the threshold parameter t = τ√log p/n for some large τ . Then we have, uniformly in Σ ∈ Gq(cn,p) ∩ U(m), ∥∥∥Ω̂− Ω∥∥∥ ∞ ≤ C(q,m, τ) (√ log p n + δr+1 ) , (5.27) 111 5.3. Proposed Estimator with probability greater than (1 − p−8τ2/C24+2) which asymptotically approaches to 1 whenever τ > C4/2; here δ = (1−m2)/(1 +m2). Remark 12. Proposition 5.3.5 states that the maximal fluctuation of the estimator {ω̂ij} about {ωij} can be well controlled. Therefore, when the magnitudes of non-zero entries of Ω are uniformly bounded away from zero, we can recover the support of Ω by cutting Ω̂ with a properly determined threshold t′ according to (5.27). The cutoff t′ can be chosen such that the recovery is successful with probability tending to 1, as more and more data are available. Based on Theorem 5.3.3, we also provide the consistency of our proposed estimator under the mean squared loss, which shall be useful to establish the upper bound for optimality in the sense of minimax risk. We mention that establishing the upper bound of estimation error is considerably more involved than the weaker claim where the same bound is stated in probability. Theorem 5.3.6. Under assumptions in Theorem 5.3.3 and in addition assuming p ≥ nξ for some ξ > 0 and r = O(log n), we have sup Σ∈Gq(cn,p)∩U(m) E ∥∥∥Ω̂− Ω∥∥∥2 . c2n,p( log pn )1−q + δ2(r+1). (5.28) One consequence of Theorem 5.3.6 is Corollary 5.3.7. With assumptions in Corollary 5.3.4, sup Σ∈Gq(cn,p)∩U(m) E ∥∥∥Ω̂− Ω∥∥∥2 . c2n,p( log pn )1−q → 0, (5.29) as n→∞. Remark 13. Since convergence in L2 implies P→, it is now clear that Corollary 5.3.4 is an immediate consequence of Corollary 5.3.7. 112 5.3. Proposed Estimator 5.3.3 Sharpness: Optimal Under Minimax Risk We have so far seen that the proposed estimator is consistent under the spectral norm. In fact, we will also show that it is rate-optimal in the sense of minimaxity among all estimators of Ω. In order to establish the minimaxity of the proposed estimator, it suffices, thanks to Corollary 5.3.7, to find a lower bound on the order of c2n,p (log p/n) 1−q for the mean squared loss. Then this would imply that our proposed estimator is sharp and thus rate-optimal. Now, we tackle this task by assembling ideas in [20–22]. More specifically, we appeal to a general lower bound argument for the minimax risk of estimating the sparse covariance matrix, see [22, Lemma 3], and then convert the optimality in terms of precision matrix as in [20]. For the completeness of our proof, it is worthy to spend a few paragraphs to describe the construction setup and recap the related lemma, from which our lower bound can follow easily. The lower bound is established via a carefully constructed finite set of the least fa- vorable multivariate normal distributions that are in the sparse uniform class Gq(cn,p)∩ U(m). For 1 ≤ J ≤ p, we denote B ⊂ Rp \ {0} as a non-zero subset in the p- dimensional vector space. Let Γ = {0, 1}J be the vertex set of the J-dimensional cube and Λ ⊂ BJ . Define the product parameter space Θ by Θ = Γ× Λ = {θ = (γ, λ) : γ ∈ Γ, λ ∈ Λ} . (5.30) Then every J × p matrix D can be identified by a mapping Θ→ RJ×p such that the parametrization of D by θ = (γ, λ) is interpreted as follows: D is formed by stacking each of the p-dimensional vector γ1λ1, · · · , γJλJ , where λ = (λ1, · · · , λJ). Moreover, a set of association operators Aj : Rp → Rp×p are defined via M = Aj(b) such that the j-th row and column of M are equal to b and other entries of M are all zeros. Now, combining the product structure of Θ and the association operators, we define 113 5.3. Proposed Estimator a function Σ : θ ∈ Θ 7→ Σ(θ) ∈ Rp×p as Σ(θ) = I + ε J∑ j=1 γjAj(λj). (5.31) For J ≤ dp/2e and a set of {λj}1≤j≤J where each λj has zeros in the first (p − J) positions and the rest entries have either 0 or 1 such that ‖λj‖0 = k (for some k to be assumed bounded appropriately), it is easy to see that Σ(θ) is an anti-block-diagonal matrix except on the main diagonal where σjj(θ)’s= 1. This constitutes the collection of the least favorable multivariate normal distributions that hopefully attain the worst estimation error. Now, we are ready to define a family of matrices that shall be used to obtain the minimax risk lower bound via (5.31): G0 = {Σ = Σ(θ) : θ ∈ Θ} . (5.32) It is not hard to verify that, for a carefully chosen k and ε small and depending on cn,p, we have G0 ⊂ Gq(cn,p) and G0 ⊂ U(m), where the latter is because of diagonal dom- inance. Finally, to complete the setup, for a given b ∈ {0, 1}, a mixture distribution P̄j,b, associated with Θ, can be defined as P̄j,b(X; Θ) = 1 2J−1|Λ| ∑ {θ∈Θ:γj(θ)=b} P (X | Σ(θ)), (5.33) where γj(θ) projects θ = (γ, λ) to the j-th element of γ. Essentially, (5.33) defines a mixture of distributions over all parameter sets with a common projection onto γj. With the above notation, [22, Lemma 3] gives a general lower bound of estimating sparse covariances under the L2 loss. Proposition 5.3.8. (Cai and Zhou [22]) Let Θ be the parameter space of θ con- structed in (5.30) and (5.31). For an arbitrary function ψ(θ), let U be any estimator 114 5.3. Proposed Estimator of ψ(θ) from data X generated from the probability family {Pθ : θ ∈ Θ}. Then max θ∈Θ Eθ ‖U − ψ(θ)‖2 ≥ α 4 J 2 min 1≤j≤J ∥∥P̄j,0 ∧ P̄j,1∥∥ , (5.34) where α = min (θ,θ′):H(γ(θ),γ(θ′))≥1 ‖ψ(θ)− ψ(θ′)‖2 H(γ(θ), γ(θ′)) (5.35) and H(γ, γ′) is the Hamming distance defined on {0, 1}J H(γ, γ′) = J∑ j=1 |γj − γ′j|. (5.36) In light of Theorem 5.3.3 and Theorem 5.3.6, we have similar results for estimating the precision matrix as those obtained for estimating covariance matrix [22, Theorem 1]. Theorem 5.3.9. Suppose cn,p ≤ C(log p/n)(1−q)/2. Then the minimax risk of esti- mating the precision matrix Ω = Σ−1 on {Σ ∈ Gq(cn,p) ∩ U(m)}, where data Xn,p are i.i.d. sub-Gaussian with covariance Σ, obeys inf Ω̂ sup Σ∈Gq(cn,p)∩U(m) E ∥∥∥Ω̂− Σ−1∥∥∥2 ≥ Cc2n,p( log pn )1−q , (5.37) where the infimum is taken over all possible estimator Ω̂ for Ω and here the constant C is independent of n and p. Now, it is clear that our main Theorem 5.1.1 is a direct consequence of Corol- lary 5.3.7 and Theorem 5.3.9. 5.3.4 Model Selection Consistency We have so far shown the matrix L2 norm consistency of the proposed estimator. Estimation consistency in the spectral norm does not imply the model selection con- 115 5.3. Proposed Estimator sistency and vice versa. Here by model selection consistency we mean that P ({ω̂ij 6= 0} = {ωij 6= 0})→ 1, (5.38) as n → ∞. Therefore, it is interesting to ask whether or not the proposed estima- tor can accurately recover the bona fide statistical structures. This is of particular importance when we need to identify the structure in graphical models. To establish the model selection consistency, it suffices to have the consistency of estimator under the entry ∞-norm. As we have seen that this is indeed the case in Proposition 5.3.5. Hence, it leads to the following theorem. Theorem 5.3.10. Let ω = min{|ωij| : ωij 6= 0} and r = ⌈ log n− log log p− 2 logC 2[log(1 +m2)− log(1−m2)] − 1 ⌉ , (5.39) for some C > 0. Suppose ω > 2t′. Then Tt′(Ω̂) is model selection consistent. In fact, it is clear that Theorem 5.3.10 implies the signs of {ωij} can also be recovered with high probability. 5.3.5 Extensions The proposed algorithm framework described in Algorithm 3 can be easily extended to broader situations where additional information regarding the sparsity is available. For instance, when there is an ordering structure between variables such as in auto- regression (AR) models, the covariance has a bandable structure. By incorporating this additional information, the proposed estimator can be modified such that even better theoretic properties can be achieved. More precisely, by replacing the thresh- olding operator in the current Algorithm 3 with the tapering operator, the optimal convergence rate min { n−2α/(2α+1) + log p n , p n } (5.40) 116 5.4. Practical Choices of η, r and τ can be accomplished by adapting our arguments in Theorem 5.3.3 and 5.3.6 to ac- commodate the optimality results found for the estimator of Σ in [20]. Here α is a sparsity control parameter specifying the rate of decay of entries moving away from diagonals. We can show, by essentially the same arguments, that this tapering esti- mator is minimax for the covariance matrices of this form. The proposed framework can also be extended to the adaptive thresholding case, which has been shown to yield better numeric performances on real data when the homoscedastic assumption appears too restrictive [17]. By allowing location-specific cutoffs which could be estimated in a data-driven means, the adaptive thresholding procedure is shown to be optimal when the heteroscedasticity is indeed present. In this case, the universal (non-adaptive) thresholding operator is suboptimal. In view of these feasible extensions which can properly handle different real prob- lems, we can see that the proposed algorithm is in fact a quite general and flexible framework. Moreover, the framework can be easily adapted in a way such that var- ious optimality may be achieved while the attractive low computational complexity is maintained. 5.4 Practical Choices of η, r and τ The construction of the proposed estimator involves determining the parameters η, r and τ . For the choice of η, we use the estimation η̂ = 2/(MΣ̃n + mΣ̃n). Note that the thresholding operator Tt does not necessarily preserve positive-definiteness which means mΣ̃n < 0; nevertheless, Tt with a carefully chosen t can preserve positive- definiteness with high probability [11]. In the exceptional case where mΣ̃n < 0, we can simply project negative eigen-values to 0 and the error bound of approximating Σ remains unchanged (except for a factor of 2). To determine r and τ , we employ a random splitting procedure on a two-dimensional grid j = (j1, j2). n data points are randomly partitioned into two sets: one training 117 5.5. Numerical Experiments set of size n1 and one test set of size n2 = (n−n1). The precision matrix is estimated on a collection of tuning parameters r and τ and the optimal tuning parameters r̂ and t̂ are then determined by maximizing the normal log-likelihood on the test data set, up to an additive constant, (ĵ1, ĵ2) = arg max j=(j1,j2) log-likelihood ( Ω̂n1,rj1 ,tj2 ; Σ ? n2 ) = arg min j=(j1,j2) { trace ( Σ?n2Ω̂n1,rj1 ,tj2 ) − log det(Ω̂n1,rj1 ,tj2 ) } , (5.41) where Ω̂n1,rj1 ,tj2 represents the proposed estimate of the precision matrix Ω on the training data with tuning parameters rj1 and tj2 , and Σ ? n2 means the sample covariance calculated on the test data. 5.5 Numerical Experiments In this section, some simulations are conducted to evaluate the performance of the proposed estimator. We consider a Toeplitz model of Ω with entries ωij’s decaying exponentially fast as they moving away from diagonals. This kind of models arises naturally in time-series data analysis where a natural ordering on variables is present. More specifically, we choose ωij = a |i−j| with a = 0.6 for our setup. This is tantamount to assume a moving average model by noting that the covariance matrix Σ has a band structure σij = (1 + a 2)/(1− a2), for i = j, −a/(1− a2), for i 6= j. (5.42) In our case, we therefore have σii = 2.1250 (except for the first and last diagonal elements) and σij = −0.9375 for i 6= j. Note that this is also one of the models considered in [18]. We compare the performance of our algorithm with several main- stream optimization-based methods including CLIME, graphical Lasso, and SCAD. Since performances of those optimization methods on this model have been thor- oughly studied and become virtual standards, e.g. in accordance to [18] and [108], in 118 5.5. Numerical Experiments Table 5.1: Estimation error under the spectral norm, specificity, and sensitivity of Ω̂, Ω̂taper, CLIME, graphical Lasso (GLasso), and SCAD for n = 100. p Spectral Norm Loss Ω̂ Ω̂taper CLIME GLasso SCAD 30 2.28 1.25 2.28 2.48 2.38 60 2.78 1.44 2.79 2.93 2.71 90 2.90 1.60 2.97 3.07 2.76 120 2.97 1.67 3.08 3.14 2.79 200 3.01 1.78 3.17 3.25 2.83 p Specificity % Ω̂ Ω̂taper CLIME GLasso SCAD 30 55.28 99.99 78.69 50.65 99.26 60 90.11 100.00 90.37 69.47 99.86 90 99.86 100.00 94.30 77.62 99.88 120 99.88 100.00 96.45 81.46 99.91 200 99.88 100.00 97.41 85.36 99.92 p Sensitivity % Ω̂ Ω̂taper CLIME GLasso SCAD 30 56.28 67.20 41.07 60.02 16.93 60 21.54 62.08 25.96 41.72 12.72 90 11.81 59.13 20.32 33.70 11.94 120 11.30 58.94 17.16 29.32 11.57 200 10.87 57.06 15.03 25.34 11.07 the same setup as ours, we only run our algorithm and directly compare with their performances reported under the same setting (see Table 1, 2, and 3 in [18]). We synthesize 100 training data points and another independent 100 test data points. p is varied over {30, 60, 90, 120, 200}. We tune the parameters t and r on a two-dimensional grid. We choose t = τ √ log p/n for τ ranging over [0, 2] evenly spaced with interval size 0.2, and choose r to be integers in the region [0, 3dlog ne]. Then the optimal parameter pair (t̂, r̂) are determined by maximizing the normal log- likelihood (5.41) on the test data. We further set the effective zero level to be 10−3, meaning that estimated entries with absolute values below this level are thought as zeros. All performance results are averaged over 100 simulations and they are shown in Table 5.1 and Figure 5.2, along with the performances reported in [18, Table 1, 2, and 3] for comparison. First, we examine the estimation performances in terms of the spectral norm 119 5.5. Numerical Experiments loss, the specificity, the sensitivity, and the Mathews correlation coefficients (MCC). From Table 5.1, we see that the proposed estimator Ω̂ improves upon CLIME and graphical Lasso, while being slightly outperformed by the SCAD approach which is most computationally expensive. Secondly, by looking at the specificity (i.e. true negative rate TNR, the proportion of claimed negatives that are true negative) and the sensitivity (i.e. true positive rate TPR, the proportion of detected positives that are true positive), we further observe that the proposed estimator behaves like SCAD for larger p. By comparing these two performance measures separately, we argue that it is difficult to arrive at a safe conclusion concerning the existence of a uniformly superior estimator, at least from a numeric perspective. To jointly examine the specificity and the sensitivity, we also calculate the MCC. The MCC is defined as TP× TN− FP× FN√ (TP + FP)× (TP + FN)× (TN + FP)× (TN + FN) . (5.43) MCC values are always between -1 and 1; with 1 being perfect prediction and -1 being the worst performance. From Figure 5.2, it is clear that for large-scale problems, the proposed estimator Ω̂ and SCAD outperform CLIME and graphical Lasso. The MCCs of the proposed estimator and SCAD are nearly identical. We note that it has been empirically observed by existing literature that SCAD often tends to produce sparser estimates. Nevertheless, in terms of computational cost, our proposed algorithm is more preferred than SCAD for ultra large-scale problems. Further, we show that the performance of the proposed estimator Ω̂ can be im- proved by taking advantage of the Toeplitz structure of the covariance and precision matrices (as additional information) and replacing thresholding by the tapering pro- cedure proposed in [20]. From Table 5.1 and Figure 5.2, we can see that, incorpora- tion of the bandable structure into the tapering version Ω̂taper can significantly and uniformly improve the estimation performances within our framework. Again, it is worthy emphasizing that in this simulation setup, Ω̂taper can be proved to remain optimal in the minimax risk sense under the spectral norm. 120 5.5. Numerical Experiments Figure 5.2: The Mathews correlation coefficients (MCC) of estimating Ω when using the proposed algorithm based on the Neumann series representation, its tapering version, CLIME, graphical Lasso (GLasso), and SCAD. (a) True Ω (b) Ω̂ (c) Ω̂taper Figure 5.3: True sparse precision matrix Ω with p = 200. Estimated precision matrix Ω̂ based on the Neumann series and its tapering version Ω̂taper with n = 100, averaged over 100 replications. 121 5.6. Application to An Real fMRI data We also plot the entries of Ω in Figure 5.3. A visual inspection shows that the pattern recovered by our proposed algorithm preserves the true structure. On one hand, the high specificity of our proposed algorithm reflects its ability to detect essentially all effective zeros in Ω. On the other hand, the low sensitivity of Ω̂, as well as CLIME, graphical Lasso, and SCAD, comes from the fact that there are many non-zero but small off-diagonal entries in Ω which makes the high accuracy of model selection more difficult to achieve than a truly sparse precision matrix. This observation can be well justified by Theorem 5.3.10, where we require the minimal magnitude of ωij be greater than a certain threshold to achieve the model selection consistency. 5.6 Application to An Real fMRI data 5.6.1 Modeling F→STN The importance of the “hyperdirect pathway” (from the frontal cortex directly to the subthalamic nucleus (STN), i.e. F→STN) is being increasingly recognized in Parkinson’s disease (PD). The hyperdirect pathway has been implicated in impulse control problem. During deep brain stimulation (DBS) surgery, a small electrode is implanted into the STN and an electrical current delivered to disrupt noisy activity in this brain structure. Traditionally, the STN was considered accessible only through DBS. More contemporary work has suggested that there is a direct connection be- tween the frontal cortex in the outer part of the brain and the STN: the hyperdirect pathway. Moreover, despite the small size of the STN, a recent fMRI study in sub- jects without PD has suggested that it is possible to measure activity in the STN. In the current study, we are interested in test the hypothesis that the strength of connection between frontal cortices and the STN will be significantly different when subjects perform a motor task involving a sudden “stop” command compared to the same motor task where a stop command is not issued. We expect there is some con- 122 5.6. Application to An Real fMRI data nections in the stop task; but not in the control task. In addition, we shall focus on inferring direct connectivity between frontal cortices and the STN, rather than the result of indirect influence, e.g. via the thalamus. Now, we formulate the model. Our goal is to construct connections A → B | C, where A, B, and C are pre-defined brain regions. Here, A = frontal cortex (F), B = STN, and C = thalamus. In words, we would like to learn the brain connectivities directly from A to B, by removing the indirect effect of connections of A to B via C; this is exactly tantamount to learn a sparse precision matrix of the three regions. Since the pixels are highly correlated within neighborhoods and their sizes are very large, we first apply PCA to reduce the dimensionality. More specifically, we look at the eigen- pixels and learn sparse connections between those eigen-pixels. In particular, in our experiments, we used 10 PCs for each region; then we combine the 10 PCs from the three regions (A, B, C) and run our proposed model to learn a sparse 30×30 precision matrix Ω. As we have seen, a nonzero entry in precision matrix implies an edge in its Gaussian graphical model representation, which in turn implies a connection between the corresponding two eigen-pixels. Therefore, we are interested in the non-zero entries of Ω between rows 1-10 and columns 11-20. We run the proposed method based on the Neumann series representation on this data set. 2-fold CV is used to determine the optimal threshold and number of terms in the Neumann series. 5.6.2 Learned F→STN Connectivities We first plot the estimated precision matrix Ω for two normal subjects N005 and N006, see Figure 5.4. There are one connections identified for each normal subject in the stop task: PC2 of A↔PC1 of B in N005 and PC1 of A↔PC2 of B in N006. In contrast, there is no connectivity detected by our model in the control task for both subjects. This is in accordance with the biological knowledge that expects connections in the stop task but not in the control task for normal individuals. Second, the patterns of identified PCs that are connected in the stop tasks for 123 5.6. Application to An Real fMRI data (a) N005, stop task (b) N006, stop task (c) N005, control task (d) N006, control task Figure 5.4: Estimated precision matrix Ω for two normal subjects N005 and N006. Each subject performs two task: stop and control. From upper left block to bottom right block: A, B, C. We are interested in non-zero entries in the upper middle block: rows 1-10 and columns 11-20. 124 5.7. Conclusion and Discussion (a) N005, stop task (b) N006, stop task Figure 5.5: The patterns of identified PCs that are connected in the stop task for two normal subjects N005 and N006. N005 and N006 are also shown in Figure 5.5. It is clear that the associated PCs between A and B are highly correlated with each other. Finally, we plot the loadings of the identified PCs in Figure 5.6. We can see that there is a clear clustering property for the original pixels associated with the identi- fied PCs, implying that there are a few clusters of spatially close pixels connecting together. 5.7 Conclusion and Discussion We presented a conceptually simple and computationally efficient algorithm on esti- mation of large sparse precision matrices. The proposed estimator is motivated by identifying a class of sparse matrices that is approximately inversely closed. Our theoretic analysis showed that the proposed estimator for this class is statistically valid and optimal in the sense of minimax risk under the spectral norm. We further showed that the proposed estimator is model selection consistent which is a direct consequence of the established convergence result on the entry∞-norm. Then, simu- lation results demonstrated the encouraging performances of the proposed estimator when compared with state-of-art optimization based methods. Finally, the proposed method was applied to learn direct brain connectivity based on fMRI data and yielded 125 5.7. Conclusion and Discussion (a) N005, PC2 of A (b) N005, PC1 of B (c) N006, PC1 of A (d) N006, PC2 of B Figure 5.6: Loadings of the identifed PCs for two normal subjects N005 and N006. 126 5.7. Conclusion and Discussion biologically plausible results. We would like to point out a few directions for future work. An interesting study following the current work can be on better determining the parameters using a data- driven approach. We will need to analyze its theoretic performance in comparison with an oracle that allows us to know the true covariance/precision matrix in advance of observing any data. We expect that the work in [11] could shed us some light on this future direction. Another future direction is to extend the current work to the complex case. So far we have developed our framework for real data, however, we note that there is nothing preventing extending the obtained results to the complex case, as long as the concentration of measure inequality (5.22) continues to be valid. This requirement is generally true (see the theoretic ground laid in [83]). Extension to the complex case will allow us to apply the proposed framework to a broad range of statistical signal processing problems [1, 87, 114]. Finally, it is also interesting to consider estimating time-varying sparse Ω(t) in the proposed framework, e.g. for modeling fMRI brain networks. 127 Chapter 6 fMRI Group Analysis of Brain Connectivity 6.1 Introduction Studying brain connectivity is crucial in understanding brain functioning and can provide significant insight into the pathophysiology of a number of neurological disor- ders. Increasingly, inferring brain connectivity using functional Magnetic Resonance Imaging (fMRI) is being explored, and many mathematical formalisms, such as struc- tural equation modeling (SEM) [93], multivariate autoregressive models (mAR) [65], dynamic causal modeling (DCM) and dynamic Bayesian networks (DBNs) [88] have been proposed. Despite significant progress during the last decade, there are still a number of challenges associated with inferring brain connectivity from fMRI. One is the curse of complexity with the above SEM and/or mAR approaches when dealing with practical fMRI data sets where the number of brain regions-of-interest (ROIs) is relatively large and the number of time points is limited. Based on a number of neu- roscience studies, the connections between brain regions generally can be considered a priori to form a sparse network, suggesting that sparsity should be incorporated into brain connectivity modeling. For instance, sparse mAR models were studied in [118] where the parameters are estimated using penalized regression. Group analysis of effective brain connectivity has long been another challenging topic, since biomed- ical research is usually conducted at a group level to extract the population features. Efficient group analysis requires appropriate handling of expected inter-subject vari- 128 6.1. Introduction ability without destroying inter-group differences. To address the above two crucial challenges, this chapter aims at developing a novel, computationally-efficient brain connectivity model that incorporates both sparsity and suitable group analysis. Several methods for meaningfully extracting group information from fMRI data have been proposed. The common structure (CS) model in the DBNs context [88] enforces the same graphical structure for all subjects within a given group, but the connection coefficients are allowed to vary on a subject-by-subject basis. However, CS inference based on DBNs inevitably requires computationally-intensive algorithms such as the Markov Chain Monte Carlo (MCMC). Another proposed method is that of Bayesian group analysis [113] where several possible model structures are considered and the posterior evidence of models for each subject is estimated. Here we present a different group linear regression model to perform group analysis while incorporating the sparsity principle. More specifically, we adopt the modeling concept on the CS level – whereby brain connections generating the fMRI observations are assumed to be structurally identical among subjects within the same group, but individual connection parameters are allowed to vary between subjects – and propose a group robust Lasso framework to perform group analysis. There are several advantages associated with the proposed novel framework: 1. The proposed model is based on the optimization of a convex objective function and thus is computationally more efficient than graphical modeling approaches such as DBNs. 2. The proposed model represents a unified framework whereby group analysis is based on networks learned directly from the time courses in fMRI data. 3. The proposed model is robust against large variance noise and outliers. Remark 14. We note that the proposed group robust Lasso approach to infer brain connectivity networks is not based on inverse covariance matrix. Nonetheless, this marginal neighborhood selection procedure has been shown to be a consistent variable 129 6.2. Methods selection method for constructing a Gaussian graphical model under certain regularity conditions [95]. 6.2 Methods 6.2.1 A Group Robust Lasso Model We propose inferring brain connectivity through a linear regression approach. The Blood Oxygen Level Contrast (BOLD) signal intensity at a target ROI is regarded as a response which is modeled by a linear combination of time courses from ROIs subjected to the corruption by certain noise: y = Xβ + e. (6.1) Here y is a response vector,X is a design matrix with columns representing predictors, β is a coefficient vector, and e is a zero-mean random error vector which is assumed to have iid elements with a common finite variance, σ2. We consider the situation where the number of potential predictors is large while the number of bona fide predictors with non-zero coefficients are only a small fraction of the total. Thus, the goal is to determine the correct underlying sub-model. The Lasso [115] is a popular linear model selection tool that continuously shrinks coefficients to zeros. The Lasso is a regularized linear model and minimization of its `1 penalized squared `2 loss is known to promote sparsity on the coefficient vector. Nevertheless, the Lasso solution can be unsatisfactory for group analysis because its selection of a predictor is relatively independent of each other and therefore the Lasso estimator in unable to incorporate the potential similarity of structures across subjects. Yet for group analysis, we have certain structural grouping information that is available to us as a priori, e.g. the subjects within the same group are assumed to share the same connectivity structure. Therefore, β is composed of G groups each of which contains pg individual coefficients 130 6.2. Methods for g ∈ {1, · · · , G}. In matrix notation, β = (β∗1, · · · ,βG∗)∗, and X = (X1, · · · ,XG) is a block design matrix of dimension n ×∑g pg. With this notation, we can refer to grouping selection to mean that the sparsity is promoted at the group level, i.e. corresponding subject-specific coefficients within one group are either all non-zeros or all zeros. Thus the Lasso is a special case of the group version when pg = 1 for all g. To promote sparsity at the group level, we choose to minimize the following ob- jective function: f(β) = L (β;y,X) + λn G∑ g=1 ∥∥βg∥∥`2 (6.2) where L(·) can be any cost function. Unlike the `1 penalty in the Lasso, summation of block Euclidean norms (a.k.a. blocked `1 norm) in the penalty term encourages grouping selection [124]. The group Lasso is a the special case where L is the standard squared `2 loss [128]. More generally, we can adopt robust losses that are less sensitive to noise that includes large variability or even outliers. For instance, the convex combination of `1 and squared `2 losses [31] or the Huber loss [34] coupled with the block `1 regularization yields a group robust Lasso. In this paper, we propose a group robust Lasso by using the convex combined loss with a robustness tuning parameter δ ∈ [0, 1] L (β;y,X) = (1− δ) ‖y −Xβ‖`1 + δ ‖y −Xβ‖ 2 `2 . (6.3) The group Lasso is thus a reduced case when δ = 1. In general, a smaller δ gives more robustness. In the case of robust Lasso (pg = 1,∀g), the asymptotic behavior of its estimator has been studied in [31] where it is shown therein that the asymptotic variance is stabilized. Furthermore, the robustness tuning parameter can be chosen by the minimal asymptotic variance criterion when the error distribution is known. The proposed group robust Lasso estimator is defined to be any minimizer of (6.2), i.e. argminβ {f(β)} . Note that there is a corresponding model for each λn, so deter- mining a proper shrinkage amount is important to make the subsequential inference. The optimal shrinkage parameter λn is determined by the BIC which is computed as 131 6.2. Methods follows: we first solve for the group robust Lasso for a fixed λn. Once the model is determined, we fit the corresponding subset of data to the selected model by unreg- ularized least squares. We then obtain an estimator of β with the shrinkage effect removed. An estimator of σ2 is given by the maximum likelihood σ̂2 = L ( β̂;y,X ) (2− δ)n . The estimator (β̂, σ̂2) is called the Gauss group robust Lasso estimator which corrects for the bias of underestimating non-zero coefficients and thus is more suitable for accurate estimation. Note that the likelihood under which β̂ is computed is assumed to be Gaussian (equivalent to the least squares estimator) while the likelihood to estimate the variance, σ2, is a blend of Gaussian(0, σ2) and Laplace(2σ2) distributions. Now the BIC can be calculated from the Gauss group robust Lasso estimate BIC = −2× log-likelihood(β̂, σ̂2) + k log(n) = 1− δ σ̂2 ∥∥∥y −Xβ̂∥∥∥ `1 + δ σ̂2 ∥∥∥y −Xβ̂∥∥∥2 `2 + 2(1− δ)n log(4σ̂2) + δn log(σ̂2) + k log(n) (6.4) where k is the number of predictors in the selected model. Finally, the optimal model is chosen by the minimal BIC value among the set of different shrinkages. To summarize, proposed procedure contains the following steps: 1. Choose a set of shrinkage parameters λn. Run the group robust Lasso for each shrinkage. 2. For each shrinkage, identify the non-zero coefficients and use this submodel to compute (β̂, σ̂2). 3. Compute BICs using estimates from step 2 and choose the model corresponding to the minimal BIC value as the optimal model. 132 6.2. Methods We use the group robust Lasso as a model selection tool and estimate parameters based on the selected model. We refer this (variant) Gauss group robust Lasso as the grpRLasso in this paper unless otherwise indicated. 6.2.2 A Group Sparse SEM+mAR(1) Model The brain connectivity model we assume has a unified SEM framework that captures both spatial and temporal brain connections, where we combine the standard SEM model [93] (to represent the relations considered instanteous at the temporal resolu- tion of fMRI) and the 1st-order mAR model [65] (to represent longitudinal temporal relations). Suppose there are p ROIs and the brain is MRI scanned at time 1, · · · , T . We also assume that there are S subjects belonging to G groups. Denote by ys,j the fMRI measurement vector of the jth ROI of subject s, for j ∈ {1, · · · , p} and s ∈ {1, · · · , S}, as the response variable. Before introducing the group SEM+mAR(1) model, we introduce a few useful notations first. For the sth subject, let Y (0) s be the (T − 1) × p matrix with the jth column containing the fMRI measurements of the jth ROI from time 2 to T, and let the (T − 1) × p matrix Y (1)s be the time-shifted version of Y (0)s with lag 1. Y (0)s,−j denotes the (T − 1)× (p− 1) matrix with the jth column ys,j being removed. With these notations, for each subject s ∈ {1, · · · , S}, we have the following SEM+mAR(1) model: ys,j = Y (0) s,−jβ (0) s,j︸ ︷︷ ︸ SEM +Y (1)s β (1) s,j︸ ︷︷ ︸ mAR(1) +es (6.5) where es means the error vector for each subject s, and β (0) s,j and β (1) s,j represent respectively the SEM and mAR connection strength coefficients to be estimated. Putting all subjects together and rewriting in matrix form, we can reach the ambient linear regression model (6.6) with a block diagonal design matrix X and a 133 6.2. Methods coefficient vector β with a group structure. y1,j y2,j ... yS,j = diag { (Y (0) s,−j, Y (1) s ) }S s=1 β (0) 1,j β (1) 1,j β (0) 2,j β (1) 2,j ... β (0) S,j β (1) S,j + e1 e2 ... eS ︸ ︷︷ ︸ def =Xβ+e (6.6) where diag { (Y (0) s,−j, Y (1) s ) }S s=1 = Y (0) 1,−j Y (1) 1 Y (0) 2,−j Y (1) 2 . . . Y (0) S,−j Y (1) S . Now for each target ROI j, the proposed grpRLasso can be applied to (6.6) to learn a sparse coefficient vector with grouping structures. Brain connectivity networks are constructed by enumerating all the ROIs and our network analysis is based on the learned grpRLasso coefficient matrices. We give the asymptotic behavior of the proposed group robust lasso estimator. The obtained theoretic result justifies its usage. Essentially, we shall prove that the proposed group robust lasso can select the correct underlying model with probability approaching to one when a large sample size is available. In other words, the group robust lasso has the oracle property for model selection, namely without knowing the support of the true coefficient vector, the correct model can be identified with probability arbitrarily close to one. Compared with the group lasso, we shall see that the group robust lasso is robust against errors with large variability. 134 6.3. A Simulation Example Lasso RLasso grpLasso grpRLasso Oracle 2.99 2.89 9.62 1.66 1.66 (1.59) (1.63) (5.33) (0.52) (0.52) Table 6.1: MSEs for the estimated coefficients with their standard deviations shown in brackets. grpLasso abbreviates for the group Lasso,RLasso for the robust Lasso with the convex combined loss, and grpRLasso for the group robust Lasso. Oracle is the maximum likelihood estimate (for the Gaussian likelihood) obtained from the model with knowing the true non-zero locations. Theorem 6.2.1. Suppose that λn/ √ n→ 0 and λnn γ−12 →∞. Assume further that 1. n−1XTX→ C, where C is a positive definite matrix; 2. {ei} have a common continuous probability density function in a neighborhood of 0. Then the group robust Lasso estimator has the asymptotic normality on non-zero components: √ n ( β̂nA − βA ) ⇒ N ( 0, (1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10 4[δ + (1− δ)f(0)]2 C −1 11 ) . (6.7) Remark 15. Under further adaptive regularization weight on group coefficients with rate λg = ∥∥∥β̂LSng ∥∥∥−γ 2 , we can show the sign consistency P ( sgn ( β̂nA ) = sgn (βA) ) → 1, (6.8) as n→∞. 6.3 A Simulation Example A synthetic example is used to demonstrate empirical evidence for the improved de- tection and estimation performance of the proposed grpRLasso over the grpLasso, Lasso, and robust Lasso. A design matrix containing 400 observations and 200 pre- dictors is realized from a Gaussian Ensemble in each synthetic data set. The true 135 6.4. fMRI Group Analysis in Parkinson’s Disease sparse coefficient vector β0 is set to have a block structure. That is, β0 contains three blocks of non-zero coefficients of different magnitudes (1.5, -3, and 2, respectively) distributed in intervals located in [20, 30], [90, 95], and [180, 188], respectively. The shrinkage amount is set on an evenly spaced grid over [0, 2000]. BIC (6.4) is used to choose the optimal model. The error e is simulated from a Student-t distribution with parameters ν = 3 and σ2 = 9 so that it has a variance 27. The mean squared errors (MSEs) computed from 100 simulations are shown in Table 6.1. We note that the grpRLasso achieves the same performance as when the non-zero locations are known in advance (the “oracle property”). Hence, this implies that the proposed grpRLasso is very accurate and robust in terms of both model selection and parameter estimation. In contrast, the grpLasso model has a very large MSE, even worse than the Lasso. This is not surprising after a careful investigation on the nature of a grouping variable selection tool. Suppose that if the group Lasso falsely identifies a non-zero coefficient, then the rest elements in the group are all non-zeros which render a high estimation error. In contrast, since selection procedure of the Lasso is independent among predictors, incorrect selection of one variable has little influence on others. 6.4 fMRI Group Analysis in Parkinson’s Disease In this section, we apply the grpRLasso to a fMRI data collected from subjects with and without Parkinson’s disease (PD) and report the group analysis results of the learned brain connectivity networks. 6.4.1 Data Description fMRI scans for ten normal people and eight subjects with PD were collected in the study. Subjects were asked to continually squeeze a bulb in their right hand to control an inflatable ring so that the ring moved through an undulating tunnel without 136 6.4. fMRI Group Analysis in Parkinson’s Disease Figure 6.1: Weighted network for the normal group: solid lines are the SEM connec- tions and dashed lines the mAR(1) connections. Node labels are the names of ROIs with the prefixal L indicating the left brain hemisphere and R the right hemisphere. Line width is proportional to the connection strength. The thicker the edge, the larger the magnitude of the estimated coefficient. Figure 6.2: Different connections between normal and “off-medication” networks: dashed edges are only present in the normal network while dotted edges only in the “off-medication” network. Solid edges exist in both networks with significantly different means (t-test with size 0.05). 137 6.4. fMRI Group Analysis in Parkinson’s Disease touching the sides. PD subjects performed the same task after been withdrawn from their L-dopa medication for 12hrs. Images were acquired at a sampling rate of 0.5Hz and a trial lasted for five minutes so that 150 data points were obtained for each subject. 6.4.2 Learned Brain Connections The robustness tuning parameter δ is fixed to 0.5. It is worth noting that this pa- rameter can be further optimized if required [31]. For each target ROI, there are 35 possible directed edges pointing to it, 17 from SEM and 18 from mAR(1). Since we have the normal (10 subjects) and PD off-medication (8 subjects), a linear regres- sion model for each target node has 2 × 35 = 70 groups partitioning the total 630 coefficients. Generally speaking, the (robust) Lasso networks yielded many more connections that are inconsistent among subjects within a group. Hence they are less useful for group studies and not reported further. The network learned from grpRLasso for the normal group is shown in Figure 6.1 and the difference between the normal and PD group networks is shown in Figure 6.2. Edges shown in Figure1 have signifi- cant non-zero coefficients with a t-test (against zero mean) of size 0.05. There are four main findings of biological significance. First, as seen in Figure 6.1, there were many reciprocal connections between homologous regions in both groups (e.g. left supplementary motor area (L SMA)↔right supplementary motor area (R SMA)). Second, there appeared to be a left↔right shift in the regions active when comparing normal subjects to PD subjects, despite the fact that all subjects were using their right hand during the motor task. For example, while normal subjects recruited there R SMA and right thalamus (R THA), PD subjects recruited their L SMA and L THA. Presumably the connections between regions homologus (Figure 6.1) pro- vide a mechanism through which PD subjects can recruit regions on the opposite side of the brain. Third, there were connections between the right prefrontal cortex 138 6.4. fMRI Group Analysis in Parkinson’s Disease (R PFC) and right caudate (R CAU) in normal subjects that were missing in PD sub- jects. This likely reflects alterations in the secondary dopaminergic pathway to medial prefrontal regions known to be affected in PD [37]. Fourth, there was enhanced con- nectivity withing basal ganglia regions (e.g. right putamen (R PUT)↔right globus pallidus (R GLP), R THA→L PUT, right caudate (R CAU)→R GLP). This may reflect that these regions become entrained in oscillations which might enhance the functional connectivity observed with fMRI [49]. 139 Chapter 7 Conclusions and Discussions 7.1 Contribution Summary In this thesis, we have studied several issues on estimating high-dimensional sparse models from both theoretic and algorithmic perspectives. Now, we summarize the main contributions of this thesis. In Chapter 2, • We proposed a convex combined loss of `1 (LAD) and `2 (LS), rather than the pure LS cost function, coupled with the `1 penalty to produce a robust version of the Lasso. Asymptotic normality was established, and we showed that the variance of the asymptotic normal distribution is stabilized. Estimation consistency was proved at different shrinkage rates for {λn} and further proved by a non-asymptotic analysis for the noiseless case. • Under a simple adaptation procedure, we showed that the proposed robust Lasso is model selection consistent, i.e. the probability of the selected model to be the true model approaches to 1. • As an extension of the asymptotic analysis of the proposed robust Lasso, we studied an alternative robust version of the Lasso with the Huber loss function, the Huberized Lasso. For the Huberized Lasso, asymptotic normality and model selection consistency were established under much weaker conditions on the error distribution, i.e. no finite moment assumption is required for preserving similar asymptotic results as in the convex combined case. The Huberized 140 7.1. Contribution Summary Lasso estimator is well-behaved in the limiting situation when the error follows a Cauchy distribution, which has infinite first and second moments. • The analysis result obtained for the non-stochastic design was extended to the random design case with additional mild regularity assumptions. These as- sumptions are typically satisfied for auto-regressive models. In Chapter 3, • We proposed a hierarchical, fully Bayesian version of the Lasso model for infer- ring sparse linear regression from high-dimensional data sets. • We developed a reversible-jump MCMC algorithm to compute the unbiased minimum variance estimates and proved its convergence. In Chapter 4, • For high-dimensional covariance estimation problems where p/n → ∞, we showed that the MMSE shrinkage oracle estimator is inconsistent under both Frobenius and spectral risks for some typical covariance matrices. Moreover, we showed that the tapering estimator is uniformly superior than the MMSE shrinkage estimator in this case. • We proposed a STO estimator that combines the advantages from both the MMSE shrinkage and tapering estimators. In particular, the proposed estimator is suitable for estimating general, high-dimensional covariance matrices. An oracle estimator in the closed-form was derived and a practical algorithm to approximate the STO estimator was presented. In Chapter 5, • We identified a class of sparse matrices that is approximately inversely closed and proposed a conceptually simple and computationally efficient algorithm on estimation of large sparse precision matrices in this class. 141 7.1. Contribution Summary Figure 7.1: This overview summarizes the challenges raised in Chapter 1, the methods proposed in this thesis, and the relationship between proposed methods and the challenges being addressed. • Our asymptotic analysis showed that the proposed estimator for this class is statistically valid and optimal in the sense of minimax risk under the spectral norm. • We also established convergence result on the entry ∞-norm and showed that the proposed estimator is model selection consistent. In Chapter 6, • We presented a group robust Lasso (grpRLasso) framework that combines SEM and mAR(1) for inferring group-level, sparse brain connectivity networks. • The grpRLasso was applied to fMRI obtained from subjects with and with- out PD and significant group differences in biologically plausible regions were found. We suggest that the proposed method provides a computationally effi- cient means to infer group brain connectivity from fMRI data. The challenges C1-C5 raised in Chapter 1 have been individually addressed by the four topics studied in this thesis. The overview picture is shown in Figure 7.1. 142 7.2. Directions for Future Research 7.2 Directions for Future Research We would like to point out a few directions for the future research based on the accomplished work in this thesis. 7.2.1 Estimation of Conditional Independence for High-Dimensional Non-Stationary Time Series The independence assumption is of critical importance in all the approaches where the goal is to estimate a (series of) static and sparse precision matrix Ω’s. Unfortunately, this assumption is overly restrictive to a broad range of real-world datasets since the data generation mechanisms are hardly static. Equivalently speaking, the underlying Ω for generating {xi} can change over time and its structures and parameters at a particular time may also depend on the previous ones. For instances, functional brain connectivities are likely to alter when different tasks are performed; genetic regula- tory networks are prone to evolving in order to adapt to changing environments. To capture the time-dependent features, the iid hypothesis must be relaxed to accommo- date the reality. Nevertheless, there have only been a few research [78, 131] devoted to the estimation of high-dimensional precision matrices of multivariate time-series data. In the proposed future research, we shall focus on the estimation of high-dimensional sparse precision matrices Ω(t) for time-series data. We model the time-varying Ω(t) under the general physical dependence measure framework proposed by [125]: xi = G(ei, ei−1, · · · ; i/n), (7.1) where G is a measurable function driving the physical data generation processes and {ei} are independent innovations at t = i/n. There are two major advantages over the current state-of-the-art approaches based on the independence hypothesis. First, (7.1) can model non-linear, non-stationary multivariate time-series {xi}. In particular, 143 7.2. Directions for Future Research mild conditions on G allow us to model locally stationary processes, a very flexible and general stochastic framework that covers a wide range of many existing time- series models including linear processes (time-varying auto-regression and moving average processes), Volterra series, non-linear transforms of linear processes, and non- linear time series. Second, combining (7.1) with sparsity in Ω(t) unifies two forms of time-dependence in a single model: changing graph of Ω(t) and dependence through auto-correlation. Both are new to the current literature. Compared with the time-varying Gaussian graphical models in [78, 131], the ap- proach (7.1) significantly generalizes their work in several aspects. First, in [131], time-varying Ω(t) is modeled by independent structural changes, rather than the stochastic paradigm we consider here. In fact, their assumption can be seen as a spe- cial case in our paradigm in the sense that xi = Σ 1/2(i/n)ei. Second, our approach has clear regression interpretation on Ω(t), whereas in [131], estimation of Ω(t) is done by the 1-norm penalized Gaussian likelihood parameterized by kernel smoothed sample covariance matrix. Third, the smoothness conditions used in [131] are based on the maximal fluctuation of the first and second order derivatives on the entries of Σ(t) and Ω(t). As pointed out in [125], however, using derivatives is not a good way of dealing with time dependence because they may not even exist if G is not well-defined. In contrast, we leverage the physical dependence measure based on the coupling idea [125]. Based on the model (7.1), my research plan contains the following stages: 1. We consider the non-parametric estimation problem of the function G(·) (and hence Ω(t)) changing smoothly over time. In linear processes, this is tantamount to estimate the time-varying coefficients. We shall derive the asymptotics, in- cluding consistency, limiting distributions, and optimality, of the proposed es- timator. We will also consider the adaptive procedures on bandwidth selection. 2. We consider the case where there exist abrupt change-points (at unknown posi- tions) in G(·); this corresponds to the structural changes, e.g. adding or deleting 144 7.2. Directions for Future Research edges, in the graphical model. 3. We shall apply the theoretic results to estimate functional brain connectivity networks from functional Magnetic Resonance Imaging (fMRI) data, a natural and practically important application of the time-dependent model (7.1). The proposed method is novel in the sense that it extends the current linear struc- tural equation modeling (SEM) and multivariate auto-regression (mAR) models to locally stationary, time-varying models. 4. We shall also consider the application of the proposed method to model genetic regulatory networks using gene expression data. It is known that cell cycle is a dynamic system and gene expression levels are adaptive to external varying conditions. 7.2.2 Estimation of Eigen-Structures of High-Dimensional Covariance Matrices So far, we have seen that the estimation performances for large covariance and pre- cision matrices are measured by the accuracy of eigen-values. A more challenging problem (and largely remains unsolved) is to estimate the eigen-vector of such high- dimensional matrices. Precise estimation of both eigen-values and eign-vectors, for which we call the eigen-structures, is of tremendous importance in many areas of statistics (e.g. PCA and linear discriminant analysis), machine learning (e.g. face recognition and classification), signal processing (e.g. beamforming), and compu- tational biology (e.g. microarray clustering and genome-wide association study). Recently, it has been shown that eigen-vectors of the sample covariance matrix is gradually orthogonal to those of Σ in high-dimensional setups and therefore they essentially contain little information about the eigen-structure of Σ [73]. Without special structures in Σ, current approaches and theory can deal with the problem size comparable with the sample size, see [103] for the spiked covariance model as 145 7.2. Directions for Future Research an example; nonetheless, I am very interested in estimating the eigen-structure in regularized subclasses of Σ in the situations where p n; for instance, p grows at sub-exponential rate of n. Tools from random matrix theory and optimization are extremely useful for this purpose. 7.2.3 Multi-Task Lasso for Group Analysis The group analysis presented in [35] requires strictly grouping structures. 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In particular, p will be the number of the coefficients in the linear regression models and the number of variables in the covariance/precision matrix estimation problems. Vector and Matrix Bold lower letters, e.g. a,x, · · · , are used to denote vectors; capital letters, e.g. A,B,Σ,Ω, · · · , for matrices; and curly capital letters, e.g. G,S, · · · , for a collection of elements such as matrices. For a generic vector a, standard norm notations are used such as ‖a‖`1 := ‖a‖1 = ∑ j |aj|, ‖a‖`2 := ‖a‖2 = √∑ j |aj|2, ‖a‖`∞ := ‖a‖∞ = supj |aj|, and ‖a‖`0 := ‖a‖0 = ∑ j I(aj 6= 0). For a generic matrix A and 1 ≤ r ≤ ∞, the matrix Lr norm is defined as ‖A‖Lr = sup‖x‖r=1 ‖Ax‖r . (2) 161 A.1. Notations For a square matrix A, the matrix L1, L2, L∞, and Frobenius norms are defined as: ‖A‖L1 = sup j ∑ i |aij| (3) ‖A‖L2 = sup‖x‖2=1 √ xTATAx (4) ‖A‖L∞ = sup i ∑ j |aij| (5) ‖A‖F = √∑ i ∑ j a2ij, (6) respectively. One verifies that ‖A‖L2 ≤ ‖A‖L1 . Note that for a symmetric square matrix A, ‖A‖L1 = ‖A‖L∞ and ‖A‖L2 = supj {|λj(A)|} where λj(A) is the j-th eigen- value of A, i.e. ‖A‖L2 is the spectral norm of A (a.k.a. the operator norm of A as a linear functional from `2 → `2). Note that, in the rest of the paper, it is assumed AT = A unless otherwise indicated since we focus on considering covariance and precision matrices. Moreover, for simplicity, we shall skip the subscript in ‖·‖2 for the Euclidean norm of a vector and the spectral norm of a matrix. We will also use the entry-wise norm on matrices, which is tantamount to regard matrices as vectors. For instances, ‖A‖∞ stands for the maximum magnitude of the entries of A and ‖A‖1 = ∑ i ∑ j |aij|. The “dist” function defined on matrices for the distance between a point and a set, which coincides with the usual definition dist (A,S) = inf{‖A−B‖ : B ∈ S}. If S is a closed subset, the infimum is attained. Let xi be the i th-row of X and xj be the jth-column of a matrix X. Let Tr(M) denote the trace of a squared matrix M . M11,M12,M21, and M22 are submatrices of M partitioned according to M = M11 M12 M21 M22 . For a generic vector u, we interchangeably use uj and u[j] to denote the j th-element of u. Define supp (u) = {j ∈ {1, · · · , p}|uj 6= 0} and define |u| to be the cardinality 162 A.2. Proofs for Chapter 2 of u. Probability Zn P→ Z refers to convergence in probability and Zn ⇒ Z convergence in distribution. Note that we shall also use capital letters to mean random variables; this should not cause confusion with the matrix notation when the context is clear. We use the phase with high probability to mean that the referred probability approaches to 1 when n → ∞ (thus, p → ∞ as well). 1(A) is the indicator function of some measurable set A. For a random variable Z, E(Z;A) = ∫ A Z dP is the expectation of Z taken on A. For two probability measures P and Q with a common dominating measure µ, let p and q be the density of P and Q, respectively. Then the total variation affinity between P and Q is defined as ‖P ∧Q‖ = ∫ min(p, q) dµ. We use |A| to denote the size or cardinality of a set A. β̂n is said to be a √ n-consistent estimator for β if √ n(β̂n − β) has an unbiased limiting distribution with respect to (w.r.t.) 0. A.2 Proofs for Chapter 2 A.2.1 Proof of Theorem 2.2.1 Let Zn(u) = n −1∑n i=1 L(u; yi,xi)+n −1λn ‖u‖`1 and β̂n minimize Zn. Then it follows that √ n(β̂n − β) minimizes Vn, where Vn(u) = n∑ i=1 [ δ (( ei − u ∗xi√ n )2 − e2i ) + (1− δ) (∣∣∣∣ei − u∗xi√n ∣∣∣∣− |ei|) ] + λn p∑ j=1 [∣∣∣∣βj + uj√n ∣∣∣∣− |βj|] . (7) Without loss of generality (WLOG), by symmetry we can assume u ∗xi√ n ≥ 0. Put Yi = (1− δ)(1(ei < 0)− 1(ei ≥ 0))− 2δei 163 A.2. Proofs for Chapter 2 and Zi = 2× ( u∗xi√ n − ei ) 1 ( 0 ≤ ei < u ∗xi√ n ) . Then Vn(u) = δ n∑ i=1 ( u∗ xix ∗ i n u ) + 1√ n n∑ i=1 u∗xiYi + (1− δ) n∑ i=1 Zi +λn p∑ j=1 [∣∣∣∣βj + uj√n ∣∣∣∣− |βj|] . (8) With the error assumption, we have EYi = (1− δ)(P (ei < 0)− P (ei ≥ 0))− 2δEei = 0 and Var (Yi) = (1− δ)2E[1(ei < 0)− 1(ei ≥ 0)]2 + 4δ2Ee2i − 4δ(1− δ) [E(ei; ei < 0)− E(ei; ei ≥ 0)] = (1− δ)2 + 4δ2σ2 + 4δ(1− δ)E|ei|, since ei has a symmetric distribution. Note that σ 2 <∞ is used here. Now by assumption 1, we have n∑ i=1 ( u∗ xix ∗ i n u ) → u∗Cu, by Lemma A.2.5 n∑ i=1 Zi P→ f(0)u∗Cu and 1√ n n∑ i=1 u∗xiYi ⇒ u∗W 164 A.2. Proofs for Chapter 2 where W ∼ N(0, ((1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10)C), and λn p∑ j=1 [∣∣∣∣βj + uj√n ∣∣∣∣− |βj|]→ λ0 p∑ j=1 [ujsgn (βj) 1(βj 6= 0) + |uj|1(βj = 0)]. Combining all terms together and applying Slutsky’s lemma (c.f. p60, [110]), we deduce that Vn(u)⇒ V (u), where V (u) = δu∗Cu+ (1− δ)f(0)u∗Cu+ u∗W + λ0 p∑ j=1 [ujsgn (βj) 1(βj 6= 0) + |uj|1(βj = 0)]. (9) It is obvious that the finite-dimensional convergence holds. Finally, since Vn is con- vex and V has a unique minimum, the epi-convergence result from [60] implies that argmin(Vn) = √ n(β̂n − β)⇒ argmin(V ). A.2.2 Proof of Corollary 2.2.2 λ0 = 0 implies that V (u) = (δ + (1− δ)f(0))u∗Cu+ u∗W, which is minimized at argmin(V ) = − C −1W 2(δ + (1− δ)f(0)) ∼ N ( 0, (1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10 4(δ + (1− δ)f(0))2 C −1 ) . 165 A.2. Proofs for Chapter 2 A.2.3 Proof of Corollary 2.2.3 By the law of large numbers and the fact that β̂LS is a √ n-consistent estimator of β under assumptions 1 and 2(c), σ̂2 = 1 n− p n∑ i=1 ( ei − ( β̂LS − β )∗ xi )2 = n n− p [ 1 n n∑ i=1 e2i − 2 ( β̂LS − β )∗ 1 n n∑ i=1 eixi + ( β̂LS − β )∗ 1 n n∑ i=1 xix ∗ i ( β̂LS − β )] P→ σ2. (10) The corollary then follows from Theorem 2.2.1. A.2.4 Proof of Theorem 2.2.4 Define Zn(u) = n −1∑n i=1 L(u; yi,xi)+n −1λn ‖u‖`1 , where L(u; yi,xi) = δ (yi − u∗xi) 2+ (1− δ) |yi − u∗xi| . It suffices to show that 1. For any compact subset K ⊆ Rp, sup u∈K |Zn(u)− Z(u)| P→ 0. (11) 2. β̂n = Op(1). Put ri = yi − u∗xi. For part 1, consider Zn(u) = δ(u− β)∗ ( n∑ i=1 xix ∗ i n ) (u− β) + δ n n∑ i=1 e2i − 2δ n n∑ i=1 (u− β)∗xiei + (1− δ) 1 n n∑ i=1 |ri|+ λn n ‖u‖`1 . (12) 166 A.2. Proofs for Chapter 2 The first three terms are easily seen to converge in probability to δ[(u − β)∗C(u − β) + σ2]. For the fourth term, note that, ∑n i=1Er 2 i n2 → 0. By a weak law of large numbers (c.f. Theorem 1.14 in [110]), we have 1 n n∑ i=1 (|ri| − E|ri|) P→ 0. So it follows that 1 n n∑ i=1 |ri| P→ 1 n n∑ i=1 E|ri|, (13) provided that the limit of 1 n ∑n i=1E|ri| exists. Now, we show that the sequence { 1 n ∑n i=1E|ri|}n∈N is Cauchy. Consider∣∣∣∣∣ 1n+ 1 n+1∑ i=1 E|ri| − 1 n n∑ i=1 E|ri| ∣∣∣∣∣ = ∣∣∣∣∣ 1n+ 1 n+1∑ i=1 E|ri| − 1 n n+1∑ i=1 E|ri|+ 1 n n+1∑ i=1 E|ri| − 1 n n∑ i=1 E|ri| ∣∣∣∣∣ ≤ 1 n(n+ 1) n+1∑ i=1 E|ri|+ 1 n E|rn+1| ≤ 1 n2 n+1∑ i=1 [M10 + |(β − u)∗xi|] + 1 n E [|en+1|+ |(β − u)∗xn+1|] . Then it follows that, with probability 1, the last quantity converges to 0 by Lemma A.2.1 and A.2.2, as max1≤i≤n |xi| = o( √ n). Thus, the limit of n−1 ∑n i=1E|ri| exists, which is denote by r. Therefore we can conclude that Zn converges in probability to a function Z, where Z(u) = δ(u− β)∗C(u− β) + δσ2 + (1− δ)r + λ0 ‖u‖`1 . 167 A.2. Proofs for Chapter 2 Since {Zn}n∈N are convex, it follows from the convexity lemma [105] that Z is nec- essarily convex and the pointwise convergence in probability can be strengthened to the uniform convergence on compact sets. Part 1 is thus proved. For part 2, since Zn(u) ≥ δ n n∑ i=1 (yi − u∗xi)2, where the minimum of the RHS is bounded in probability, it follows that β̂n = Op(1). As a particular case that λn = o(n), we can see that lim inf n→∞ Zn(u) ≥ δ(u− β)∗C(u− β) + δσ2 + (1− δ)M10 (14) and lim sup n→∞ Zn(u) ≤ δ(u− β)∗C(u− β) + δσ2 + (1− δ)M10 + 2(1− δ) lim sup n→∞ 1 n n∑ i=1 |(u− β)∗xi|. (15) It is then clear that lim supn→∞ Zn(β) ≤ lim supn→∞ Zn(u). Since we have shown Zn P→ Z, it follows that Z(β) ≤ Z(u). I.e. β minimizes Z(u). The proof is complete. A.2.5 Proof of Proposition 2.2.5 Define Zn(u) = n∑ i=1 [δ(u− β)∗xix∗i (u− β) + (1− δ) |(u− β)∗xi|] + λn ‖β‖`1 168 A.2. Proofs for Chapter 2 and Vn(u) = n∑ i=1 [δu∗xix∗iu+ (1− δ) |u∗xi|] + λn p∑ j=1 [|βj + uj| − |βj|] = δnu∗Cnu+ (1− δ) n∑ i=1 |u∗xi|+ λn p∑ j=1 [|βj + uj| − |βj|] . Let β̂ 0 n be a minimizer of Zn(u) over u. Then h = β̂ 0 n − β minimizes Vn(u). Note that Vn(0) = 0, therefore Vn(h) ≤ 0. Since δnh∗Cnh+ (1− δ) n∑ i=1 |h∗xi| ≥ 0, we have ∑ j /∈A |hj|+ ∑ j∈A (|βj + hj| − |βj|) ≤ 0 such that ‖h‖`1(Ac) ≤ ∣∣∣∣∣∑ j∈A (|βj + hj| − |βj|) ∣∣∣∣∣ ≤∑ j∈A ||βj + hj| − |βj|| ≤ ∑ j∈A |hj| = ‖h‖`1(A) . Since ‖h‖`0(A) ≤ S, it follows that ‖h‖`1(A) ≤ √ S ‖h‖`2(A) ≤ √ S ‖h‖`2 , (16) whereas ‖h‖`1 = ‖h‖`1(A) + ‖h‖`1(Ac) ≤ 2 ‖h‖`1(A) ≤ 2 √ S ‖h‖`2 . (17) Now we can bound Vn(u) from below as Vn(u) ≥ δnu∗Cnu+ (1− δ) √ n (u∗Cnu) 1/2 + λn p∑ j=1 [|βj + uj| − |βj|] , (18) where we used ‖Xu‖`1 ≥ ‖Xu‖`2 = √ n (u∗Cnu) 1/2 . 169 A.2. Proofs for Chapter 2 Substituting h into (18), we get 0 ≥ Vn(h) = δnh∗Cnh+ (1− δ) √ n (h∗Cnh) 1/2 + λn p∑ j=1 [|βj + hj| − |βj|] . (19) So by (16) and (18), we obtain λn √ S ‖h‖`2 ≥ λn ‖h‖`1(A) ≥ δnh∗Cnh+ (1− δ) √ n (h∗Cnh) 1/2 . (20) Now, we bound h∗Cnh from below. WLOG, we can assume that h is in decreasing order of magnitudes. Let T1 be the S0-largest positions of h. Decompose h = h(T1)+ h(T c1 ), where h(T1) is the p × 1 vector that is a restricted version of h to the set T1 and 0 elsewhere. We note that ‖h(T c1 )‖2`2 ≤ p∑ j=S0+1 ‖h‖2`1 j2 ≤ ‖h‖2`1 p∑ j=S0+1 ( 1 j − 1 − 1 j ) ≤ ‖h‖ 2 `1 S0 ≤ 4S ‖h‖ 2 `2 S0 , where the third inequality follows from the telescope sum. So it follows that h(T1) ∗Cnh(T1) ≥ φmin(S0) ‖h(T1)‖2`2 = φmin(S0) (‖h‖2`2 − ‖h(T c1 )‖2`2) ≥ φmin(S0) ‖h‖2`2 ( 1− 4S S0 ) . Also, since ‖h‖`0(T c1 ) ≤ p− S0, we have h(T c1 ) ∗Cnh(T c1 ) ≤ φmax(p− S0) ‖h‖2`0(T c1 ) ≤ φmax(p− S0)4S S0 ‖h‖2`2 . 170 A.2. Proofs for Chapter 2 So applying Minkowski’s inequality, we conclude that h∗Cnh ≥ (h(T1)∗Cnh(T1)− h(T c1 )∗Cnh(T c1 ))2 ≥ ‖h‖2`2 ( 1− 4 √ Sφmax(p− S0) S0φmin(S0) ) , (21) where we used φmax(p − S0) ≥ φmin(S0). Set D0 = 1 − 4 √ Sφmax(p−S0) S0φmin(S0) > 0 where the positivity of D0 is a consequence of the inherent design. By inserting this estimate of h∗Cnh into (20), we get λn √ S ‖h‖`2 ≥ δnh∗Cnh+ (1− δ) √ n (h∗Cnh) 1/2 ≥ δnD0 ‖h‖2`2 + (1− δ) √ nD0 ‖h‖`2 . (22) Canceling ‖h‖`2 on both sides yields ‖h‖`2 ≤ λn √ S − (1− δ)√nD0 δnD0 . (23) This completes the proof. A.2.6 Proof of Theorem 2.3.1 As in the proof of Theorem 2.2.1, we let Zn(u) = n −1∑n i=1 L(u; yi,xi)+n −1λn ∑p j=1 ŵj |uj| and β̂n minimize Zn. Define Vn(u) = n∑ i=1 [ δ (( ei − u ∗xi√ n )2 − e2i ) + (1− δ) (∣∣∣∣ei − u∗xi√n ∣∣∣∣− |ei|) ] + λn p∑ j=1 ŵj [∣∣∣∣βj + uj√n ∣∣∣∣− |βj|] . (24) 171 A.2. Proofs for Chapter 2 Then √ n(β̂n − β) minimizes Vn. Rewrite Vn as Vn(u) = δ n∑ i=1 ( u∗ xix ∗ i n u ) + 1√ n n∑ i=1 u∗xiYi + (1− δ) n∑ i=1 Zi + λn√ n p∑ j=1 ŵj √ n [∣∣∣∣βj + uj√n ∣∣∣∣− |βj|] , (25) where Yi = (1− δ)(1(ei < 0)− 1(ei ≥ 0))− 2δei and Zi = 2× ( u∗xi√ n − ei ) 1 ( 0 ≤ ei < u ∗xi√ n ) . We have already seen that the first three terms together converge in distribution to (δ + (1− δ)f(0))u∗Cu+ u∗W, where W ∼ N(0, ((1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10)C). For the last term, we divide into two cases. If βj 6= 0, then ŵj P→ |βj|−γ by the continuous mapping theorem (CMT). So it follows that λn√ n︸︷︷︸ →0 × ŵj︸︷︷︸ P→|βj |−γ ×√n (∣∣∣∣βj + uj√n ∣∣∣∣− |βj|)︸ ︷︷ ︸ →ujsgn(βj) P→ 0. If βj = 0, then √ n (∣∣∣βj + uj√n∣∣∣− |βj|) = |uj| and λn√ n ŵj = λnn γ−1 2︸ ︷︷ ︸ →∞ ∣∣∣√nβ̂LS[j]∣∣∣−γ︸ ︷︷ ︸ =Op(1) P→∞. Applying the Slutsky’s lemma, we deduce that Vn ⇒ V pointwise, where V (u) = (δ + (1− δ)f(0))u ∗Cu+ u∗W if uj = 0∀j /∈ A, ∞ otherwise. (26) 172 A.2. Proofs for Chapter 2 Since Vn is convex and V has unique minimum, it follows from the standard epi- convergence results ([60] and [77]) that √ n(β̂n − β) = argmin(Vn) ⇒ argmin(V ). That is, √ n(β̂n − β)[A]⇒ C−111 WA and √ nβ̂n[A c]⇒ 0, where WA = N ( 0, (1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10 4[δ + (1− δ)f(0)]2 C11 ) . Then part 1 of the theorem follows and it is ready to see that argmin(Vn) converges in probability to 0 on Ac. For part 2, it suffices to show the following two cases: 1. For j ∈ A, the asymptotic normality proved in part 1) yeilds β̂n[j] P→ βj. Since |βj| > 0, it follows that P (j ∈ An)→ 1. 2. For j ∈ Ac, we want to show that P (j ∈ An) → 0. Observe that on the event {j ∈ An}, the first-order sub-differential optimality condition implies that ∣∣∣∣∣∣2δ xj ∗ ( y −Xβ̂n ) √ n + (1− δ) ∑ i∈B xijsgn ( yi − β̂∗nxi ) √ n − λnŵjsgn ( β̂n[j] ) √ n ∣∣∣∣∣∣ ≤ (1− δ) ∑ i/∈B |xij|√ n , (27) where B = {i ∈ {1, · · · , n} : yi−β̂∗nxi 6= 0}. Since the RHS of (27) is bounded in probability while the LHS diverges in probability, it follows that the probability with which (27) holds vanishes as n→∞. Hence, it follows that P (j ∈ An)→ 0. 173 A.2. Proofs for Chapter 2 A.2.7 Proof of Theorem 2.4.1 Let Zn(u) = 1 n ∑n i=1 L(u; yi,xi) + λn n ‖u‖`1 and β̂ H n minimize Zn. Anticipating 1/ √ n convergence rate so that we define Vn as Vn(u) = n∑ i=1 [( ei − u ∗xi√ n )2 1 (∣∣∣∣ei − u∗xi√n ∣∣∣∣ ≤ δ) − e2i 1 (|ei| ≤ δ) ] + n∑ i=1 [( 2δ ∣∣∣∣ei − u∗xi√n ∣∣∣∣− δ2) 1(∣∣∣∣ei − u∗xi√n ∣∣∣∣ > δ) − (2δ|ei| − δ2)1(|ei| > δ) ] + λn p∑ j=1 [∣∣∣∣βj + uj√n ∣∣∣∣− |βj|] . (28) Then √ n(β̂ H n −β) minimizes Vn. WLOG, by symmetry we can assume u ∗xi√ n ≥ 0. We can decompose Vn as Vn(u) = n∑ i=1 (Si + Ti + Yi + Zi) + λn p∑ j=1 [∣∣∣∣βj + uj√n ∣∣∣∣− |βj|] , (29) where Si = u∗xix∗iu n 1 (∣∣∣∣ei − u∗xi√n ∣∣∣∣ ≤ δ) , Ti = (|ei| − δ)2 ( 1 (∣∣∣∣ei − u∗xi√n ∣∣∣∣ ≤ δ)− 1 (|ei| ≤ δ)) , Yi = 2u∗xi√ n [ δ(1(ei < 0)− 1(ei ≥ 0))1 (∣∣∣∣ei − u∗xi√n ∣∣∣∣ > δ) −ei1 (∣∣∣∣ei − u∗xi√n ∣∣∣∣ ≤ δ)] , (30) 174 A.2. Proofs for Chapter 2 and Zi = 4δ ( u∗xi√ n − ei ) 1 ( 0 ≤ ei < u ∗xi√ n ) × 1 (∣∣∣∣ei − u∗xi√n ∣∣∣∣ > δ) . First, we observe that ESi = u∗xix∗iu n P (∣∣∣∣ei − u∗xi√n ∣∣∣∣ ≤ δ) . Since f is continuous, the continuity of the probability measure P implies that P (∣∣∣∣ei − u∗xi√n ∣∣∣∣ ≤ δ)→ K0δ. So it follows from Lemma A.2.2 that E ( n∑ i=1 Si ) → K0δu∗Cu. Since Si is a Bernoulli r.v., we have Var (Si) = ( u∗xix∗iu n )2 P (∣∣∣∣ei − u∗xi√n ∣∣∣∣ ≤ δ) × P (∣∣∣∣ei − u∗xi√n ∣∣∣∣ > δ) . So we deduce that n∑ i=1 Var (Si) ≤ n∑ i=1 u∗xix∗iu n ( u∗xi√ n )2 → 0, whereas Chebyshev’s inequality implies that n∑ i=1 Si P→ K0δu∗Cu. (31) 175 A.2. Proofs for Chapter 2 Next, we show that the second term Ti stochastically vanishes, i.e. n∑ i=1 Ti P→ 0. (32) To see this, it is straightforward to check that ETi = o ( n− 3 2 ) so E ∑n i=1 Ti = o ( n− 1 2 ) . Furthermore, by the symmetry of ei we can express ET 2 i as ET 2i = E ( (ei − δ)4 ; δ − u ∗xi√ n ≤ ei ≤ δ + u ∗xi√ n ) = 2 5 f(δ) ( u∗xi√ n )5 + o ( n− 5 2 ) such that Var (Ti) = 2 5 f(δ) ( u∗xi√ n )5 + o(n− 5 2 ), whereby Lemma A.2.1 implies that ∑n i=1Var (Ti) → 0. So it follows that ∑n i=1 Ti = op(1). The last two terms converge as in the proof of Theorem 2.2.1. Specifically, a routine calculation shows that EYi = 2u∗xi√ n [ δP ( δ − u ∗xi√ n < ei < δ + u∗xi√ n ) −E ( ei; δ − u ∗xi√ n < ei < δ + u∗xi√ n )] = 2u∗xi√ n [ 2δf(δ) u∗xi√ n + o ( n−1/2 )− 2δf(δ)u∗xi√ n ] = o ( n−1 ) , (33) 176 A.2. Proofs for Chapter 2 which implies that E ∑n i=1 Yi = o(1). Note that the two components of Yi are orthog- onal and it can be easily shown that Var ( n∑ i=1 Yi ) = n∑ i=1 [4u∗xix∗iu n ( δ2M0δ +K2δ + 2δf(δ) ( u∗xi√ n )2 + o(n−1/2) )] → (δ2M0δ +K2δ)C. (34) So the CLT implies that n∑ i=1 Yi ⇒ 2u∗W, (35) where W ∼ N (0, (δ2M0δ +K2δ)C). Finally, note that Zi = 0 for large n, so we can deduce that n∑ i=1 Zi P→ 0. (36) Combining (31), (32), (35), and (36) together and applying Slutsky’s lemma, we deduce that Vn(u)⇒ V (u) where V (u) = K0δu ∗Cu+ 2u∗W + λ0 p∑ j=1 [ujsgn (βj) 1(βj 6= 0) + |uj|1(βj = 0)]. Since Vn is convex and V has a unique minimum, it follows from [60] that argmin(Vn) = √ n(β̂ H n − β)⇒ argmin(V ). A.2.8 Proof of Theorem 2.5.1 The proof is essentially similar to that of Theorem 2.2.1, with additional complexity from the extra randomness ofX. We use the same notation as in the proof of Theorem 2.2.1 unless otherwise indicated. Let Vn be defined as in (8). Recall that Yi = (1− δ)(1(ei < 0)− 1(ei ≥ 0))− 2δei, 177 A.2. Proofs for Chapter 2 Zi = 2× ( u∗xi√ n − ei ) 1 ( 0 ≤ ei < u ∗xi√ n ) , and Vn(u) = δ n∑ i=1 ( u∗ xix ∗ i n u ) + 1√ n n∑ i=1 u∗xiYi + (1− δ) n∑ i=1 Zi + λn p∑ j=1 [∣∣∣∣βj + uj√n ∣∣∣∣− |βj|] . First, we observe that {u∗xiYi}i∈N forms a martingale difference sequence since E (u∗xiYi | Fi−1) = u∗xiE (Yi) = 0, where the second equality follows from the hypotheses that xi ∈ Fi−1 and σ(ei) is independent of Fi−1. Put Tn = ∑n i=1E ( (u∗xiYi) 2 | Fi−1 ) . A simple calculation shows that Tn n = ((1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10) × u∗ ( 1 n n∑ i=1 xix ∗ i ) u P→ ((1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10)u∗Cu. Moreover, consider n−1 n∑ i=1 E ( u∗xix∗iuY 2 i ) = E [ u∗ ( 1 n n∑ i=1 xix ∗ i ) uE ( Y 2i |Fi−1 )] = E ( Y 2i ) E [ u∗ ( 1 n n∑ i=1 xix ∗ i ) u ] → ((1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10)u∗Cu, where the last conclusion follows from the dominated convergence theorem. So it 178 A.2. Proofs for Chapter 2 follows that ∑n i=1 n −1 (u∗xiYi) 2 ∈ L1. By Lemma A.2.3, we have 1 (|u∗xiYi| > √n) = 1(∣∣∣∣u∗xiYi√n ∣∣∣∣ > ) ↓ 0, P -a.s. Again by the dominated convergence and Lemma A.2.1, we deduce that 1 n n∑ i=1 E ( (u∗xiYi)2; |u∗xiYi| > √ n )→ 0 as n→∞. Now the martingale central limit theorem (p414, [44]) implies that 1√ n n∑ i=1 u∗xiYi ⇒ N(0, ((1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10)u∗Cu). Secondly, the term involving Zi is similar to the non-stochastic design case. That is, we need to show that ∑n i=1 Zi P→ f(0)u∗Cu, then the rest of the proof proceeds as in Theorem 2.2.1. Observe that by Lemma A.2.5 and the measureability assumption on xi, ∑n i=1E (Zi|Fi−1) = f(0) ∑n i=1 u ∗ xix∗i n u P→ f(0)u∗Cu. So it follows from the Skorokhod representation (p83-84, [44]) that n∑ i=1 EZi = E ( n∑ i=1 E (Zi|Fi−1) ) → f(0)u∗Cu. Similarly, we can show that ∑n i=1Var (Zi)→ 0. This completes the proof. A.2.9 Proof of Auxiliary Lemmas The following two useful lemmas (Lemma A.2.1 and A.2.2) concerning the conver- gence of two real sequences will be used repeatedly in the proofs of the theorems. Lemma A.2.1. Let 0 ≤ c1 < c2 < · · · and 0 ≤ d1 < d2 < · · · with cn → ∞ and dn → ∞. Let {an}n∈N and {bn}n∈N be two real sequences such that an ≥ 0, 1 cn ∑n i=1 ai → a ≥ 0, and 1dn max1≤i≤n |bi| → 0. Then 1cndn ∑n i=1 aibi → 0. Proof. Fix an > 0. Since 1 dn max1≤i≤n |bi| → 0, there is an N = N() such that 179 A.2. Proofs for Chapter 2 n ≥ N implies that |bi| dn ≤ for all i ∈ {1, · · · , n}. But then ∣∣∣∣∣ 1cndn n∑ i=1 aibi ∣∣∣∣∣ ≤ 1cndn N−1∑ i=1 |aibi|+ n∑ i=N ∣∣∣∣aicn ∣∣∣∣ ∣∣∣∣ bidn ∣∣∣∣ ≤ + ( n∑ i=1 ai cn ) ≤ + (a+ ) = (a+ 1)+ 2 for sufficiently large n. Since is arbitrary, it follows that 1 cndn ∑n i=1 aibi → 0 as n→∞. Lemma A.2.2. Let 0 ≤ c1 < c2 < · · · with cn → ∞. Let {an}n∈N and {bn}n∈N be two real sequences such that an ≥ 0, 1cn ∑n i=1 ai → a ≥ 0 and bn → b. Then 1 cn ∑n i=1 aibi → ab. Proof. Fix an > 0. Since bn → b, there is an N = N() such that n ≥ N implies that |bn − b| ≤ . Consider ∣∣∣∣∣ 1cn n∑ i=1 aibi − ab ∣∣∣∣∣ = ∣∣∣∣∣ n∑ i=1 ai cn bi − n∑ i=1 ai cn b+ n∑ i=1 ai cn b− ab ∣∣∣∣∣ ≤ n∑ i=1 ∣∣∣∣aicn (bi − b) ∣∣∣∣+ ∣∣∣∣∣ n∑ i=1 ai cn − a ∣∣∣∣∣ b ≤ N−1∑ i=1 ∣∣∣∣aicn (bi − b) ∣∣∣∣+ n∑ i=N ai cn |bi − b|+ b ≤ + ( n∑ i=1 ai cn ) + b ≤ + (a+ )+ b = (a+ b+ 1)+ 2 for sufficiently large n. Since is arbitrary, the lemma follows. 180 A.2. Proofs for Chapter 2 Lemma A.2.3. Under assumption 1, we have 1 n max 1≤i≤n x∗ixi → 0 and max 1≤i≤n x∗i ( n∑ i=1 x∗ixi )−1 xi → 0 as n→∞. Proof. Let > 0 and T = tr (C). Since Cn → C and C is positive definite, we have tr (Cn) = 1 n ∑n i=1 x ∗ ixi → T > 0. So there is an N = N() such that∣∣∣∣∣ 1n n∑ i=1 x∗ixi − T ∣∣∣∣∣ ≤ for all n ≥ N . But then 0 ≤ 1 n max N≤i≤n x∗ixi = 1 n max N≤i≤n ( i∑ j=1 x∗jxj − i−1∑ j=1 x∗jxj ) ≤ 1 n max N≤i≤n ((T + )i− (T − )(i− 1)) ≤ max N≤i≤n ( T + 2i n ) ≤ 3 for sufficiently large n. Letting n→∞, the first part of the lemma follows since 0 ≤ lim sup n→∞ 1 n max 1≤i≤n x∗ixi ≤ 3, whereas > 0 is arbitrary. The second claim is an easy consequence of the first part by the finite-dimensionality of xi’s. Lemma A.2.4. Suppose X1, X2, · · · is a sequence of i.i.d. r.v.’s with a continuous, positive probability density function f around 0 (w.r.t. the Lebesgue measure). Let F be the common cumulative distribution function with F (0) = 1 2 . Let gn(u) = 181 A.2. Proofs for Chapter 2 ∑n i=1 (∣∣∣Xi − u√n∣∣∣− |Xi|). Then gn(u) ⇒ g(u) where g(u) = uZ + f(0)u2 with Z ∼ N(0, 1). Proof. First assume u ≥ 0 and rewrite gn as following gn(u) = u√ n n∑ i=1 Yi + n∑ i=1 Zi, (37) where Yi = 1(Xi < 0)− 1(Xi ≥ 0) and Zi = 2× ( u√ n −Xi ) 1 ( 0 ≤ Xi < u√ n ) . Since F (0) = 1/2, i.e. Xi has median 0, we have EYi = 0 and Var (Yi) = EY 2 i = 1. Hence the central limit theorem (CLT) implies that ∑n i=1 Yi√ n ⇒ N(0, 1). Now consider the second term concerning Zi. First observe that for |u′| small, by the continuity assumption of f about 0, we have P (0 ≤ Xi < u′) = ∫ u′ 0 f(xi) dxi = ∫ u′ 0 (f(0) + o(1)) dxi = f(0)u′ + o(u′), E(Xi; 0 ≤ Xi < u′) = ∫ u′ 0 xi(f(0) + o(1)) dxi = 1 2 f(0)u′2 + o(u′2), E(X2i ; 0 ≤ Xi < u′) = ∫ u′ 0 x2i (f(0) + o(1)) dxi = 1 3 f(0)u′3 + o(u′3). 182 A.2. Proofs for Chapter 2 Applying u′ = u/ √ n, we deduce that EZi = E ( 2 ( u√ n −Xi ) ; 0 ≤ Xi < u√ n ) = 2u√ n P ( 0 ≤ Xi < u√ n ) − 2E ( Xi; 0 ≤ Xi < u√ n ) = 2u√ n ( f(0) u√ n + o ( u√ n )) − f(0)u 2 n + o (( u√ n )2) = f(0) n u2 + o ( u2 n ) . (38) So ∑n i=1EZi = f(0)u 2 + o(1). It can be similarly shown that Var (Zi) = 4f(0) 3 ( u√ n )3 + o (( u√ n )3) so that ∑n i=1Var (Zi) = O(n −1/2). So it follows from Chebyshev’s inequality that P (∣∣∣∣∣ n∑ i=1 Zi − n∑ i=1 EZi ∣∣∣∣∣ > ) ≤ ∑n i=1Var (Zi) 2 → 0, i.e. n∑ i=1 Zi P→ f(0)u2. Invoking Slutsky’s lemma, we obtain gn ⇒ g pointwise. The case for u < 0 follows the similar lines. Lemma A.2.5. Let X1, X2, · · · be an i.i.d. sequence of r.v.’s with a common con- tinuous, positive p.d.f. f around 0 and the median of Xi equal 0 for all i. Let u,vi ∈ Rp and gn(u) = ∑n i=1 (∣∣∣Xi − u∗vi√n ∣∣∣− |Xi|). Then gn(u) ⇒ g(u) where g(u) = u∗W + f(0)u∗Cu with W ∼ N(0, C) and C = limn→∞ 1n ∑n i=1 viv ∗ i . Proof. The proof is an easy consequence of Lemma A.2.4 and we use the same notation unless explicitly indicated. Without loss of generality, we can assume that u∗vi ≥ 0 for all i by symmetry. As in the proof of Lemma A.2.4, we have EZi = f(0) n (u∗viv∗iu)+ 183 A.3. Appendix for Chapter 3 o ( u∗viv∗i u n ) . So n∑ i=1 EZi = f(0)u ∗ ∑n i=1 viv ∗ i n u+ o(1)→ f(0)u∗Cu. Moreover, observe that n∑ i=1 Var (Zi) = 4f(0) 3 ( n∑ i=1 u∗viv∗iu n u∗vi√ n ) + o( 1√ n ). Applying Lemma A.2.1 with ai = u ∗viv∗iu, bi = u ∗vi, cn = n, and dn = √ n, it then follows from Lemma A.2.3 that ∑n i=1Var (Zi) → 0 as n → ∞. Then Chebyshev’s inequality implies that ∑n i=1 Zi P→ f(0)u∗Cu. Let Ỹi = (u∗vi)Yi. The CLT and Slutsky’s lemma together imply that ∑n i=1 Ỹi√ n ⇒ N(0,u∗Cu) = u∗W, where W ∼ N(0, C). The conclusion follows by invoking Slutsky’s lemma once again. A.3 Appendix for Chapter 3 A.3.1 Derivation of the Joint Posterior Distribution in (3.3) Denote the likelihood of the model by L(y|X, ·). We can factorize the joint posterior distribution according to the conditional independence relationships encoded in the hierarchical model as: p(β, τ, σ2,γ, λ|y, X) 184 A.3. Appendix for Chapter 3 ∝ p(β, τ, σ2,γ, λ)× L(y|X,β, τ, σ2,γ, λ) ∝ 1( p k )p(λ)p(τ, σ2)p(k|λ)∏ j∈γ p(βj|γ, τ)× (σ2)− n 2 exp ( −||y −Xβ|| 2 2 2σ2 ) ∝ 1( p k )(λτσ)−1 e−λλk k! τ−k exp ( − ∑ j∈γ |βj| τ ) × σ−n exp ( −||y −Xβ|| 2 2 2σ2 ) , which gives (3.3). A.3.2 MCMC Algorithm for the Binomial-Gaussian Model We apply the proposed fully Bayesian framework to the Binomial-Gaussian model proposed in [59]. More specifically, let γ be a p-length binary vector representing an active set. The active set is assumed to be binomially distributed p(γ|w) = wqγ (1− w)p−qγ (39) where w is the probability of taking value one and qγ means the number of ones in γ. Conditioning on γ, Zellner’s g-prior is used for the coefficient: p(βγ |γ, g) = N ( 0, gσ2(XTγXγ) −1 ) . (40) Then, assigning a conjugate beta prior on w and Jeffrey’s non-informative priors on σ2 and g, we have p(w) = Beta(a, b), p(σ2, g) ∝ (σ2g)−1. 185 A.3. Appendix for Chapter 3 Therefore we can analytically integrate out βγ and obtain p(y, g, σ2, w,γ) ∝ p(w)p(g)p(σ2)p(γ|w)p(y|g, σ2,γ) ∝ wa−1(1− w)b−1(σ2)−1g−1wqγ (1− w)p−qγ × (2piσ2)−n/2(1 + g)−qγ/2 exp ( −y Ty 2σ2 + g 2σ2(1 + g) Rγ ) , (41) where Rγ = β̂ T γX T γXγβ̂γ and β̂γ is least squares estimate of model γ. By further integrating out σ2 and w, we obtain p(y, g,γ) ∝ B(a+ qγ , b+ p− qγ)g−1(1 + g)−qγ/2 ( −y Ty 2 + g 2(1 + g) Rγ )−n/2 (42) where B(a+ qγ , b+ p− qγ) = Γ(a+qγ )Γ(b+p−qγ ) Γ(a+b−p) . Now, it follows that the full conditional probabilities p(γ|y, g) ∝ B(a+ qγ , b+ p− qγ)(1 + g)−qγ/2 ( −y Ty 2 + g 2(1 + g) Rγ )−n/2 , (43) p(g|y,γ) ∝ g−1(1 + g)−qγ/2 ( −y Ty 2 + g 2(1 + g) Rγ )−n/2 I(g ≥ 0). (44) Thus the updating probability from γ → γ ′ is the MH ratio min { p(γ ′|y, g) p(γ|y, g) × p(γ ′ → γ) p(γ → γ ′) , 1 } . (45) We note that the above BG-MCMC algorithm can be interpreted as an RJ-MCMC approach with the birth-and-death proposal. Here since the Jacobian for the birth- and-death proposal is 1 for model jumping, there is no explicit Jacobian term reflected in the generalized MH ratio and thus it is coincides with the ordinary MH ratio with γ being considered as a regular parameter. 186 A.3. Appendix for Chapter 3 Regarding the parameter g, we employ a symmetric random walk sampler for updating. More specifically, we let g′ = g + u where u ∼ N(0, 2) and accept with probability min { p(g′|y,γ) p(g|y,γ) , 1 } . The MCMC algorithm for the BG model is summarized in Algorithm 4. Algorithm 4: BG-MCMC algorithm Input: The number of iterations T . Random walk step size . Data: X and y. Output: {{ γ(t), g(t) } ∣∣t ∈ {0, · · · , T}}. begin1 Initialization: choose { γ(0), g(0) } and t = 1.2 repeat3 if k(t−1) = 1 then4 k(t) ← k(t−1) + 1.5 else if k(t−1) = p then6 k(t) ← k(t−1) − 1.7 else8 k(t) ← k(t−1) + U({−1, 1}).9 end10 if k(t) = k(t−1) + 1 then11 Propose a γ ′ such that |γ ′| = |γ|+ 1 (birth move);12 Accept γ ′ with the probability given by MH ratio.13 else14 Propose a γ ′ such that |γ ′| = |γ| − 1 (death move);15 Accept γ ′ with the probability given by MH ratio.16 end17 Sample g with the Gaussian random walk with step size . t← t+ 1.18 until t = T.19 end20 187 A.4. Proofs for Chapter 4 A.4 Proofs for Chapter 4 A.4.1 Proof of Theorem 4.2.2 By definition, LHS of (4.11) can be written as E‖Σ̂o − Σ‖2F = E{Tr[(Σ̂o − Σ)2]} = Tr(EΣ̂2o)− Tr[(EΣ̂o)Σ]− Tr[Σ(EΣ̂o)] + Tr(Σ2) = Tr(EΣ̂2o)− 2 [ (1− ρo)Tr(Σ2) + ρoTr 2(Σ) p ] + Tr(Σ2). We analyze the first term in the last equation and find that it equals I = (1− ρo)2E[Tr(Ŝ2)] + p−1ρ2oE[Tr2(Ŝ)] + 2p−1(1− ρo)ρoE[Tr2(Ŝ)]. Let A = E [ Tr(Ŝ2)− Tr(Ŝ) p ] and B = Tr(Σ2)− Tr 2(Σ) p . Note that for any square matrix M , we have Tr(M2) ≥ Tr2(M)/p; therefore A ≥ 0 and B ≥ 0. Expanding further by substituting ρ̂o in (4.8) into I, we can obtain that I = B(1− B A ) + E [ Tr2(Ŝ) p − Tr 2(Σ) p ] . Now, by the Gaussian assumption, we have ETr(Ŝ2) = n+ 1 n Tr(Σ2) + 1 n Tr2(Σ) ETr2(Ŝ) = Tr2(Σ) + 2 n Tr(Σ2). Plugging this expression into I, we see the theorem follows. 188 A.4. Proofs for Chapter 4 A.4.2 Proof of Theorem 4.2.3 By the definition of the matrix spectral norm and noting that EŜ = Σ and ETr(Ŝ) = Tr(Σ) = p, we can obtain the following chain inequalities E‖Σ̂o − Σ‖2 = E ( sup ‖x‖2=1 xT (Σ̂o − Σ)x )2 ≥ ( E sup ‖x‖2=1 xT (Σ̂o − Σ)x )2 ≥ sup ‖x‖2=1 (ExT (Σ̂o − Σ)x)2 = sup ‖x‖2=1 ( xT (EΣ̂o − Σ)x )2 = sup ‖x‖2=1 ( xT ((1− ρo)EŜ + ρoEF̂ − Σ)x )2 = sup ‖x‖2=1 ( xT (p−1ρoTr(Σ)I − ρoΣ)x )2 = ( ρo − ρo inf‖x‖2=1 xTΣx )2 = ρ2o(1− λmin(Σ))2. Here we used Jensen’s inequality twice at the second and third steps. A.4.3 Proof of Theorem 4.3.1 Let ρ ∈ [0, 1] and note that ETr(Σ̂Σ) = (1− ρ)Tr(Σ2) + ρTr((W ◦ Σ)Σ) = (1− ρ)‖Σ‖2F + ρ‖V ◦ Σ‖2F , (46) where V = (vij) with vij = √ wij. Moreover, since Tr(Σ̂2) = (1− ρ)2Tr(Ŝ2) + 2ρ(1− ρ)Tr((W ◦ Ŝ)Ŝ) + ρ2Tr((W ◦ Ŝ)2), taking expectation on both sides, we can obtain that ETr(Σ̂2) = (1− ρ)2E‖Ŝ‖2F + 2ρ(1− ρ)E‖V ◦ Ŝ‖2F + ρ2E‖W ◦ Ŝ‖2F . (47) 189 A.4. Proofs for Chapter 4 To calculate ρ̂STO, we need to find the minimizer of E[Tr(Σ̂2)] − 2Tr((EΣ̂)Σ). Ex- panding this using (46) and (47), we get E [ Tr(Σ̂2)− 2Tr((EΣ̂)Σ) ] = (1− ρ)2E‖Ŝ‖2F + 2ρ(1− ρ)E‖V ◦ Ŝ‖2F + ρ2E‖W ◦ Ŝ‖2F − 2(1− ρ)‖Σ‖2F − 2ρ‖V ◦ Σ‖2F . Differentiating this function w.r.t. ρ and finding its solution to zero, we immediately get (4.20). Further, if it is assumed that {xi} follow i.i.d. N(0,Σ), by Wick’s theorem, we have Eŝ2ij = 1 n2 E (∑ k=1 xkixkj )2 = 1 n2 E ( n∑ k=1 x2kix 2 kj + ∑ 1≤k 6=k′≤n xkixkjxk′ixk′j ) = 1 n Ex2kix 2 kj + 1 n2 ∑ 1≤k 6=k′≤n E(xkixkj)E(xk′ixk′j) = 1 n ( σiiσjj + 2σ 2 ij ) + n(n− 1) n2 σ2ij = σiiσjj n + n+ 1 n σ2ij. Now, it follows from direct calculation that E‖V ◦ Ŝ‖2F = E p∑ i=1 p∑ j=1 wij ŝ 2 ij = p∑ i=1 p∑ j=1 wijEŝ 2 ij = p∑ i=1 p∑ j=1 wij ( n+ 1 n σ2ij + 1 n σiiσjj ) = n+ 1 n ‖V ◦ Σ‖2F + 1 n Tr(DV 2D), where D = diag(Σ); so similarly E‖W ◦ Ŝ‖2F = n+ 1 n ‖W ◦ Σ‖2F + 1 n Tr(DW 2D). 190 A.5. Proofs for Chapter 5 Substituting into (4.20), we have ρ̂STO = ‖Σ‖2F + Tr2(Σ)− ‖V ◦ Σ‖2F − Tr(DV 2D) (n+ 1)(‖Σ‖2F + ‖W ◦ Σ‖2F − 2‖V ◦ Σ‖2F ) + Tr2(Σ) + Tr(DW 2D)− 2Tr(DV 2D) . (48) If ρ̂STO /∈ [0, 1], we modify ρ̂STO to be either 0 or 1, whichever gives a smaller MSE since it is a quadratic function of ρ and therefore attains the minimum value at one of its boundary points. A.5 Proofs for Chapter 5 A.5.1 Proof of Lemma 5.2.1 Suppose A ∈ Gq(cn,p), for ∀j, we have |a(i)j|q ≤ cn,p i , (49) where (i) is the decreasingly ordered i-th index (in magnitude) of aj, the j-th column of A. Fix an ε > 0 and choose B ∈ Sk such that bij = aij, for |(i)| ≤ k.0, otherwise. (50) Then, it follows that ‖A−B‖ ≤ ‖A−B‖L1 = sup j ∑ i |aij − bij| = sup j ∑ i>k |a(i)j| ≤ sup j ∑ i>k (cn,p i )1/q = c1/qn,p sup j ∑ i>k i− 1 q ≤ q 1− q c 1 q n,pk 1− 1 q ≤ ε (51) for k being large enough, since −q−1 < −1. For part 2), we will use the following Lemma: 191 A.5. Proofs for Chapter 5 Lemma A.5.1. For each k ∈ N, 0 ≤ q < 1, and x1 ≥ x2 · · · ≥ xk ≥ 0, the following holds ( k∑ i=1 i 1 q −1xi )q ≤ k∑ i=1 xqi . (52) Proof. Without loss of generality, let us prove an equivalent statement with yi := x q i , q′ := 1 q , k∑ i=1 iq ′−1yq ′ i ≤ ( k∑ i=1 yi )q′ . This follows from ( k∑ i=1 yi )q′ = ( k−1∑ i=1 yi + yk )q′ = ( k−1∑ i=1 yi )q′( 1 + yk∑k−1 i=1 yi )q′ ≥ ( k−1∑ i=1 yi )q′[ 1 + ( 1 + ∑k−1 i=1 yi yk )q′−1( yk∑k−1 i=1 yi )q′] . At the last line above, for q′ > 1, we use (1 + x)q ′ ≥ 1 + (1 + 1 x )q ′−1xq ′ , (53) which can be shown by integrating for 0 ≤ t ≤ x, q′(1 + t)q ′−1 ≥ q′tq′−1(1 + 1/x)q′−1. Therefore, ( k∑ i=1 yi )q′ ≥ ( k−1∑ i=1 yi )q′ + ( 1+ ∑k−1 i=1 yi yk )q′−1 yq ′ k ≥ ( k−1∑ i=1 yi )q′ +kq ′−1yq ′ k (since yi yk ≥ 1). The desired inequality follows from induction on k. Now consider two matrices A = (aij)p×p ∈ Gq(cn,p), A′ = ( a′ij ) p×p ∈ Gq(c′n,p). Fix 192 A.5. Proofs for Chapter 5 1 ≤ j ≤ p. Then, for AA′ = (a′′ij)p×p, ∑ i ∣∣a′′ij∣∣q ≤∑ i (∑ ` |ai`a′`j| )q ≤ ∑ i (∑ ` |aiσ[`]a′(`)j| )q , where (`) is the ordered index as in (49), and σ is a permutation of {1, 2, · · · , p} due to the reordering. Thus, ∑ i ∣∣a′′ij∣∣q ≤∑ i [∑ ` ( c′n,p ` ) 1 q |aiσ[`]| ]q = c′n,p ∑ i (∑ ` `− 1 q |aiσ[`]| )q . Reorder `− 1 q aiσ[`]’s according to their magnitude. Denote this new sequence by z(`): z(1) ≥ z(2) ≥ z(3) ≥ · · · ≥ z(p). Notice that since 1q − 1 > 0, z(`) ` 1 q −1 ≥ z(`+1) (`+ 1) 1 q −1 , ` = 1, 2, · · · , p− 1. Therefore, we can use (52) to see ∑ i (∑ ` `− 1 q |aiσ[`]| )q = ∑ i (∑ ` z(`) )q = ∑ i (∑ ` ` 1 q −1 z(`) ` 1 q −1 )q ≤ ∑ i ∑ ` zq(`) `1−q = ∑ i ∑ ` (σ′[`])q−1`−1|aiσ[`]|q for some permutation σ′. By interchanging the summations and using Hölder’s in- equality, this last term is bounded by cn,p ∑ ` (σ′[`])q−1`−1 ≤ cn,p (∑ ` (σ′[`])−2 ) 1−q 2 (∑ ` `− 2 1+q ) 1+q 2 = cn,p (∑ ` `−2 ) 1−q 2 (∑ ` `− 2 1+q ) 1+q 2 ≤ cn,pC(q) for some constant C(q) depending only on q; here, C(q) < ∞ follows from q < 1. 193 A.5. Proofs for Chapter 5 Thus, we have the estimate ∑ i |a′′ij|q ≤ C(q)cn,pc′n,p. (54) Then, we deduce by induction that for Ar = (a (r) ij ), sup j ∑ i ∣∣∣a(r)ij ∣∣∣q ≤ C(q)rcrn,p. (55) Equivalently, we have Ar ∈ Gq(C(q)rcrn,p) if A ∈ Gq(cn,p). Hence, we can construct B′ ∈ Sk′ as in (50) such that b′ij = a(r)ij for |(i)| ≤ k′ and b′ij = 0 otherwise, where k′ is defined in (5.17). The proof is then complete because‖Ar −B′‖ ≤ ε for all k′ ≥ [ q (1− q)εC(q) r q c r q n,p ] q 1−q = kminC(q) r 1−q c r−1 1−q n,p . A.5.2 Proof of Theorem 5.2.2 Suppose Σ ∈ F(q,m) and let Ω = Σ−1. Then m ≤ λmin(Σ) ≤ λmax(Σ) ≤ m−1. Since λmin(Σ) = λ −1 max(Ω) and λmax(Σ) = λ −1 min(Ω), we have m ≤ λmin(Ω) ≤ λmax(Ω) ≤ m−1. (56) Thus, Ω ∈ U(m). Write Ω in terms of its Neumann series as Ω = Σ−1 = Br +Rr, where Br is defined in (5.10) and Rr = η ∞∑ j=r+1 (I − ηΣ)j. (57) 194 A.5. Proofs for Chapter 5 It then follows from (5.9) that, for any ε > 0, there is an r0 = r0(m, ε) ∈ N such that ‖Rr‖ ≤ ε for all r ≥ r0. Also, we have by assumption that Σ ∈ Gq(cn,p) for some cn,p ≥ 0 which in turn implies by Lemma 5.2.1 that, for any j ≥ 1, (I − ηΣ)j ∈ Gq ((1 + ηcn,p)jC(q)j). Then it is obtained (e.g. using (52)) that Br0 ∈ Gq (ηqr0C(q)r0(1 + ηcn,p)r0) ⊂ Gq. (58) The second claim of the theorem follows from (5.17). A.5.3 Proof of Proposition 5.3.1 By definition, σ?jk = n −1∑n i=1 xijxik. We begin with P (|σ?jk − σjk| ≥ t) = P ( 1 n ∣∣∣∣∣ n∑ i=1 xijxik − σjk ∣∣∣∣∣ ≥ t ) = P ( 1 n ∣∣∣∣∣ n∑ i=1 x̃ijx̃ik − ρjk ∣∣∣∣∣ ≥ t√σjjσkk ) , (59) where x̃ij = xij/ √ σjj (similar definition for x̃ik) and ρjk = σjk/ √ σjjσkk. Because x̃ij x̃ik i.i.d.∼ N 0 0 , 1 ρjk ρkj 1 , we deduce that x̃ij + x̃ik i.i.d.∼ N(0, 2(1 + ρjk)), x̃ij − x̃ik i.i.d.∼ N(0, 2(1− ρjk)). Therefore, from the polarization identity ∑ i (x̃ijx̃ik − ρij) = 1 4 ∑ i [ (x̃ij + x̃ik) 2 − 2(1 + ρjk)− (x̃ij − x̃ik)2 + 2(1− ρjk) ] , 195 A.5. Proofs for Chapter 5 the expression (59) can be bounded by ≤P ( 1 4n ∣∣∣∣∣∑ i (x̃ij + x̃ik) 2 − 2(1 + ρjk) ∣∣∣∣∣ ≥ t2√σjjσkk ) + P ( 1 4n ∣∣∣∣∣∑ i (x̃ij − x̃ik)2 − 2(1− ρjk) ∣∣∣∣∣ ≥ t2√σjjσkk ) ≤2P ( 1 n ∣∣∣∣∣∑ i ( V 2i − 1 )∣∣∣∣∣ ≥ t(1 + |ρjk|)√σjjσkk ) , where Vi are i.i.d. N(0, 1). Note that (1+ |ρjk|)√σjjσkk ≤ 2m−1 since Σ ∈ U(m) and V 2i are χ 2(1) with exponential tail, it follows from [120, Corollary 17, page 16] that there are constants C4, C5 > 0, depending only on C1, C2 and m, such that P ( 1 n ∣∣∣∣∣∑ i ( V 2i − 1 )∣∣∣∣∣ ≥ t(1 + |ρjk|)√σjjσkk ) ≤ P ( 1 n ∣∣∣∣∣∑ i ( V 2i − 1 )∣∣∣∣∣ ≥ mt2 ) ≤ 2 exp [ −C4min ( t2 C25 , t C5 ) n ] . The proposition is therefore proved. A.5.4 Proof of Theorem 5.3.3 By choosing ε = O(cn,p(log p/n) (1−q)/2) in the forth step of our algorithm, it suffices to show ∥∥∥Ω̃− Ω∥∥∥ obeys the upper bound (5.24). By (5.23) and in [11, Theorem 1], ∥∥∥Σ̃n − Σ∥∥∥ ≤ Ccn,p( log p n )(1−q)/2 (60) 196 A.5. Proofs for Chapter 5 with probability greater than (1− C6p−8τ2/C24+2) that approaches to 1 whenever τ > C4/2. With such a high probability, ∥∥∥Ω̃− Ω∥∥∥ = η ∥∥∥∥∥ r∑ j=0 (I − ηΣ̃n)j − ∞∑ j=0 (I − ηΣ)j ∥∥∥∥∥ ≤η ∥∥∥∥∥ r∑ j=0 [ (I − ηΣ̃n)j − (I − ηΣ)j ]∥∥∥∥∥+ η ∞∑ j=r+1 ∥∥(I − ηΣ)j∥∥ ≤η2 r∑ j=1 C(j, τ) ∥∥∥Σ̃n − Σ∥∥∥+ 1 m ( 1−m2 1 +m2 )r+1 ≤C(q,m, τ) ∥∥∥Σ̃n − Σ∥∥∥+ 1 m ( 1−m2 1 +m2 )r+1 , (61) where the last two inequalities follow from Lemma A.5.2 (see below) and ‖I − ηΣ‖ = (1−m2)/(1 +m2) ≤ δ < 1. Note that cn,p(log p/n)(1−q)/2 → 0 as n→∞, thus from (60) the theorem follows. The following lemma says that the matrix power operation Ar is a contraction mapping for ‖A‖ uniformly bounded up by 1 on appropriate subsets. Lemma A.5.2. Fix a δ ∈ (0, 1) and ε > 0. Let A and B be any two square matrices such that ‖B‖ ≤ δ and ‖A−B‖ ≤ ε. Then for all r ∈ N, there exists a constant, C(r, δ + ε), depending only on r and δ + ε such that ‖Ar −Br‖ ≤ C(r, δ + ε) ‖A−B‖ . (62) Moreover, if δ+ ε < 1, then the sequence {C(r, δ+ ε)}∞r=1 is summable; in particular, ∞∑ r=1 ‖Ar −Br‖ . ‖A−B‖ , (63) where the constant here depends on (δ + ε). Proof. The proof is standard and is based on induction. By assumptions, ‖A‖ ≤ δ + ε < 1. For r = 1, select C(1, δ + ε) = 1. Suppose (62) holds for r. Observe the 197 A.5. Proofs for Chapter 5 following identity Ar+1 −Br+1 = A(Ar −Br) + (A−B)Br, (64) which in turn implies that ∥∥Ar+1 −Br+1∥∥ ≤‖A‖ ‖Ar −Br‖+ ‖A−B‖ ‖B‖r ≤(δ + ε)C(r, δ + ε) ‖A−B‖+ δr ‖A−B‖ = [(δ + ε)C(r, δ + ε) + δr] ‖A−B‖ , (65) where the induction hypothesis is used in the second inequality. Choose C(r + 1, δ + ε) = (δ + ε)C(r, δ + ε) + δr (66) and (62) follows. Now, we analyze the property of C(r, δ + ε). By (66), simple recursion yields C(r, δ + ε) = r−1∑ k=0 (δ + ε)kδr−1−k ≤ r(δ + ε)r−1. Since δ + ε < 1, the lemma is hence proved. A.5.5 Proof of Proposition 5.3.5 The proof essentially is similar to the proof of Theorem 5.3.3 except changing the norm. Thus, we only sketch the important steps and emphasize the differences. First, we invoke a simple fact about the equivalence between the spectral and entry- ∞ norms of a matrix An×m; that is, ‖A‖∞ ≤ ‖A‖ ≤ √ mn ‖A‖∞ . (67) 198 A.5. Proofs for Chapter 5 For the proposition, we only need the left inequality and thus prove it. Indeed, let (i0, j0) be the index pair with its corresponding entry attaining ‖A‖∞. Let x0 = (0, · · · , 1, · · · , 0)T where 1 is in the j0-th position. Then by definition of the spectral norm, we have ‖A‖ = sup ‖x‖=1 ‖Ax‖ ≥ ‖Ax0‖ = ‖aj0‖ ≥ ‖aj0‖∞ = |ai0j0| = ‖A‖∞ , where aj0 is the j0-th column of A. Furthermore, by (5.23) and the union bound ‖Σ?n − Σ‖∞ ≤ τ √ log p n (68) with probability greater than 1−p−8τ2/C24+2 that approaches to 1 whenever τ > C4/2. But this implies that ∥∥∥Σ̃n − Σ∥∥∥∞ ≤ ‖TtΣ?n − TtΣ‖∞+‖TtΣ− Σ‖∞ ≤ ‖Σ?n − Σ‖∞+2τ √ log p n = Op (√ log p n ) , (69) since ‖Tt(A)− Tt(B))‖∞ ≤ ‖A−B‖∞ + t. Now, the argument proceeds as in The- orem 5.3.3. Since ‖I − ηΣ‖∞ ≤ ‖I − ηΣ‖ < 1, we thus have, with high probability, ∥∥∥Ω̃− Ω∥∥∥ ∞ ≤ C(q,m, τ) ∥∥∥Σ̃n − Σ∥∥∥∞+ 1m ( 1−m2 1 +m2 )r+1 ≤ C(q,m, τ) (√ log p n + δr+1 ) . (70) A.5.6 Proof of Theorem 5.3.6 Before proving Theorem 5.3.6, we present a few technical lemmas that bound the magnitudes E ∥∥∥Σ̃n − Σ∥∥∥r on some “bad set”. Define B = {∥∥∥Σ̃n − Σ∥∥∥ > Ccn,p( log p n )(1−q)/2} . (71) 199 A.5. Proofs for Chapter 5 Lemma A.5.3. Under the assumptions in Theorem 5.3.6, there is a constant C(q,m, τ) > 0 independent of r, n, and p, such that we have E (∥∥∥Σ̃n − Σ∥∥∥r ;B) . [C(q,m, τ)cn,p( log p n )(1−q)/2]r . (72) Proof. By the proof of [11, Theorem 1], we extract the consequence that P (∥∥∥Σ̂n − Σ∥∥∥ ≤ C1cn,pt1−q) ≥ 1− C2p2 exp(−C3nt2), (73) where t = τ √ log p/n is the thresholding parameter (tending to 0); or equivalently, write P (∥∥∥Σ̂n − Σ∥∥∥ > t′) < C2p2 exp[−C4n( t′ cn,p )2/(1−q)] , (74) for t′ = C1cn,pt1−q. Applying Lemma A.5.4 with X = ∥∥∥Σ̂n − Σ∥∥∥ and t′, we therefore have that E (∥∥∥Σ̃n − Σ∥∥∥r ;B) = t′rP (X > t′) + ∫ Cr5 t′r P (X > u1/r) du+ ∫ ∞ Cr5 P (X > u1/r) du def = R1 +R2 +R3. We now develop upper bounds for R1, R2, and R3. For R1, we see from (74) that P (X > t′) ≤ C2p−C5+2 → 0 (75) and therefore it follows that for large enough τ R1 = o (( C1cn,pt 1−q)r) . (76) Writing R2 by definition and applying Lemma A.5.5 with α = 2/[r(1 − q)] and 200 A.5. Proofs for Chapter 5 C2 = C4c −2/(1−q) n,p , we deduce that R2 ≤ C2p2 ∫ ∞ t′r exp [ −C4n ( u1/r cn,p )2/(1−q)] du = C2p 2 r(1− q) 2 crn,p(C4n) −r(1−q)/2 ∫ ∞ C6 log p v r(1−q) 2 −1e−v dv, (77) where the last term is an upper incomplete gamma function Γ ( r(1− q) 2 ;C6 log p ) p−C6(C6 log p) r(1−q) 2 −1, as p→∞. But this now implies that R2 . [ C7cn,p ( log p n ) 1−q 2 ]r , (78) since p ≥ nξ and r = O(log n). Invoking LemmaA.5.5 twice, we see that R3 can be bounded in a similar means as in R2, except for the difference that the tail is sub-exponential rather than sub-Gaussian. The lemma now follows from (76) and (78). Lemma A.5.4. Let X be a non-negative r.v., t ≥ 0, and r ∈ N. Then E (Xr;X > t) = trP (X > t) + ∫ ∞ tr P (X > u1/r) du. (79) Proof. The lemma is an easy consequence of Fubini’s theorem. Lemma A.5.5. Let X be a non-negative r.v. and α > 0. Suppose that there are absolute constants C1 and C2 such that P (X > t) ≤ C1 exp (−C2ntα) (80) 201 A.5. Proofs for Chapter 5 for all t ∈ [a, b] where 0 ≤ a ≤ b ≤ ∞. Then we have ∫ b a P (X > u) du ≤ C1α−1(C2n)−1/α ∫ C2nbα C2naα vα −1−1e−v dv. (81) Proof. Let v = C2nu α. Then the lemma follows from a direct application of the change of variables. Proof of Theorem 5.3.6: By Theorem 5.3.3 and Lemma A.5.3, we have E ∥∥∥Ω̂− Ω∥∥∥2 = E (∥∥∥Ω̂− Ω∥∥∥2 ;Bc)+ E((∥∥∥Ω̂− Ω∥∥∥2 ;B) ≤ C(q,m, τ) [ c2n,p ( log p n )1−q + δ2(r+1) ] + E( (∥∥∥Ω̂− Ω∥∥∥2 ;B) . (82) ≤ C(q,m, τ) [ c2n,p ( log p n )1−q + δ2(r+1) ] +O ( 2r∑ j=2 Cjcjp,n ( log p n )j(1−q)/2) = O ( c2n,p ( log p n )1−q + δ2(r+1) ) , (83) since cn,p(log p/n) (1−q)/2 → 0. This is the content of theorem. A.5.7 Proof of Theorem 5.3.9 Considering the general lower bound in Proposition 5.3.8 with ψ(θ) = (Σ(θ))−1, similarly as in [22], we have inf Ω̂ sup Σ∈Gq(cn,p)∩U(m) E ∥∥∥Ω̂− Σ−1∥∥∥2 ≥ inf Ω̂ max θ∈Θ Eθ ∥∥∥Ω̂− Σ(θ)−1∥∥∥2 ≥α 4 p 4 min 1≤i≤J ∥∥P̄i,0 ∧ P̄i,1∥∥ , (84) where α = min (θ,θ′):H(γ(θ),γ(θ′))≥1 ‖Σ(θ)−1 − Σ(θ′)−1‖2 H(γ(θ), γ(θ′)) . (85) 202 A.6. Proofs for Chapter 6 Since Σ(θ) and Σ(θ′) belong to U(m), it easily follows that ∥∥Σ(θ)−1 − Σ(θ′)−1∥∥ ≥ m2 ‖Σ(θ)− Σ(θ′)‖ . (86) Now, the theorem follows from [22, Lemma 5 and Lemma 6]: that is, we have inf Ω̂ sup Σ∈Gq(cn,p)∩U(m) E ∥∥∥Ω̂− Σ−1∥∥∥2 ≥ C ′c2n,p( log pn )1−q (87) for some constant C ′ > 0. A.6 Proofs for Chapter 6 A.6.1 Proof of Theorem 6.2.1 The proof of the estimation and sign consistency for the group robust lasso estimator is based on the argument developed in [31, Theorem IV.1]. Let Zn be the objective function to minimized. Define Vn(u) = δ n∑ i=1 ( u∗ xix ∗ i n u ) + 1√ n n∑ i=1 u∗xiYi + (1− δ) n∑ i=1 Zi + λn G∑ g=1 [∥∥∥∥βg + ug√n ∥∥∥∥ 2 − ∥∥βg∥∥2] , (88) where Yi = (1− δ)(1(ei < 0)− 1(ei ≥ 0))− 2δei and Zi = 2× ( u∗xi√ n − ei ) 1 ( 0 ≤ ei < u ∗xi√ n ) . Then it follows that √ n(β̂ H n − β) minimizes Vn and the first three terms together converge in distribution to (δ + (1− δ)f(0))u∗Cu+ u∗W, 203 A.6. Proofs for Chapter 6 where W ∼ N(0, ((1− δ)2 + 4δ2σ2 + 4δ(1− δ)M10)C). For the last term, we divide into two cases. For βg 6= 0, we have √ n (∥∥∥∥βg + ug√n ∥∥∥∥ 2 − ∥∥βg∥∥2)→ ugTβg∥∥βg∥∥2 (89) by identifying the derivative of the Euclidean norm. So it follows that λn√ n︸︷︷︸ →0 ×√n (∣∣∣∣βj + uj√n ∣∣∣∣− |βj|)︸ ︷︷ ︸ →ug Tβg∥∥∥∥βg ∥∥∥∥ 2 P→ 0. (90) where we have used our assumption on the shrinkage rates λn/ √ n→ 0. For βg = 0, it is obvious that ∥∥∥∥βg + ug√n ∥∥∥∥ 2 − ∥∥βg∥∥2 = n− 12 ‖ug‖2 so we obtain that λnn − 1 2 ‖ug‖2 = λnn γ−1 2︸ ︷︷ ︸ →∞ ∥∥∥√nβ̂LSng ∥∥∥−γ 2︸ ︷︷ ︸ Op(1) ‖ug‖2 P→∞. (91) Putting all the terms together and applying Slutsky’s lemma, we deduce that Vn(u)⇒ V (u) for each u ∈ Rp where V (u) = (δ + (1− δ)f(0))u ∗Cu+ u∗W if ug = 0∀g /∈ A, ∞ otherwise. (92) Since Vn is convex and V has unique minimum, it follows from the standard epi- convergence results that √ n(β̂n−β) = argmin(Vn)⇒ argmin(V ) which is equivalent to say that √ n ( β̂nA − βA ) ⇒ C−1AAWA and √ nβ̂nAc P→ 0 with uA is the restriction of W to the support of the true coefficient vector. Now the theorem follows. 204
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