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Modeling latent correlation structures with application to agricultural and environmental science Bornn, Luke 2012

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Modeling Latent Correlation Structures with Application to Agricultural and Environmental Science by Luke Bornn B.Sc. Mathematics and Statistics, University of the Fraser Valley, 2006 M.Sc. Statistics, The University of British Columbia, 2008 A THESIS SUBMITTED IN PARTIAL FULFILLMENT OF THE REQUIREMENTS FOR THE DEGREE OF Doctor of Philosophy in THE FACULTY OF GRADUATE STUDIES (Statistics) The University Of British Columbia (Vancouver) July 2012 c© Luke Bornn, 2012 Abstract In this thesis, we explore the issue of latent correlation structure in spatial and other correlated systems. Firstly, we propose a class of prior distributions on decompos- able graphs, allowing for improved modeling flexibility. While existing methods solely penalize the number of edges, the proposed work empowers practitioners to control clustering, level of separation, and other features of the graph. Emphasis is placed on a particular prior distribution which derives its motivation from the class of product partition models; the properties of this prior relative to existing priors is examined through theory and simulation. We then demonstrate the use of graphical models in the field of agriculture, showing how the proposed prior distri- bution alleviates the inflexibility of previous approaches in properly modeling the interactions between the yield of different crop varieties. Secondly, we describe how spatial dependence can be incorporated into sta- tistical models for crop yield along with the dangers of ignoring it. In particular, approaches that ignore this dependence suffer in their ability to capture (and pre- dict) the underlying phenomena. Prior distributions are developed to accommo- date the spatial non-stationarity arising from distinct between-region differences in agricultural policy and practice. As a result, the model developed has improved ii prediction performance relative to existing models, and allows for straightforward interpretation of climatic effects on the model’s output. Lastly, we propose a novel approach to modeling nonstationary spatial fields. The proposed method works by expanding the geographic plane over which these processes evolve into higher dimensional spaces, transforming and clarifying com- plex patterns in the physical plane. By combining aspects of multi-dimensional scaling, group lasso, and latent variable models, a dimensionally sparse projection is found in which the originally nonstationary field exhibits stationarity. Following a comparison with existing methods in a simulated environment, dimension expan- sion is studied on a classic test-bed data set historically used to study nonstationary models. Following this, we explore the use of dimension expansion in modeling air pollution in the United Kingdom, a process known to be strongly influenced by rural/urban effects, amongst others, which gives rise to a nonstationary field. iii Preface A version of Chapter 2 has been published. Bornn, L., Caron, F. (2011) Bayesian Clustering in Decomposable Graphs. Bayesian Analysis. Vol. 6, No. 4, Pages 829- 846. I conducted all computation and numerical work, and hence generation of figures. The idea was jointly developed by myself and Dr. Caron, and I conducted the majority of writing and editing. A version of Chapter 4 has been published. Bornn, L., Zidek, J. (2012) Efficient Stabilization of Crop Yield Prediction in the Canadian Prairies. Agricultural and Forest Meteorology. Vol. 152, Pages 223-232. I conducted all computation and numerical work, and hence generation of figures. The idea was jointly developed by myself and Dr. Zidek, and I conducted the majority of writing and editing. A version of Chapter 5 has been published. Bornn, L., Shaddick, G., Zidek, J. (2012) Modelling Nonstationary Processes Through Dimension Expansion. Jour- nal of the American Statistical Association. Vol. 107, Pages 281-289. I conducted all computation and numerical work, and hence generation of figures. The idea was jointly developed by myself, Dr. Shaddick, and Dr. Zidek, and I conducted the majority of writing and editing. iv Table of Contents Abstract . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . ii Preface . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . iv Table of Contents . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . v List of Tables . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . viii List of Figures . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . ix Acknowledgments . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . xi Dedication . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . xiii 1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 1 1.1 Structure in Correlated Systems . . . . . . . . . . . . . . . . . . 3 1.1.1 Graphical Models . . . . . . . . . . . . . . . . . . . . . . 4 1.1.2 Correlation Functions . . . . . . . . . . . . . . . . . . . 7 2 Bayesian Clustering in Decomposable Graphs . . . . . . . . . . . . 9 2.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 9 v 2.2 Bayesian Inference on Decomposable Graphs . . . . . . . . . . . 11 2.3 Priors on Decomposable Graphs . . . . . . . . . . . . . . . . . . 13 2.3.1 Previous work . . . . . . . . . . . . . . . . . . . . . . . 13 2.3.2 A new prior distribution on decomposable graphs . . . . . 16 2.3.3 Extensions . . . . . . . . . . . . . . . . . . . . . . . . . 19 2.4 Example: Modeling Agricultural Output of Different Species . . . 20 2.5 Example: Modeling 20th Century American Voting Patterns . . . 24 2.6 Discussion . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 25 3 Spatial Statistics and Nonstationarity . . . . . . . . . . . . . . . . . 29 3.1 Geostatistics . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 30 3.2 Lattice Data . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 32 3.3 Nonstationarity . . . . . . . . . . . . . . . . . . . . . . . . . . . 33 4 Efficient Stabilization of Crop Yield Prediction in the Canadian Prairies 35 4.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 35 4.2 Materials and Methods . . . . . . . . . . . . . . . . . . . . . . . 38 4.2.1 Incorporating soil water . . . . . . . . . . . . . . . . . . 38 4.2.2 Incorporating temperature . . . . . . . . . . . . . . . . . 49 4.2.3 A context-specific spatial Bayesian approach . . . . . . . 53 4.3 Results . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 61 4.4 Conclusion . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 63 5 Nonstationary Modeling Through Dimension Expansion . . . . . . . 66 5.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 66 5.2 Dimension Expansion . . . . . . . . . . . . . . . . . . . . . . . . 69 vi 5.2.1 Illustrative example . . . . . . . . . . . . . . . . . . . . . 74 5.2.2 Image warping and folding . . . . . . . . . . . . . . . . . 76 5.3 Applications . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 78 5.3.1 Solar radiation . . . . . . . . . . . . . . . . . . . . . . . 78 5.3.2 Air pollution . . . . . . . . . . . . . . . . . . . . . . . . 80 5.4 Discussion . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 81 5.5 Optimization of Equation (5.2) . . . . . . . . . . . . . . . . . . . 85 6 Conclusion . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 87 6.1 Summary . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 87 6.2 Future Work . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 89 6.2.1 Nonstationarity . . . . . . . . . . . . . . . . . . . . . . . 89 6.2.2 Sparsity and group correlation . . . . . . . . . . . . . . . 90 6.2.3 Structural health monitoring . . . . . . . . . . . . . . . . 90 6.2.4 Monte Carlo . . . . . . . . . . . . . . . . . . . . . . . . 91 Bibliography . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 92 vii List of Tables Table 2.1 Predictive density for various binomial priors . . . . . . . . . . 24 Table 2.2 Predictive density for various PGM priors . . . . . . . . . . . 24 Table 4.1 Comparison of crop yield models . . . . . . . . . . . . . . . . 53 viii List of Figures Figure 2.1 Samples from binomial and PGM priors . . . . . . . . . . . . 15 Figure 2.2 Parameters of PGM prior . . . . . . . . . . . . . . . . . . . . 17 Figure 2.3 Posterior samples from agricultural model . . . . . . . . . . . 23 Figure 2.4 Highest posterior density for graphical model priors . . . . . . 26 Figure 4.1 Least squares residuals . . . . . . . . . . . . . . . . . . . . . 41 Figure 4.2 Correlation of SI and yield over time . . . . . . . . . . . . . . 43 Figure 4.3 Screeplot for SI . . . . . . . . . . . . . . . . . . . . . . . . . 45 Figure 4.4 Principal components and mean for SI . . . . . . . . . . . . . 46 Figure 4.5 Fit of crop yield models . . . . . . . . . . . . . . . . . . . . 48 Figure 4.6 Cross-validation of crop yield models . . . . . . . . . . . . . 49 Figure 4.7 Correlation of GDD and yield over time . . . . . . . . . . . . 51 Figure 4.8 Regression coefficient surfaces . . . . . . . . . . . . . . . . . 60 Figure 4.9 Cross-validation of Bayesian model . . . . . . . . . . . . . . 62 Figure 4.10 Cross-validation by region . . . . . . . . . . . . . . . . . . . 64 Figure 5.1 Empirical variograms from 3-D process and 3-D projection . . 75 Figure 5.2 Learned latent locations and corresponding empirical variogram 75 ix Figure 5.3 Warped grids and corresponding variograms . . . . . . . . . . 77 Figure 5.4 Locations and variogram for solar radiation data . . . . . . . 79 Figure 5.5 Dimension expansion of the solar radiation surface . . . . . . 80 Figure 5.6 Cross-validation on UK black smoke data . . . . . . . . . . . 82 Figure 5.7 Learned dimension of UK black smoke field . . . . . . . . . . 83 Figure 5.8 Variograms on UK black smoke data . . . . . . . . . . . . . . 84 x Acknowledgments I gratefully acknowledge the tremendous support of Arnaud Doucet, Raphael Got- tardo, and Jim Zidek, who as supervisors over my time at UBC filled me with an insatiable passion for research and a discerning eye towards quality. I am forever in debt for their continual mentorship, insights, and wisdom in helping me estab- lish an independent research programme and lay the groundwork for an academic career. I am honored to have been able to work with them, and am proud to call them friends. I had the tremendous pleasure to work with Marian Anghel, Sezer Atamturktur, François Caron, Julien Cornebise, Pierre Del Moral, Chuck Farrar, Todd Graves, Dave Higdon, Roman Holenstein, Pierre Jacob, Gyuhae Park, Gavin Shaddick, Ingo Steinwart, and Aline Tabet throughout my PhD studies. The friendships and collaborations developed with this fine group have not only improved my research, but also enriched my life. I also thank my supervisory committee, Claudia Tebaldi, Paul Gustafson, and Alexandre Bouchard-Côté, without whom you would not be reading this. I wish to thank Charles Serele, Harvey Hill, and Nathaniel Newlands, of Agri- culture and Agri-foods Canada for providing the data used in Chapter 4 as well as xi useful suggestions throughout the chapter’s development. Additionally, we thank numerous referees and editors for valuable input in the published works which drive this thesis. This work was funded by an NSERC PGS-D as well as a Michael Smith Research Trainee Award. Most importantly, I am forever grateful to my family for their unceasing love and support. xii Dedication To Katie xiii Chapter 1 Introduction While introductory statistics courses often begin with the topic of summarizing and visualizing single variables, attention inevitably turns to discussion of the relation- ship between two or more variables. Starting with correlation, focus quickly shifts to regression, classification, and other techniques. All of these tasks, however, rely on having several variables which are related in such a way that we can express their relationship mathematically. Take for example weight and height. If we are told a person’s height we can not guess their weight exactly, but it allows us to better estimate a person’s weight. In this way we say that while there is no perfect formula relating height to weight, the variables are correlated. Formally, we define the correlation (or more formally, Pearson’s correlation coefficient) between two random variables X and Y with expected values µX and µY and standard deviations σX and σY , respectively, as: ρX ,Y = corr(X ,Y ) = cov(X ,Y ) σXσY = E(X−µX)E(Y −µY ) σXσY . 1 Intuitively, if high values of X correspond to high values ofY , the two variables will be highly correlated, with a value close to 1. If high values of X correspond to low values of Y , the two variables will be negatively correlated, with a value close to −1. If there is no relation between values of X and Y , the correlation coefficient will be 0. We may calculate an empirical version of Pearson’s correlation coefficient, of- ten notated r, by substituting empirical means and standard deviations for the pop- ulation versions above. rX ,Y = 1 n−1 n ∑ i=1 ( Xi− X̄ sX )( Yi− Ȳ sY . ) Correlation is closely related to the concept of independence, which intuitively means that the outcome of one event does not influence the outcome of a second event. In fact, if two random variables X and Y are independent, then ρX ,Y = 0. Note that while independence implies zero correlation, the converse is not true, as Pearson’s correlation coefficient only captures linear relationships. While the explicit definitions above may seem overly pedantic, they are worth stating simply for the realization that a large portion of methodological development in the field of statistics (and this thesis!) is focused on this simple notion of correlation, extended to complex systems. When dealing with data which arrives sequentially in time, or across a spa- tial domain, observations will not be independent. For example, the average tem- perature tomorrow is largely influenced by the temperature today. Likewise in space, we expect the amount of precipitation which falls at the University of British Columbia to be related to the amount which falls in downtown Vancouver, as the 2 two locations are only 10km apart. As such, observations which are obtained spa- tially or temporally must be modeled appropriately, as they will not satisfy the independence assumption required by many methods such as standard linear re- gression. 1.1 Structure in Correlated Systems The way in which correlation arises in space and time is structured, a phenomena which is intuitively obvious. Using again the example of average temperature, the value tomorrow is likely to be largely influenced by the value today. However, the temperature tomorrow is likely less influenced by the temperature a year earlier. In this way, we intuitively understand that the correlation between values in time decreases as more time separates the observations. In contrast, sometimes patterns are cyclical. For example, whether a house is decorated with Christmas lights is uncorrelated with whether that house had Christmas lights a month earlier. How- ever, it would be highly correlated with whether the house had Christmas lights 365 days earlier. Similarly, whether a person attends a church service on Sunday is only weakly correlated with whether that person attended church the day before, on Saturday. A person’s attendance of a Sunday church service would be more highly correlated with their presence or absence in church a week previous. Similarly in space, we expect locations which are closer together to have more similar values. For instance, the temperatures in Seattle, WA and Vancouver, BC, or Boston, MA and Providence, RI are likely to be quite similar on a given day. In contrast the temperatures in Seattle, WA and Boston, MA are less likely to be close. Similarly for Vancouver, BC, and Providence, RI. Likewise with time, correlation is not always perfectly connected to the amount of spatial separation (as we will 3 detail later in our discussions of nonstationarity). For example, we would expect the proportion of people living in apartments to be more alike in cities separated by hundreds or thousands of kilometers than a city and one if its neighboring (lower- density) suburbs. Several methods exist for mathematically encoding this correlation structure. The first, through graphical models, allows us to specify which variables (loca- tions, or time points for instance) are correlated, and which are not. Through a graphical model we can encode, for example, the knowledge that given all neigh- boring counties the poverty rate in one county is independent of all other counties. Another method for mathematically expressing structured correlation is by assign- ing it functional form. This is the technique often taken in time series, for example in AR models, which mathematically express the distribution of the current value given the set of previous values. The two techniques of graphical models and cor- relation functions actually go hand in hand. Graphical models encode the structure of zero/non-zero correlation between variables, whereas correlation functions can express the value of correlation between variables which are correlated. 1.1.1 Graphical Models Graphical models are a convenient way to encode independence relationships be- tween variables in a multivariate distribution. This convenience derives from clear visualization techniques, as well as straight-forward computation of conditional and marginal densities [55]. Undirected graphical models, or Markov random fields, have a simple interpretation regarding independence: given a setC, nodes A and B are conditionally independent if all paths between A and B pass through the 4 set C. Formally, P(A∩B|C) = P(A|C)P(B|C) This class of graphical models is often used in spatial statistics, imaging, and physics to represent for example the independence relationships between regions, pixels, and molecules, respectively. Undirected graphical models used in this way have a direct correspondence with the inverse covariance matrix, Σ−1, in that a lack of an edge between these variables in the graphical model (indicating conditional independence) leads to a zero in the precision matrix Σ−1. Graphical models may also have directed edges, which allows for the modeling of more advanced relationship, including causation and deterministic relationships. For our purposes, however, we focus on undirected graphical models. We now introduce some notation; see, for example Dawid and Lauritzen [23], Lauritzen [55]. Let G = (V,E) be a graphical model with vertices or nodes V = {1, . . . ,n} and pairwise edges E. The pair of nodes {i, j} ∈ V are adjacent or neighbors if (i, j) ∈ E, and a subset C ⊂V is said to be complete if all its elements are adjacent to each other. A cycle is a sequence of unique adjacent nodes, where the first node is connected to the last node. In this way a cycle represents the path starting at a given node, through a sequence of unique nodes, and returning to the starting location. A cycle is chordless if all pairs of non-adjacent nodes in the cycle are not neighbors. In other words, a chordless cycle has no chord, or shortcut. A graph with no chordless cycles of length greater than two is said to be chordal, or decomposable. As will be shown in Chapter 2, decomposable graphs have several unique properties, one of which is that the likelihood of the graphical 5 model may be factorized in a computationally efficient way. Specifically, if P satisfies the conditional independencies implied by a decomposable graph G , then the likelihood of the graphical model specified by P can be factorized according to the graph’s cliques and separators p(y|G ,θ) = ∏ nc i=1 p(yCi |θCi) ∏nsj=1 p(yS j |θS j) where θ is a quantity parameterizing the graphical model P over the graph G and satisfying some consistency conditions with respect to G ([23]). Undirected graphical models often arise in spatial statistics, where one is in- terested in modeling counts, proportions, or some other quantities in neighboring regions. Each region, or areal unit, has a measurement attached to it, and corre- lation is typically modeled through the neighborhood structure. Specifically, re- gions which are physically neighbors are also neighbors in the undirected graph, or Markov random field. Hence given its neighbors, a given region is independent of all other regions. When conducting Bayesian inference on undirected graphical models, one must assign a prior distribution over the space of all possible graphs G for a given set of variables, or nodesV . For a handful of variables, one may assign a prior probability to each individual graph. However, as the number of graphs scales exponentially with the number of nodes, automated methods must be found. Current approaches have been limited in their ability to accommodate varying forms of prior infor- mation on the graph. For instance, in an effort to encourage interpretable graphs, the standard approach has been to penalize the number of edges (conditional de- pendencies) in the graph. However, many situations exist where one might expect 6 variables to be clustered together and the graph to exhibit block structure. At the moment no such prior distribution exists to handle this problem. In Chapter 2, we propose a class of prior distributions motivated from the class of product partition models which will allow improved flexibility in the specification of prior informa- tion on the graph. 1.1.2 Correlation Functions We may also mathematically encode correlation structure through functional forms. For example, consider a sequence of observations X1,X2, . . . ,XT , modeled with the class of autoregressive (AR) models. A simplified AR(p) model may be written as Xt = p ∑ i=1 βiXt−i+ εt where β1,β2, . . . ,βp are parameters of the model, and εt is the noise, or error, term. Through this form, we have implicitly defined a correlation function. For an AR(1) process, it is corr(Xt ,Xt−k) = β k1 . Correlation functions describe the correlation between random variables at differ- ent points in space or time. Typically, as with the AR example above, the correla- tion decreases with space or time. Or, more generally, the correlation is a function of distance in space or time, although in space it can be also modeled as a function of the vector between locations. In both time-series and spatial modeling, the value of the correlation modeling becomes particularly obvious when one turns to prediction. Because nearby obser- 7 vations in space or time are similar to each other, one may use the values at nearby points to predict at the unobserved location or time. In Chapter 4, we discuss a model for crop yield which uses this basic idea to stabilize prediction. Subse- quently, Chapter 5 explores more complex correlation functions where modeling correlation as a function of spatial distance is no longer justified. 8 Chapter 2 Bayesian Clustering in Decomposable Graphs1 2.1 Introduction This chapter is concerned with the inference of the conditional independence graph G of a multivariate random vector Y of dimension n, a problem sometimes re- ferred to as structure learning. We focus here on undirected decomposable graphs, whose popularity is mainly due to the tractable factorization they allow for the likelihood ([23, 55]); related work for directed graphical models can be found in [53]. Learning the conditional independence graph G is an onerous task due to the large number of graphs on a set of n nodes, or variables. It is possible us- ing optimization methods to find the graph which best fits the data according to some metric [31, 61, 88]; alternatively Bayesian model averaging may be used to 1A version of Chapter 2 has been published. Bornn, L., Caron, F. (2011) Bayesian Clustering in Decomposable Graphs. Bayesian Analysis. Vol. 6, No. 4, pages 829-846. [9] 9 accommodate uncertainty in the estimated graph, or maximum a posteriori estima- tion may be used to select a given model from the posterior over graphs. Such an approach relies on a prior distribution pi(G ) over the set of decomposable graphs of a given size; through Bayes theorem, this prior is updated based on the data to give an a posteriori estimate of the distribution over graphs. Current approaches have been limited in their ability to accommodate varying forms of prior information on the graph. For instance, in an effort to encourage interpretable graphs, the standard approach has been to penalize the number of edges (conditional dependencies) in the graph. However, many situations exist where one might expect variables to be clustered together and the graph to exhibit block structure. At the moment no such prior distribution exists to handle this problem. Our contribution in this thesis is to propose a class of prior distributions motivated from the class of product partition models, which will allow improved flexibility in the specification of prior information on the graph. The field of agriculture is particularly susceptible to the application of graph- ical models. Due to large spatial domains as well as multifarious crop varieties, it is valuable to have models which both handle the complexity of the biophysical process as well as allow straightforward interpretation. In particular, one might examine the set of zero/non-zero correlations between crop varieties’ yields, using the presence or absence of edges to make decisions regarding crop management, marketing, and insurance policies. In addition, due to small sample sizes in many agricultural applications, the choice of prior distribution becomes particularly im- portant. 10 2.2 Bayesian Inference on Decomposable Graphs We begin with a brief overview of graphical models, following the exposition in [23]; see also [55] for further details on graphical models. Let G = (V,E) be a graphical model with vertices V = {1, . . . ,n} and pairwise edges E. The pair of nodes {i, j} ∈ V are adjacent if (i, j) ∈ E, and a subset C ⊂ V is said to be complete if all its elements are adjacent to each other. A complete subgraph that is maximal (i.e. not contained within another complete subgraph) is called a clique. An ordering of the cliques of an undirected graph, (C1, . . . ,Cnc) is said to be perfect if the vertices of each clique Ci also contained in any previous clique C1, . . . ,Ci−1 are all members of one previous clique; that is, for i= 2,3, . . . ,nc Hi =Ci∩∪i−1j=1C j ⊆Ch for some h∈{1,2, . . . , i−1}. The sets Hi, i= 1, . . . ,nc−1 are called separators. We write S1, . . . ,Sns for the non-empty separators (some might appear multiple times). If an undirected graph admits a perfect ordering it is said to be decomposable. We associate to each vertex i a random variable Yi, with realizations yi. For A ⊆ V , let YA = {Yi|i ∈ A}. A distribution P over V is Markov with respect to G if, for any decomposition (A,B) of G , XA is independent of XB given XA∩B. The widespread use of decomposable models is due to the resulting factorization of densities. Specifically, if P satisfies the conditional independencies implied by a decomposable graph G , then the likelihood of the graphical model specified by P can be factorized according to the graph’s cliques and separators p(y|G ,θ) = ∏ nc i=1 p(yCi |θCi) ∏nsj=1 p(yS j |θS j) (2.1) 11 where θ is a quantity parameterizing the graphical model P over the graph G and satisfying some consistency conditions with respect to G ([23]). Traditionally, focus has been on Gaussian graphical models, also known as covariance selection models ([26]) where P = Nn(µ,Σ) is a n-dimensional multi- variate Gaussian distribution and θ is the n× n covariance matrix Σ. Conditional independence structure is represented by the precision matrix Σ−1. If the edge (i, j) /∈ E, then the variables Yi and Yj are conditionally independent given the re- maining variables, and Σ−1(i, j) = Σ −1 ( j,i) = 0. As such, the Gaussian graphical model may be factorized as (2.1) with the covariance Σ replacing θ , and the corresponding likelihood terms written as p(yB|ΣB) = (2pi)−|B|/2 det(ΣB)−|B|/2 exp[−12 tr(SB(ΣB) −1)] (2.2) for each complete set B, where |B| denotes the cardinality of B and SB is the em- pirical covariance matrix of yB. From a Bayesian perspective, we are interested in the posterior distribution p(θ ,G |y) ∝ p(y|θ ,G )p(θ |G )pi(G ). Much work has been dedicated to specifying proper priors p(θ |G ), see e.g. ([23, 36]). The main focus of this chapter is the specification of a prior distribution pi(G ) over the space of decomposable graphs. As this space is very large compared to the number of observations, it is crucial to add as much prior information as possible on the structure of the unknown graph G . Moreover, we are generally interested in obtaining sparse graph estimates for needs of interpretation and prediction. Up until now, the specification of pi(G ) has been limited to the uniform distribution, or priors which penalize the complexity as measured by the number of edges. This brings us to the focus of this work, namely 12 a class of prior distributions pi(G ) which subsumes control over the structure and features of G . 2.3 Priors on Decomposable Graphs 2.3.1 Previous work While early work on inference in decomposable models often assumed a uniform prior over graphs (i.e. [36]), such priors put considerable mass on models of in- termediate size. In an effort to put more weight on smaller graphs, several authors have proposed using a binomial prior distribution with parameter ρ on the number of edges r in the graph. This yields priors of the type [27, 50] pi(G ) ∝ ρr(1−ρ)m−r (2.3) where m = n(n−1)2 is the maximal number of possible edges on n nodes. When ρ = 1/2, it reduces to the forementioned uniform prior over graphs. [50] suggest the use of ρ = 2/(n− 1), motivated from the resulting density’s peak at n edges in the unconstrained graph. Some authors also consider adding a hierarchical Beta prior ρ ∼ Be(a,b) ([19]), giving the marginal prior on the graph as pi(G ) = ∫ 1 0 pi(G |ρ)pi(ρ)dρ ∝ β (a+ k,b+m− k) β (a,b) where β (·, ·) is the beta function. [19] suggest a default choice of a = b = 1, implying a uniform prior on ρ . Interestingly, the resulting prior on G is pi(G ) = 1 m+1 ( m r )−1 , 13 which penalizes medium-sized graphs as desired. Such a prior weights each graph according to the number of graphs in the unrestricted space with the same number of edges. However, as shown by [3], the space of decomposable graphs can be con- siderably different than the unrestricted space. To address this, [3] have proposed a uniform prior on decomposable graphs given the number of edges. However, calculating the number of decomposable graphs of a given size is an arduous task: there exists no list in the literature of decomposable graphs and their breakdown in terms of number of edges, nor are there straightforward ways of computing such quantities. As a result, [3] proposes an MCMC estimation scheme, testing its ac- curacy up to 12 nodes, although such a scheme will likely become prohibitive in higher dimensions. While the priors in the above references allow one to control the size of the resulting graphs through the number of edges, often doing so results in undesirable graph structures, namely those with a high number of separators and long strings of nodes. Figure 2.1(top) shows random samples from a binomial prior over 20- node graphs with ρ = 0.1 (closely echoing the choice of [50], namely ρ = 2/(n− 1) ≈ 0.1) and ρ = 0.5 (the uniform prior). We see from this plot that there is no clustering of the cliques, making interpretation difficult. In addition, the long strings/trees seen for ρ = 0.1 do not mesh with reality in most cases. Clearly such a class of priors is not suitable if one suspects clustering amongst the variables, clique sizes to be upper (or lower) bounded, or nearly full separation between cliques. Our focus therefore is on moving beyond priors which focus on the number of edges to priors which focus on graph (clique and separator) structure. 14 !" # $ % & ' ( ) !*!! !" !# !$ !% !& !' !( !) "* ! " # $ % &' ( ) !* !! !" !# !$ !% !& !' !( !) "* ! " # $ % & ' ( ) !* !! !" !# !$ !% !& !' !( !) "* ! " # $ % & ' ( ) !* !! !" !# !$ !% !& !' !( !) "* ! " # $ % & ' ( ) !* !! !" !# !$ !% !& !' !( !) "* ! " # $ % & ' ( ) !* !! !" !# !$ !% !& !' !( !) "* ! " # $ % & ' ( ) !* !! !" !# !$ !% !& !' !( !) "* ! " # $ % & ' ( ) !* !! !" !# !$ !% !& !' !( !) "* ! " # $ % & ' ( ) !*!! !"!# !$ !% !& !' !( !) "* ! " # $ % & ' ( ) !* !! !" !# !$ !% !& !' !( !) "* ! " # $ % & '( ) !* !! !" !# !$!% !& !' !( !) "* ! 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"* Clique Sizes: 2(17), 1(2) Separator Sizes: 1(11) Clique Sizes: 2(14), 1(4) Separator Sizes: 1(6) Clique Sizes: 9, 8, 6, 3 Separator Sizes: 6 Clique Sizes: 9, 6, 3, 2 Separator Sizes: N/A Clique Sizes: 10, 5, 4, 1 Separator Sizes: N/A Clique Sizes: 5(2), 4, 3, 2, 1 Separator Sizes: N/A Clique Sizes: 6, 2(2), 1(10) Separator Sizes: N/A Clique Sizes: 4, 3(2), 2 (3), 1(3) Separator Sizes: N/A Clique Sizes: 7, 3(2), 1(8) Separator Sizes: 1 Clique Sizes: 4, 3, 2(3), 1(7) Separator Sizes: N/A Clique Sizes: 8(2), 7(3), 6(4), 5(2), 4(2) Separator Sizes: 7, 6(3), 5(4), 4(2), 3(2) Clique Sizes: 8, 7(3), 6(5), 5(3), 4 Separator Sizes: 6(3), 5(5), 4(3), 3 Clique Sizes: 8(3), 7, 6(5), 5(3), 4 Separator Sizes: 7(2), 6, 5(5), 4(3), 3 Clique Sizes: 10, 9(4), 7(4), 4(2) Separator Sizes: 8(4), 6(4), 3(2) Clique Sizes: 2(14), 3(2), 1 Separator Sizes: 1(9) Clique Sizes: 3, 2(8),1(5) Separator Sizes: 1(4) Figure 2.1: Four random samples from binomial and product graphical model (PGM) priors. Clique and separator sizes for each graph are also shown (“Clique Sizes: 2(3)” implies 3 cliques of size 2). 4 million samples were generated using Markov chain Monte Carlo, and every millionth is shown. While the binomial is characterized by large strings and many separators, the product graphical model allows one to induce clustering by setting b small. 15 2.3.2 A new prior distribution on decomposable graphs Motivated from the class of product partition models ([7, 8, 42]), we consider prior distributions of the form pi(G ) ∝ ∏ncj=1ψC(C j) ∏nsj=1ψS(S j) (2.4) where ψC and ψS are respectively called the clique/separator cohesion functions, with the convention that ψS( /0) = 1. Evidently one could choose to penalize only cliques or separators by setting ψC or ψS to constant values. Alternatively, one could simply penalize clique sizes by setting ψB = a|B|. Motivated from the class of product partition models, consider the cohesion functions ψC(B) = a(|B|− 1)! and ψS(B) = 1b(|B|−1)!, a> 0, b> 0, hence pi(G ) ∝ ancbns ∏ncj=1(|C j|−1)! ∏nsj=1(|S j|−1)! (2.5) The factorial terms result in predilection towards large cliques and small separators – a desirable trait in terms of interpretability of the resulting graph. For instance, even if a = b = 1 with 20 nodes, the completely connected graph would be pre- ferred over the complete independence graph by a factor of 20!. The parameters a and b respectively tune the number of cliques and separators in the decompos- able graph. For a small, the prior will favour a small number of large cliques. Likewise for b, with small values favouring fewer separators. Figure 2.1 (bottom) shows samples from this prior. Because of its relation to product partition models (described later), we term this prior the product graphical model prior. To clearly demonstrate the control the product graphical model prior (2.5) gives relative to the binomial prior, we set b= 1/1000, highly penalizing the number of separators and 16 Figure 2.2: Log ratio of priors over two graphs for product graphical model prior for various a, b (solid, bottom axis) and binomial prior for various ρ (dashed, top axis). While the binomial prior allows one to control the number of edges, for instance choosing G1 over G2, the same parameter would seldom choose G3 over G4, despite G3 having a sparse covariance matrix, and G4 having a saturated covariance matrix. hence resulting in highly separated cliques. In addition, we look at two different values for a; a= 0.1, resulting in fewer and larger cliques, and a= 10, resulting in more (but smaller) cliques. Fig. 2.1 demonstrates the ability of the prior to induce clustering of the cliques, and therefore sparsity in correlation. We have seen some general properties of the prior (2.5), namely the ability to control the number of cliques and separators. Figure 2.2 shows log10(pi(G )/pi(G ′))) for different graphs G ,G ′. Specifically, decreases in b result in increased prior probability on models with few separators; in addition we see that as a is increased, more mass is put on models with many cliques. In contrast, we also plot the same ratio for the binomial prior (2.3). From this, one can see the limited control such a prior gives, favouring small models in terms of number of edges, but putting very little mass on models, for example, which feature clusters of fully-connected nodes (as in G3) and therefore have sparse covariance matrices. 17 Selecting the appropriate cohesion functions in equation (2.4) is a difficult problem, but one for which we may gain insight from the existing literature on product partition models ([21, 75, 76]). For instance, one may use Figure 2.2 to select a and b to best fit with prior intuition regarding the features of the graph, then verify the choice through generation of Monte Carlo samples from the prior as in Figure 2.1. Alternatively, cross-validation or related methods may be used to select a and b; due to the potential computational cost of such methods, sequential Monte Carlo approaches may be used to speed up prior distribution selection ([12]). Given that the likelihood decomposes as (2.1) and the prior is of the form (2.4), the posterior will also be of the form (2.4) with cohesions ψC(C j)p(yC j) and ψS(S j)p(yS j). The prior admits several other attractive properties and connec- tions with well-known clustering methods as well. If ψS(S j)→ ∞ for all S j 6= /0, then Equation (2.4) reduces to the following model pi(G ) ∝ nc ∏ j=1 ψC(C j) if ns = 0 and 0 otherwise. The resulting prior puts only positive mass on graphs with no separators. It has been introduced as a prior over partitions by [42] and [7, 8] under the name of product partition models. In the particular case of (2.4) with b→ 0, the prior over G reduces to pi(G ) = ancΓ(a) Γ(a+n) nc ∏ j=1 (|C j|−1)! As shown by [76] (see also [75]), this is the distribution over partitions induced by 18 a Dirichlet Process [1, 30]. We also have E(nc) = n−1 ∑ i=0 a a+ i ' a log(1+n/a)+ γ, var(nc) = n−1 ∑ i=1 ai (a+ i)2 where γ is Euler’s constant and pr(nc = k) = s(n,k)akΓ(a)/Γ(a+n) where the coefficients s(n,k) are the absolute values of Stirling numbers of the first kind [1]. In this limiting case, the number of cliques increases logarithmically with the number of nodes. 2.3.3 Extensions Motivated by the larger class of exchangeable partition functions [54, 70], we can also consider four-parameter models, allowing more control over the relative sizes of the cliques/separators pi(G ) ∝ ∏ncj=1(a2+a1( j−1))Γ(|C j|−a1)Γ(1−a1) ∏nsj=1(b2+b1( j−1))Γ(|S j|−b1)Γ(1−b1) where a2 >−a1,0≤ a1 < 1, likewise for b1,b2. The above model reduces to (2.5) when a1 = b1 = 0. We can also consider models that control the maximal number of cliques/separators pi(G ) ∝ ∏ncj=1(c1− j+1)Γ(c2+|C j|)Γ(c2) ∏nsj=1(d1− j+1)Γ(d2+|C j|)Γ(d2) 19 where c1,c2,d1,d2 > 0, and c1 > d1 are the maximal number of cliques/separators. These two models respectively admit as limiting cases the distribution over par- titions induced by the two-parameter Poisson-Dirichlet distribution and the finite Dirichlet-multinomial distribution, see e.g. [54] for further details on these distri- butions. Using such extensions, one is able to both extend the product graphical model prior to control relative sizes and the maximal number of cliques and sep- arators, as well as borrow from the wealth of literature on Dirichlet and related distributions to gain insight into the prior distribution’s characteristics. 2.4 Example: Modeling Agricultural Output of Different Species Determining agricultural policies to govern crop production, harvesting, and export is a challenge fraught with high variability both temporally and spatially. Enabling effective crop management, handling, and marketing techniques thus requires ac- curate understanding of crop yield to account for and explain these variations. While much effort has been made in developing models for predicting single crops ([10, 72]), little effort has been made in understanding statistically the relationship between crop yield of different crop varieties. Understanding the connection between yields of different crop varieties is valu- able for a multitude of reasons. Firstly, because crops are planted and harvested at different times, the management of one crop might benefit from knowledge ob- tained from harvesting a similar crop earlier in the year. Additionally, by account- ing for correlation between different crops, insurers might better cover themselves against extreme events and better control insurance rates for farmers. Lastly, farm- ers themselves might wish to ensure some level of stability in their income, and 20 therefore might prefer to plant crops which are uncorrelated in yield. Through such a practice, a farmer would be proactive in preventing disasters across his en- tire crop portfolio. Simply by looking at the resulting undirected graph, a farmer could select two crops which do not have a path connecting them, and are therefore uncorrelated. We examine the total production (in thousands of bushels) of 24 crops in the state of California from the years 1990 to 2009 (20 years). The data is compiled from the U.S Department of Agriculture website, where a considerable database is available for viewing and analysis. The 24 crops include, for example, several va- rieties of wheat, rice, and beans. We use the now-standard Gaussian hyper-inverse Wishart model: the likelihood of yield is given in (2.1) and (2.2), and the prior for the covariance matrix Σ is hyper-inverse Wishart, which factorizes similarly to (2.1), as a ratio of inverse Wishart distributions over cliques and separators ([37]). To be specific, for each cliqueC (or, equivalently, separator S), the covariance over the clique is distributed as ΣC ∼ IW (b,DC), with parameters b and DC and density pi(ΣC|b,DC) ∝ |ΣC|−(b+2|C|)/2 exp { −1 2 tr(Σ−1C DC) } . See [19] for some alternative marginal likelihoods based on fractional Bayes factors which can help to induce parsimony. The parameters chosen for the hyper- inverse Wishart distribution are as described in [50]; we focus on the specification of pi(G ). Looking at the list of crops, one would expect that there will be clustering of the yields according to crop characteristics. For instance, it would be reasonable to expect the yield of beans to be correlated with each other. We also seek an interpretable graph, namely one with small complexity (in terms of number of 21 edges and/or separators). The first such prior we examine is the binomial prior of [50] with ρ = 2/(n− 1), chosen due to its prevalence in the literature. While such a prior allows for penalization on the number of edges, no control is available over clustering. In contrast, by using the prior (2.5), we can set b = .01 to put strong penalization on the number of separators (and hence induce separation of the cliques and therefore sparsity in the correlation matrix), and set a = .01 to encourage a small number of cliques in the pursuit of simplicity in the resulting graph. We run MCMC of length 10 million over the space of decomposable graphs ([36]) for both the binomial and product graphical model priors, thinning to every 100 samples. With both priors, one may save computational resources by making local moves, merging and splitting cliques within the Markov chain. As a result, one need not re-determine the structure of the entire graph at each move. Figure 2.3 shows the 4 graphs with highest posterior probabilities from each prior. The product graphical model prior results in the top 4 graphs having pos- terior density values in the range 0.11 to 0.49, whereas for the binomial the range is 0.04 to 0.06, indicating that the binomial prior spreads mass much more evenly across distributions relative to the product graphical model prior with a= b= 0.01. Immediately evident from the figure are the different forms resulting from each prior. Specifically, the binomial prior induces long strings of nodes with many separators, whereas the product graphical model posterior reflects our prior beliefs that variables will cluster together, resulting in sparsity in the correlations between variables. A commercial farmer desiring to plant two plots with uncorrelated crops to minimize the risk of loss might reach quite different conclusions from each prior. Specifically, the large strings of nodes from the Binomial prior suggest correlation 22 12 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24 1 2 3 45 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24 binom(2/(n-1)) 1 2 34 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 1718 19 20 21 22 23 24 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 241 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 1920 21 22 23 24 PGM(0.01, 0.01) 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 1819 20 2122 23 24 1 2 3 4 5 6 7 89 10 11 12 1314 15 16 17 18 19 20 21 22 23 24 Crop Varieties 1: wheat (winter)   2: wheat (durum)   3: rice (long grain)   4: rice (medium grain)   5: rice (short grain)   6: corn (grain)   7: corn (silage) 8: oats   9: barley   10: cotton (upland)   11: cotton (pima)   12: beans (large lima)   13: beans (baby lima)   14: beans (light kidney) 15: beans (dark kidney)   16: beans (blackeye)   17: beans (other)   18: hay (alfalfa)   19: hay (other)   20: potatoes (winter) 21: potatoes (spring)   22: potatoes (summer)   23: potatoes (fall)   24: sweet potatoes Figure 2.3: Four samples with highest posterior probability from crop yield model using binomial and product graphical model (PGM) priors. We see the bean yields (nodes 12 through 17) seem to cluster together, as do summer and fall potatoes (nodes 22 and 23). We also observe that the product graphical model prior induces separated cliques, whereas the binomial prior results in long strings and trees of connected variables. As a result, the product graphical model prior will induce sparsity in the resulting posterior covariance. between the majority of crops. The farmer might not plant winter wheat (planted in late fall) and a strain of beans (harvested in early fall) on his two plots, despite their very different growing seasons, due to their connection in two of the highest poste- rior probability graphs in Figure 2.3. In contrast, the separation of cliques from the product graphical model prior (2.5) would allow these crops to be planted together. Such decisions could be made from the highest posterior graph, or by conducting Bayesian model averaging to obtain the expected utility of a given decision. To gain an understanding of the product graphical model prior’s prediction 23 Table 2.1: Log predictive density evaluated on test data using various bino- mial priors Distribution: Binomial Parameters: 2/(24−1) 0.5 Avg. Log Predictive: −688 −707 Avg. Number of Edges: 17.8 29.6 Table 2.2: Log predictive density evaluated on test data using various PGM priors Distribution: PGM Parameters: (0.01,0.01) (0.1,0.1) (1,1) Avg. Log Predictive: −675 −677 −686 Avg. Number of Edges: 18.1 16.4 20.3 performance, we split the data into a training set (first 12 years) and testing set (last 8 years). After simulating from the posterior distribution arising from the binomial and product graphical model priors, we use Bayesian model averaging via the marginal likelihood evaluated on the test data to judge the model’s prediction performance. We evaluate the resulting posterior predictive evaluated on the test set in Tables 2.1, 2.2; indeed, the product graphical model prior provides better prediction in this example, even over a variety of parameter choices. We also show the number of edges for each model, indicating that sparsity in terms of edges alone is not responsible for the improved prediction. 2.5 Example: Modeling 20th Century American Voting Patterns In an effort to demonstrate the product graphical model prior in higher dimensions, we now turn to the modeling of American voting data by state. For each federal 24 election from 1904 to 1976, occurring every four years, we measure the proportion of votes for the republican party in each of the 50 states ([18]). Our goal is to model and visualize correlation in voting pattern changes over the last century. Some immediate questions come to mind: “Do certain states have an important role in determining election outcomes?”, “Are there groups of states which vote together, operating independently from the US as a whole?” We proceed by exploring the posterior distribution resulting from the binomial prior with edge probability 0.1, and the product graphical model prior with param- eters a = 10,b = 10−3, in an effort to make the overall number of edges resulting from each model comparable. Figure 2.4 shows the two graphs with highest pos- terior density from each model. As expected, the binomial graphs contain long strings of variables, while the product graphical model prior demonstrates cluster- ing and grouping of variables. While the binomial prior results in similar variables placed along the same string, the grouping from the product graphical model allows for clearer interpretation. For instance, we immediately observe that the southern states (SC, MS, LA, AL, GA, TX, VA, FL) generally vote in a group. Other pat- terns of interest also arise, including a close connection between AR, NC, and TN. Also, notice that NY and KS are consistently the single node connecting clusters of variables. As such, these states might be considered as key indicators of voting behavior. 2.6 Discussion The product graphical model’s implementation relies on Markov chain Monte Carlo, whereby samples of graphs from the posterior distribution are generated and infer- ence is made from their empirical distribution. Because of the sheer number of 25 0 0.1 0.2 0.3 0.4 0.5 0.6 0.7 0.8 0.9 1 0 0.1 0.2 0.3 0.4 0.5 0.6 0.7 0.8 0.9 1 AL AZ AR FL GA HI ID LA ME MD MA MS NE NV NM NY NC ND OH OK RI SC SD TN TX UT VT VA WA WV CO OR KY WY MT AK WI IA KSNJ DE INCT NH MN CA ILMO PA MI (a) PGM(10,10−3): HPD Graph 1 0 0.1 0.2 0.3 0.4 0.5 0.6 0.7 0.8 0.9 1 0 0.1 0.2 0.3 0.4 0.5 0.6 0.7 0.8 0.9 1 AL AK AZ AR CA CO CT DEFL GA HI ID ILIN IA KS KY LA ME MD MA MI MN MS MO MT NE NV NH NJ NM NY NC ND OH OK OR PA RI SC SD TN TX UT VT VA WA WV WI WY (b) Binom(0.01): HPD Graph 1 0 0.1 0.2 0.3 0.4 0.5 0.6 0.7 0.8 0.9 1 0 0.1 0.2 0.3 0.4 0.5 0.6 0.7 0.8 0.9 1 AL AK AZ CT DE GA HI INKY LA ME MD MA MS MT NH NJ ND OR SC TN UT VT WV PA CA MI OH KS CO WA IA NE AR MO NC IL FL OK TX VA NYNM RI ID NVWY MN SD WI (c) PGM(10,10−3): HPD Graph 2 AL AK AZ AR CACO CT DE FL GA HI ID IL IN IA KS KY LA ME MD MA MI MN MS MO MT NE NV NH NJ NM NY NC ND OH OK OR PA RI SC SD TN TX UT VTVA WAWV WI WY (d) Binom(0.01): HPD Graph 2 Figure 2.4: Voting example: two graphs with highest posterior density (HPD) from binomial and product graphical model priors. 26 graphs, one must be careful in interpreting the results of this approximation, hence our use of model averaging (for prediction) as well as studying a suite of graphs with high posterior probability. While fast and easily implementable software for the above models has been generously released by [50], one must still conduct checks for convergence of the Markov chain. This is exacerbated by the complex- ity of the space of decomposable models, whereby two highly probable models might be separated by a series of models of low probability. This is particularly true for the product graphical model prior; if one chooses a prior which highly pe- nalizes the number of separators (say, using b very small), transversing the space of graphs may be difficult, as the fully disconnected graph and fully connected graph both have no separators (and therefore high probability), but necessarily have a considerable number of graphs with many separators between them. While we have focused on the Bayesian approach to covariance selection, sig- nificant work has also been done in a non-Bayesian framework. A common ap- proach involves placing an `1 penalization on the precision matrix Σ−1, which leads to sparse estimates ([31, 61, 88]). Closer to the heart of this chapter, [60] examine the case of estimating G when clustering is expected, and therefore Σ−1 exhibits block structure. However, these models are neither decomposable nor generative. While we have focused in this article on Gaussian graphical models, the prior defined in this article is far more general and can be used with any type of model for handling discrete or mixed data, see e.g. [55, 57]. We have also considered the hy- perparameters a and b to be known constants. Estimating them within the MCMC sampler would require one to compute the normalizing constant in (2.5), which is in general not tractable. An exception of interest is the case b→ 0, where we can assign a gamma prior to a and use the data augmentation algorithm described 27 in [86] to update a given the other variables. In conclusion, the proposed product graphical model prior improves flexibility in modeling decomposable graphical models and borrows strength from the im- mense literature on product partition and related models. The product graphical model prior allows one to encourage (or discourage) clustering of the graphs, and therefore can induce sparsity in the correlation matrix through clique separation; consequently, the product graphical model empowers practitioners to encapsulate their true prior beliefs to build a model more attuned to the problem at hand. 28 Chapter 3 Spatial Statistics and Nonstationarity In spatial statistics, the role of correlation plays a particularly crucial role. Here observations are tied together through coordinates of geographic systems. Appli- cations range from modeling astrophysical systems by way of their locations within galaxies, to modeling agricultural output or air pollution over large and diverse re- gions. The defining characteristic of spatial statistics is that each entity under study has attached to it a spatial location; how that location is defined, whether latitude/- longitude or arbitrary coordinates, varies depending on the example at hand. Our focus is on point-referenced geostatistics and lattice, or areal, data. The latter are typically modeled using Markov random fields, or undirected graphical models, as discussed earlier. In contrast, the former is employed to model spatial processes over continuous domains. 29 3.1 Geostatistics The origins of geostatistics is Kriging, where the goal is to interpolate the value of a random field at an unobserved location using observations from nearby locations. Kriging works by computing the best linear unbiased estimator at the unobserved location, using information about the correlation present in the system. The Krig- ing estimator is a linear combination of the observed values Z(x1), . . . ,Z(xs) with weights w1, . . . ,ws, ˆZ(x0) = s ∑ i=1 wi(x0)Z(xi). The weights wi, i= 1, . . . ,s are chosen to minimize the Kriging variance Var ( Ẑ(x0)−Z(x0) ) while maintaining the unbiasedness condition, E ( Ẑ(x)−Z(x))= s∑ i=1 wi(x0)µ(xi)−µ(x0) = 0 Classically, the correlation in the system was modeled through variograms, which describe the spatial dependence in a spatial random field through the vari- ance of the difference between field values at two locations xi,x j: 2γ(xi,x j) = var(Z(xi)−Z(x j)) 30 or if the field has a constant mean, 2γ(xi,x j) = E(|Z(xi)−Z(x j)|2). For those more familiar with the notation of Gaussian processes, where covariance functions are used, variograms may be related as follows: 2γ(xi,x j) =C(xi,xi)+C(x j,x j)−2C(xi,x j)+(E(Z(xi))−E(Z(x j)))2 For observation z1, . . . ,zs, the empirical variogram is calculated as γ̂(h) = 1 |N(h)| ∑i, j∈N(h) |zi− z j|2 where N(h) is the set of observations within a small tolerance of the distance h. It is straightforward to see that the empirical variogram provides an unbiased estimate of the theoretical variogram. E [γ̂(h)] = 1 2|N(h)| ∑i, j∈N(h) E [Z(xi)−Z(x j)]2 = 1 2|N(h)| ∑i, j∈N(h) 2γ(x j− xi) = 2|N(h)| 2|N(h)|γ(h) Many variants on Kriging exist. Simple Kriging assumes a constant trend, µ(x) = 0. Ordinary Kriging assumes an unknown, but constant, trend µ(x) = µ , which is extended by Universal Kriging to polynomial trend models. By assigning a parameteric form to the variogram and learning the parameters, 31 Kriging is able to provide the best linear unbiased estimate of the field at unob- served locations. While the shape of the parametric variogram may be controlled by many parameters, the primary parameters of interest are the nugget, which de- scribes the measurement error, the sill, which describes the limit as the variogram tends to infinite lag distances, and the range, which describes where the variogram has nearly reached the sill. 3.2 Lattice Data The primary goal of areal, or lattice, models is to detect spatial patterns. For ex- ample, are areas closer to eachother more similar than areas which are far apart? An additional goal can be to smooth the data. For example, if modeling health counts, we may have low population in some regions, and as such counts may be unreliable. As a result, we may wish to smooth observations using counts from (higher population) neighboring regions. Lastly, we sometimes want to change the boundaries of areas, or extend the region to include new areas. In this setting, we’d like to extend the results from the current areas to predict value in the new region. One of the first steps in areal data is defining the neighborhood structure. This may be done straightforwardly using graphical models, as discussed in the previous chapter. Alternatively, or more classically, we may think of a weight matrix W consisting of elements wi j. Typically, wii = 0 and wi j = 1 if areas i and j are neighbors. From this weight matrix, we can then calculate the areal analogue of a variogram, called Geary’s C: C = (n−1)∑i∑ jwi j(Yi−Yj)2 ∑wi j∑i(Yi−Y )2 . 32 For the purposes of the subsequent chapter, however, we model the areal data as point-referenced data, with the spatial reference point defined from the centroid of each region. 3.3 Nonstationarity Informally, if a spatial (or temporal) random process has the same joint probability distribution at all points in space (or time), it is said to be stationary. Specifically, suppose {Z(xi)} is a stochastic process defined at points xi, i= 1, . . . ,s with cumu- lative distribution function F (Z(x1), . . . ,Z(xs)). If for all δ , F (Z(x1), . . . ,Z(xs)) = F (Z(x1+δ ), . . . ,Z(xs+δ )) . As such, in these situations the field’s mean and variance do not change over time or space. In fact, if only the first and second moments of the field are constant over space and time, we say that the process is weak-sense stationary. If this holds, the semi-variogram (γ) may be represented as a function, γ(xi,x j) = γ(xi− x j). Further, if the process is isotropic, then the variogram may be represented solely as a function of the distance between locations, γ(xi,x j) = γ(||xi− x j||). 33 If the process is stationary, the relationship between variogram and covariance function is also simplified: 2γ(xi,x j) =C(xi,xi)+C(x j,x j)−2C(xi,x j) In the following chapter, Chapter 4, we model crop yields in the Canadian Prairies, and discover that differences between provinces leads to an inherently nonstationary spatial field. Subsequently, in Chapter 5, we propose a new method- ology for handling nonstationary spatial fields, whereby nonstationary fields are expanded into a higher-dimensional space where stationarity holds. 34 Chapter 4 Efficient Stabilization of Crop Yield Prediction in the Canadian Prairies1 4.1 Introduction This chapter presents a method for forecasting wheat crop yields in the Canadian Prairie Provinces – a challenging task due to dramatic variability in yield over space and time. Its importance, however, should not be understated: wheat is one of Canada’s primary exports, accounting for 12 percent of wheat and barley traded in the world market. Thus variation in yield has considerable impact both within and beyond Canadian borders ([81]). Enabling effective crop management, han- dling, and marketing thus requires accurate predictions of crop yield that account 1A version of Chapter 4 has been published. Bornn, L., Zidek, J. (2012) Efficient Stabilization of Crop Yield Prediction in the Canadian Prairies. Agricultural and Forest Meteorology. Vol. 152, Pages 223-232. [10] 35 for and explain these variations. For example, these forecasts are helpful in set- ting insurance premiums and futures prices as well as in managing grain transport. Since spatial and temporal climate variability affect crop yields ([84]; [72]), a crop yield forecasting method must include climate as an essential component if it is to be successful. Several process–based models have been successfully used for crop yield pre- diction including the Agricultural Production Systems Simulator (APSIM) in Aus- tralia ([51]) as well as a web-based tool developed by the United States’ Southeast Climate Consortium ([49]). These process–based models typically employ tunable and user adjustable deterministic and stochastic models to simulate biological and physical processes related to crop yield. While these models use knowledge per- taining to the individual processes, they often require significant input from the user, including a wide range of meteorological and environmental variables which may be difficult or expensive to obtain. In contrast to the above, traditional statistical techniques are purely empiri- cal. While these methods may result in accurate predictions, they typically lack the interpretability of process–based models ([6]). As a result of this criticism, recent years have seen the development of statistical models that also provide in- terpretation of the underlying biophysical process (see, for example, [83], [41]). One such process knowledge-based approach involves water stress indices ([71], [72]; [73], [74]), the result of which has been of tremendous use and benefit to stakeholders, allowing for prediction and understanding of crop yield anomalies. While these models have improved the prediction of crop yield, there exists scope for improvement through a) providing an efficient dimension reduction of explana- tory variables; b) accounting for uncertainty in the estimated technology trend; c) 36 modelling spatial correlation between regions. This chapter describes the results of a project coordinated by Agriculture and Agri-foods Canada to develop a model that explains and predicts wheat yield and its relation to climatic variables. With plans for an online implementation in the future, efficiency was required as a feature of the model, as was the ability to stabi- lize the effects of noisy measurements. Building on earlier work, we employ a crop water stress index (SI) to provide explanatory power for a new crop yield predictor ([25]). To improve prediction over existing approaches, we extract a sensitive yet low-dimensional summary of this stress index, comparing various alternatives and bases before ultimately selecting principal components. We then demonstrate its improved prediction performance compared to currently used windowed average approaches. In contrast to previous work which models each agricultural region separately, we create a unified model that allows strength to be borrowed from adjacent and nearby regions, thus stabilizing both inference and prediction. By employing a spatially-motivated context-specific prior distribution on the param- eters of interest, we account for and use spatial correlation between sites while smoothing and consequently improving predictions. Following this introduction, Section 2 describes the crop yield forecasting prob- lem and available data. This section works through a series of successively im- proved models, eventually leading to a Bayesian model in Section 3 which jointly models all regions simultaneously. Model testing and diagnostics are explored in Section 4. Lastly, Section 5 concludes the work. 37 4.2 Materials and Methods This chapter models crop yield in the Canadian Prairies as a function of climate- related explanatory variables. The data include annual wheat yields (in bushels per acre) along with associated measurements of a crop water stress index and growing degree day (both described later) for 40 agricultural regions (plotted in Figure 4.1) across the Canadian Prairies from 1976-2006. The agricultural regions are those used in the 2006 Canadian Census of Agriculture, through which the data are also obtained, and are determined from climate and soil information. For each of the 31 years and 40 regions, yield is an aggregated average across the region. Likewise, stress index and growing degree day are calculated regionally, but on a daily basis throughout the growing season (April 1 to September 30). 4.2.1 Incorporating soil water The well recognized influence of soil water on crop yields dictates its inclusion in any yield prediction model ([25]). However, due to the time-consuming and costly process of measuring soil water content, in practice its effects must be in- ferred from more widely available environmental variables such as precipitation, temperature, and easily measured crop and soil-related factors. A suite of mod- els have been developed which attempt to understand soil water availability in the context of these environmental variables. Beginning with simple water balance ap- proaches that balance precipitation and soil water storage with evapotranspiration and water runoff, these models have increased in their complexity over the years ([25, 85]). For the reasons given below we focus on budget models, which build on the premise that above a certain threshold (called the ‘field capacity’), soil cannot absorb any more water and therefore any additional water is drained off through 38 runoff or drainage. Also, if the soil water fails to be replenished through precipita- tion, irrigation, or other sources, the soil reaches a point where plant roots are no longer capable of uptaking water. This stage is known as the ‘wilting point’. Evapotranspiration, which describes the sum of evaporation and plant transpi- ration, measures the water lost from plants, soil, and other land surfaces into the atmosphere. There are two key components in the budget model, potential evap- otranspiration (PET) and actual evapotranspiration (AET). PET represents the at- mospheric demand for evapotranspiration; specifically, it accounts for the energy available to evaporate water and transport it into the lower atmosphere. AET is the actual water content available for evaporation and transpiration, and relies on plant physiology and soil characteristics for its calculation. When the soil has ample water, the actual evapotranspiration (AET) can equal the PET. However when the soil is not at its field capacity, AET will be less than PET. More details on these concepts and soil science in general may be found in [16]. Budget models are straightforward to implement since they require a minimum of meteorological data as well as soil field capacities and wilting points. While more advanced models have been built which include soil hydraulic characteristics and more complex relationships between soil, plant, and meteorological systems, these models require considerably more information from the user, including de- tailed soil and plant characteristics. Because of the additional variables required by these models, we employ a budget model in the remainder of this work. Our model uses crop water stress index (SI) over agricultural land, defined as 1-AET/PET ([73, 74]). This quantity will be near 0 when water is plentiful in the soil and near 1 when the plant is stressed by a lack of available moisture. Intuition might sug- gest directly including precipitation, temperature, soil and plant information into 39 the model. However, doing so would add a large number of variables, especially considering that many of these variables are observed for every day of the grow- ing season. Using the SI instead provides an economical reduction in the dimen- sionality of the description space in a way that respects the biophysical processes involved in soil water movement and availability. Predicting yield with SI We begin by detailing the process of fitting a regression model to crop yield using least squares (LS). First let y j,t , j = 1, . . . ,40 be the yield from region j for years t = 1976, . . . ,2006. Since SI is a daily value, we create an annual average for each year and region; let si j,t denote the vector of these means in year t for each region j. We begin by fitting a common regression model to all regions, specifically y j,t = β0+β1t+β2si j,t + ε j,t . (4.1) Here ε j,t for year t and region j represents a combination of model and measure- ment error. While previously developed statistical models for crop yield account for a technology trend by first fitting a regression on time and then modelling the residuals, such approaches yield little understanding about the uncertainty associ- ated with forecasting. In particular, while forecasts that use detrended data may be similar, their associated variances will be biased as uncertainty in the trend model is ignored. In fact, to properly account for all sources of variability the technology trend should be an integral part of any forecasting model. To begin, note that the simple model in Equation 4.1 relies on only 3 parameters – all regions are described by the same equation. The validity of inference for 40 Figure 4.1: Mean residuals from model (1). We observe that the model resid- uals are spatially correlated. such a model relies on assumptions including for instance that the errors ε j,t are stochastically independent for all j, t. To test this assumption, we plot the mean residual (averaged over the 31 years) for each of the 40 stations in Figure 4.1. This figure makes it clear that the residuals are spatially correlated. For instance, the residuals in Alberta (the western-most Prairie Province) are much larger than the other two provinces, highlighting the fact that the model is biased, particularly in central Alberta. Considering the mean and standard deviation of crop yield across the prairies are 30.9 and 8.2 respectively, the average residual value of 13.5 in this region indicates that the model is consistently underestimating the crop yield there. To gain descriptive power, researchers have expanded the above model by fit- ting a different regression model to each region, specifically y j,t = β0, j+β1, jt+β2, jsi j,t + ε j,t . (4.2) 41 The expanded model now accounts for 61% of crop yield variation, compared to 33% for (1), albeit at the expense of additional parameters in its mean structure. In fact, by assigning a unique parameter to each region, this expanded model has 3× 40 = 120 parameters. By using such models, albeit with potentially modi- fied/additional explanatory variables, several authors have been able to create fairly accurate predictions of crop yield ([71]; [74]). It is important to note that the large number of predictor variables (120) makes this model prone to overfitting; while some authors have used cross-validation to prevent this (i.e. [74]), others have sought to further improve model fit by conducting extensive calibration to tune the explanatory variables (i.e. [71]). It is well understood that smoothed, or penal- ized, models have better prediction properties than larger, more variable models ([43]). This leads us to prefer the most parsimonious model yielding accurate fore- casts and to select explanatory variables which provide optimal prediction power for crop yield. While earlier models have been examined with regards to their model fit (as measured through R2), a much preferred metric is model prediction performance (as measured through cross-validation). While the availability of SIs for every day of the growing season (in our case April 1 to September 30) means its vector of measured values is of very large dimension, good modelling practice requires that this dimension be reduced before introducing the vector into the regression model. At one extreme, we could do what we did previously, and use just the mean of these daily SI values over the growing season, a one-dimensional feature, as our explanatory variable. However, that would oversimplify the SI’s role, since plant growth is influenced more at certain times than others during the growing season. As an extreme example, if the crop is harvested in early September, the SI values in late September would 42 Figure 4.2: Correlation of SI and yield over time. Correlations smoothed with Lowess smoothing. From this we see that SI is most correlated with yield in an intermediate part of the season, namely days 80 through 160. aid little in predicting crop yield. To find a low-dimensional feature that provides good predictive power for crop yield, we could average over a reduced window, that is, exclude SI values early and late in the season ([74]). This reflects the point just made that SIs early and late in the season may not be correlated with crop yield. Figure 4.2 shows this correlation between SI and crop yield for each day in the summer, organized by province. This figure suggests we average over days 80 through 160, rather than the entire growing season. However, this produces only a modest improvement, 60.72% of crop yield’s variability now being explained instead of 60.56% using the average over the entire season as before. This plot also reveals large spatial variability, particularly between provinces. We explore this issue in more detail later. There exists considerable scope for tuning this window; for instance [71] select 43 unique window start and end points for each region to achieve an excellent fit – over 75% of variation explained. However such tuning entails much attention to detail. On top of the upper and lower limits for the averaging to take place, [71] calibrate potential available soil water capacities, the maximum number of sowings and the rainfall amount triggering planting in each region. In other words, in addition to the corresponding regression coefficients, this tuning in effect adds 5 additional parameters per region, which in our case would increase the number of parameters being fitted in (2) from 120 to 320, leading most likely to serious over-fitting when considering that such an approach still uses an average SI over the growing season, not accounting for temporally-varying impact of SI. To quote John von Neumann: “With four parameters I can fit an elephant and with five I can make him wiggle his trunk.” As such, a preferred alternative would be a lower dimensional feature which cap- tures the key components of the stress index. In addition, we would like to include information which allows for the impact of SI on yield to vary over the growing season. To capture more information from the SI values than would be available from simple averaging, we extract the principal components and hence main sources of variation from the stress index. To be more precise, after subtracting the average SI from each day, the first principal component is the linear combination of growing season SI values that accounts for the most variability in the SI values. The second, which is orthogonal to the first, explains the next largest amount of variation, and so on. Each observation, in this case each region – year combination, also has a set of loadings that, when multiplied by the corresponding principal components, 44 Figure 4.3: Screeplot showing the principal component variances in decreas- ing order. From this, we see that the first 4 principal components explain much of the variation in SI. return the original observation. Figure 4.4 shows the subtracted mean process as well as the first four principal components that together show the SIs history over the growing season of our study. Four principal components are chosen as a result of examining a screeplot of the principal components, see Figure 4.3. Figure 4.4(a) reveals firstly the primary shape of the stress index, showing that initially – from April 1 – the stress is moderate, increasing until May, followed by a gradual decline until it bottoms out in July. It then returns to its highest values by the end of September. The first component (Figure 4.4(b)), which describes 46.9% of the variation in SI, captures a valley in the SI cycle around late August. The second component, which accounts for 14.6% of the variation, shows SI’s decline into its July valley followed by its rise to its early September peak. The orthog- 45 Figure 4.4: Principal components and mean for SI. This figure depicts the major patterns in the variation of the stress index (unitless) over the growing season. Observe how the first four principal components pick up deviations from the overall pattern in (a), and reveal the peaks and valleys of the stress cycle over the course of the summer induced by things like patterns in precipitation and temperature. Together these four components capture most of the variation in stress in a very economical way and eliminate the need for the high dimensional vector of daily SI values. 0 50 100 150 0. 2 0. 3 0. 4 0. 5 0. 6 Principal Component Mean Time Starting April 1 St re ss  In de x (a) 0 50 100 150 − 0. 2 − 0. 1 0. 0 0. 1 First 4 Principal Components Time Starting April 1 St re ss  In de x PC1 PC2 PC3 PC4 (b) onality of the first two components is apparent from (b). The third component of SI’s variation captures its low April start. Altogether, the first 4 principal compo- nents account for 78.5% of the variation in SI over the growing season. Beyond 4 principal components we observed little in terms of improved modelling fit and a reduction in prediction performance, as shown in figures 4.5 and 4.6. Thus by including the loadings for these 4 principal components as explanatory variables, we have created a 4-dimensional feature which accounts for a large proportion of variation in the stress index. Note that the first SI principal components aren’t necessarily the best predictors 46 of yield. However, LASSO – a penalized least squares variable selection method – in fact selects these same four principal components as the best four ([43]). This choice of feature also has a natural biophysical interpretation. For instance, a large and positive regression coefficient for the loadings corresponding to principal com- ponent 3 would imply that a reduction in stress in early April is highly connected with increased crop yield. By using this approach, the explained variance of the re- gression model increases from 60.56% from averaging SI over the growing season to 70.06%. In addition, as discussed above, the inclusion of principal components allows the user to gain intuition about the effect different seasonal patterns in the stress index will have on crop yield in a way averaging across the season can not. Using these principal component loadings, our new model is y j,t = β0, j+β1, jt+β2, jPC1 j,t + · · ·+β5, jPC4 j,t + ε j,t . (4.3) where PC1 j,t indicates the loading for principal component 1 in region j and year t. Alternative bases and levels of sparsity Because of their widely documented ability to model complex nonlinear signals while maintaining sparsity, we briefly explore wavelet bases as an alternative to principal components ([58]). Specifically, we examine a variety of different wavelet bases and levels of sparsity both in terms of cross-validated prediction error as well as R2. Because SI is inherently piece-wise smooth due to the impact of pre- cipitation resulting in changepoints in the SI values, wavelets provide a valuable alternative to principle components. Figure 4.5 plots R2 of the yield model for 47 Figure 4.5: R2 of the crop yield model for a range of bases and sparsity levels. From this we notice that principal components (PCA) provide better model fit for all sparsity levels. 5 10 15 20 0.6 0.7 0.8 0.9 Fit of PCA and Wavelet Methods Sparsity (number of wavelets or principal components) R2 PCA Daubechies 4 Daubechies 8 Daubechies 16 Daubechies 20 Least Asymmetric 8 Least Asymmetric 20 Best Localized 14 Coiflet 6 Coiflet 30 various bases and levels of sparsity. From this plot we see that principal compo- nents dominate in terms of model fit. While R2 measures how well a model fits to data, it is not a good indicator of a model’s prediction abilities. As such, Figure 4.6 plots cross-validation root mean squared error (in bushels per acre) for each basis and sparsity level. Once again we observe that principal components outper- form wavelets. From these figures we conclude that principal components lead to a model with better fit and prediction performance. This example highlights the need to be selective in the choice of basis to represent stress index and other variables in such a model. While wavelets excel at representing piece-wise smooth models in a very sparse way (requiring the storage of only 1 vector – the mother wavelet – as well as a series of indices), this is also their downfall in some circumstance such as this one which require a richer representation. 48 Figure 4.6: Cross-validation RMSE (bushels per acre) of the crop yield model for a range of bases and sparsity levels. From this we notice that principal components (PCA) provide better prediction than the wavelet bases for all sparsity levels. 5 10 15 20 6 8 10 12 14 16 CV RMSE of PCA and Wavelet Methods Sparsity (number of wavelets or principal components) CV  R MS E PCA Daubechies 4 Daubechies 8 Daubechies 16 Daubechies 20 Least Asymmetric 8 Least Asymmetric 20 Best Localized 14 Coiflet 6 Coiflet 30 4.2.2 Incorporating temperature Temperature affects a plant’s development and growth in a variety of ways, in particular its photosynthesis and respiration. In general, temperature affects plant functioning through its action on enzymatic reactions. At low temperatures, en- zyme proteins are not sufficiently flexible to complete the conformation necessary for enzymatic reaction. Conversely, high temperatures can coagulate the enzyme leading to similar barriers to the reaction. Alongside a minimum and maximum temperature to allow growth, most plants have an optimum temperature to encour- age growth. For instance, [77] conclude that the minimum and optimum tem- peratures for wheat are respectively 0 and 20-25 degrees celsius. As a result of temperature’s influence on plant development, we suspect that its inclusion into the model will result in prediction performance gains. In addition, by directly in- 49 cluding temperature effects interpreting impacts of climate on yield is made more straightforward. Growing degree day While temperature could go directly into the model, its measurement in hourly or smaller increments creates a considerable amount of data. As a result, some dimension reduction is needed to limit the number of explanatory variables. One could do this using just the maximum and minimum daily temperatures or better still, a one dimensional summary that combines the two. Thus ‘growing degree day’ (GDD) measures the heat accumulation in a region based on local weather by taking an average of the daily minimum and maximum and subtracting a base temperature as follows: GDD= max ( 0, Tmax+Tmin 2 −Tbase ) . Thus the GDD measures the daily average temperature but in a way that reflects the extremes more sensitively. The base temperature represents the physiological temperature below which development would be zero. A day with a high and low of 30 and 15 degrees celsius and a base temperature of 10 degrees would have a GDD value of 12.5 degrees celsius. Thus GDD is a simple, single-dimensional summary for describing the plant’s exposure to heat. While GDD is a simple heuristic, it is commonly used by horticulturists to estimate the stages of a plant’s growth. As an example, the maturation of wheat corresponds to about 1600 GDDs ([28]). Thus GDD provides us with a simple low-dimension summary of temperature, allowing for comparison of the thermal time available in 50 Figure 4.7: Correlation of GDD and Yield over Time. Smoothed with lowess smoothing. From this we see that a reduced window average may be appropriate. different climatic zones. While SI gives scientific insight into the moisture available for plant growth, it says little directly about the heat available to the crop. Thus to improve our model we can also include GDD, which up until now has been used primarily in this context for tuning the explanatory variables ([72]). Like SI, GDD is a daily value, and hence can be treated similarly. Thus through the correlations plotted in Figure 4.7 we look at the time of season where GDD is most correlated with yield. This figure tells us that an appropriate window would be the one bounded by days 50 through 160. Using a cumulative average over the whole season, the explained variation in yield increases from 70.06% to 73.20%, with the shortened window performing similarly. As emphasized by others (see, for instance, [72]), includ- ing GDD accounts for the biophysical phenology of the crop as well as improves 51 interpretability of the model. Hence while the prediction improvements are mini- mal, the variable’s inclusion is an important step in the development of a crop yield model. In addition to averaging over the whole season or a shorter window, we can also use principal components as we did for SI above. While using the first 4 prin- cipal components only increases this to 76.48%, the additional 120 variables result in reduced cross-validation prediction performance, hence we prefer using just the windowed average. We emphasize that this is a user and case-specific choice, and for alternate purposes different choices may be preferred. The expanded LS model 4.3 then becomes y j,t = β0, j+β1, jt+β2, jPC1 j,t + · · ·+β5, jPC4 j,t +β6, jGDD j,t + ε j,t , (4.4) where GDD j,t is the windowed average of GDD in region j, year t. It is worth noting that temperature is a component of SI; however, the addition of GDD into the model improves both model fit and prediction, thereby eliminating concerns about the deleterious effects that collinearity between the two covariates might introduce. We compare the previous models as well as those developed later in the chapter in Table 4.1, showing the features and performance of each successive model. The traditional regression models represented in Table 4.1, fitted for each re- gion separately, ignore a considerable amount of information. Specifically, because of the close spatial proximity of the regions, considerable strength may be gained by exploiting the correlation among regions. For instance, use of neighbouring SI values can help stabilize predictions based on SI values, since the latter come from a small set of regional monitoring stations and hence can be fairly noisy. The 52 Table 4.1: Features of various models. We see that while model 4 has the best fit to the data (R2 = .73), the Bayesian model gives the best predic- tion performance in terms of cross-validated root mean squared error (in bushels per acre). Effective parameters is defined as tr(S), where ŷ= Sy, and may be cosidered a measure of model complexity ([43]). Model Parameters Effective Parameters R2 CV RMSE 1: Single LS 3 3 .33 6.83 2: LS (SI) 120 120 .61 5.79 3: LS (PCA) 240 240 .70 5.72 4: LS (PCA+GDD) 280 280 .73 5.69 5: Bayes 283 139 .70 5.35 amount of borrowed strength can be considerable when the correlation between stations is high. In addition, modelling all stations jointly while incorporating spa- tial information allows us to continue to make predictions even in the presence of missing or noisy data. If a measuring station goes out of operation temporarily, its missing values may be inferred from data collected at nearby regions to yield accurate forecasts. This idea leads into our next section, which focuses on spatial models that look at all regions together in a unified manner. 4.2.3 A context-specific spatial Bayesian approach Classical regression methods rely on the assumption that their model residuals are uncorrelated. Indeed violation of that assumption can have very serious delete- rious effects on parameter estimates compared, for example, to violations of the assumption that those residuals have a Gaussian distribution ([24]). In our case the residuals are most certainly spatially dependent and thus the actual amount of in- formation in the data can be much less than the assumptions underlying (1) would suggest. The unwary analyst would then be led to make overconfident forecasts 53 with parameter estimates which vary considerably from one region to the next, yet have unduly small standard errors. One work-around would model the regions separately. However, this wastes the benefits spatial dependence provides by borrowing strength, telegraphing in- formation across the regions through the wires of correlation for the mutual im- provement of all their forecasts. This progression naturally leads us to a Bayesian framework for handling this problem, one which jointly models all regions simulta- neously while accounting for their spatial dependence. Thus we move from the fre- quency paradigm of classical statistics to the Bayesian paradigm of modern statis- tics. These two paradigms, which tend to give similar inferences at least for fairly large datasets, are very different in concept. Frequentists see data as being gener- ated by a system governed by some true but unknown parameters. They commonly seek to estimate these true parameters well in some sense, for a variety of inferen- tial purposes such as forecasting. The central tenet of their theory is repeated sam- pling – in the long run the parameters can be estimated to arbitrarily high levels of precision if the system producing the data were unperturbed. However, Bayesian statisticians reject the notion of repeated sampling as a fundamental construct in their theory, recognizing realistically that most systems cannot remain unperturbed and pump out replicate data over an extended sequence of trials. Although their models involve uncertain parameters, these parameters like all uncertain objects such as future data values, are characterized by a probability distribution. Initially that distribution, called a prior, simply reflects the Bayesian’s own knowledge. An abundance of such knowledge would mean a prior concentrated around a single point and a state of near certainty. The information in the data adds to the state 54 of knowledge through the celebrated Bayes theorem. The latter relies on the like- lihood function of the uncertain parameters which captures all the information in the data. A likelihood tightly concentrated around a single value would mean the data has eliminated much of the uncertainty about the parameters. However gen- erally, Bayes rule needs to be applied to get the combined effect of data and prior knowledge; this yields the Bayesian’s updated prior, or the so-called posterior dis- tribution. Due to its adaptability and ease of use, Bayesian inference has become a prominent fixture in modern spatial statistics, and in particular the modelling of random spatio-temporal fields ([5]; [56]). Available prior information Consider, for example, the spatial structure discussed above. Even before estimat- ing the parameters in equation 4.3, we expect parameters in adjacent regions to be similar. Thus we would be surprised if the parameters relating GDD to yield had completely opposite signs in two neighbouring regions. This reflects our prior beliefs about those parameters, namely that knowledge of one would tell us some- thing about the other. More simply, we would see them as stochastically dependent in the language of the probability distribution that characterizes our beliefs about them. We might even have some idea of their approximate magnitudes. For in- stance, a magnitude of 100 (bushels per acre/degree celsius) for the coefficients β6, j for GDD would be completely untenable, since it would mean that chang- ing one cold day to a warm one (adding, say, 10 GDD over the entire cumulative season), would increase the yield by roughly 10 bushels per acre. Thus even with- out formalizing our beliefs in a prior distribution, loose bounds on parameters are almost always apparent. 55 Application of the Bayesian approach starts by characterizing our beliefs about the parameters in the form of a prior distribution. In the regression models intro- duced above, this would amount to a joint prior distribution on each β to account for our belief in their dependence (similarity) for adjoining regions. For simplic- ity, stack all of the coefficients into a vector β , the first 7 coefficients being for all variables in region 1, the next 7 for region 2, and so on. Assuming a Gaussian distribution as a convenient prior form, we can explicitly write the prior as follows: β ∼ N (0,Σ0⊗gΩ) . (4.5) By using such a Kronecker structure, Σ0 models the correlation within a given coefficient across space, while gΩ corresponds to Zellner’s g-prior ([89]) with Ω the 7×7 empirical covariance between explanatory variables. As such, this choice of prior fits within the empirical Bayes paradigm. While specifying the coefficients (particularly the intercept) to have zero-mean seems restrictive, we note that a priori it does not seem unreasonable to assume that in the presence of full stress (signaling the complete absence of water) the crop yield would be zero. We now specify Σ0, the correlation between regions, as Σ0 = exp(−D/φ), (4.6) with a slight abuse of notation where D is the matrix with element (i, j) the Eu- clidean distance between regions i and j (as measured from the center of the re- gion). Here φ is a parameter controlling the rate of decay of correlation as distance increases. In this way, φ controls how spatially smooth the coefficients are, while 56 g controls how tight around zero the coefficients are. While we are not convinced that wheat is planted in more than the southern section of the three northern-most regions, including the region centers rather than some more southerly geographic location is conservative, as the true geographic center is further removed from adja- cent regions, and therefore the correlation expressed through Σ0 is decreased. Oth- ers have proposed more complex spatially-varying models which rely on Markov chain Monte Carlo for inference (e.g. [5, 35]); however, our goal of an online implementation restricts us to models with analytic tractability. While we suspect neighbouring regions to be similar, Figure 4.2 highlights the differences between provinces. While one could include an indicator variable to allow for provincial effects, accounting for provincially-varying coefficients would involve a considerable number of interaction terms. In fact, the varying irriga- tion and technology policies in each province result in a sharp boundary between provinces for several of the mean parameters. As such, it is not entirely logical to use a stationary prior ([20]) which assigns correlation between regions solely based on distance without any respect for political boundaries. As a result, we adjust our prior distribution to have reduced correlation between regions in dif- ferent provinces. While the obvious approach is to scale down the prior correla- tion between regions in different provinces with a constant value, this may lead to non-positive definiteness of Σ0; alternative methods which do not suffer from this problem are therefore needed. We accomplish this task by deforming the physical space, in effect pushing neighbouring provinces apart. Motivated by [78], this ar- tificial distortion of the space results in a stationary prior in the deformed space, yet a nonstationary one in the original space. The distance d (measured in degrees latitude/longitude) by which the provinces are pushed apart in the artificial space 57 is selected through cross-validation. Searching over the integers from 1 to 10, we find d = 4 to give the best prediction performance (CV RMSE of 5.35 vs 5.39 for d = 0), intuitively meaning that Alberta and Manitoba are pushed respectively west and east from Saskatchewan by 4 degrees longitude in the artificial space. The end result is a reduction in the off-diagonal elements of Σ0 corresponding to between-province regions while maintaining positive definiteness. Note that the prior parameters φ , g, and d result in an additional 3 parameters in the Bayesian model, as reflected in Table 4.1. Likelihood and posterior distributions We begin by employing the likelihood corresponding to (4.4), namely y j,t ∼ N ( β0, j+β1, jt+β2, jPC1 j,t + · · ·+β5, jPC4 j,t +β6, jGDD j,t ,σ2 ) . (4.7) In keeping with common practice, we also assign an Inverse-Gamma prior distribu- tion on σ2 with shape and scale parameters a and b set to be highly noniformative. Before proceeding, we introduce the notation y, the column vector of stacked y j, and X , the (31×40)×240 block-diagonal matrix of explanatory variables. Using Bayes theorem to combine our initial knowledge (in the form of prior distributions) and the information provided by the data (in the form of the likelihood), we can obtain the posterior distribution of the parameters. Specifically, for the regression coefficients β , the marginal posterior is obtained using Bayes Theorem as follows: pi(β |X ,y) ∝ ∫ pi(y|X ,β ,Σ)pi(β |Σ)pi(Σ)dΣ. (4.8) 58 Due to the conjugate nature of the prior and likelihood, we are able to analytically complete this integral. The resulting distribution is a multivariate Student-T, β ∼ T (β f ,Ψ,n+2a) (4.9) where β f = (X TX+(Σ0⊗gΩ)−1)−1(XT y) Ψ= (XTX+(Σ0⊗gΩ)−1)(SS+2b)/(n+2a) SS= yT y−β Tf (XTX+(Σ0⊗gΩ)−1)β Tf . From this last expression, we get the posterior mean β f , which may be used as a simple estimator for β . In fact, comparing β f = (XTX +(Σ0⊗ gΩ)−1)−1(XT y) to the LS estimate (XTX)−1(XT y), we readily see how the prior covariance affects the parameter estimates. In particular, a diffuse prior distribution adjusts the es- timate little, whereas an informative prior distribution – one that is fairly tightly concentrated around zero – shrinks the posterior estimate considerably. Setting g= 10 and φ = 106, we obtain coefficient estimates as shown in Figure 4.8, which also shows the corresponding least squares estimate using (4). We see that the spatial information used in the Bayesian model causes the coefficients to be more correlated across space. In addition, the zero-mean prior distribution leads to some shrinkage in the coefficient estimates. Interestingly, we notice little shrinkage in the estimated coefficient for technology trend, suggesting that the data contains considerable information on this quantity. 59 Figure 4.8: Coefficient surfaces for intercept, technology trend, and PC1. Other coefficients are similarly smoothed. (a) (b) (c) (d) (e) (f) 60 4.3 Results We proceed by comparing the prediction performance of the least squares and Bayesian methods. To accomplish this we use leave-one-out cross-validation, re- moving years one at a time in succession to compare each model’s predictive abil- ity. More specifically, we successively remove each year in turn, using the remain- ing years to find the posterior mean, notated β̂ i if year i is removed. This posterior mean is then used to perform prediction on the removed year. From this the root mean squared error (RMSE) is calculated as the square root of the sum of squared prediction errors for each year and region. RMSE = √√√√ 31∑ i=1 40 ∑ j=1 (yi, j−Xi, jβ̂ i)2/(31×40). (4.10) Figure 4.9 shows the cross-validation root mean squared error (RMSE) of the pos- terior mean estimate for various settings of g and φ . As g→ ∞ and φ → 0, the Bayesian model converges to the least squares solution, as evidenced by converg- ing cross-validation errors. However, if g is too small, the prior on the regres- sion coefficients is too informative towards zero, and hence the resulting posterior means are overly shrunken, resulting in poor prediction (RMSE > 6). While one could assign prior distributions to these parameters, we prefer finding them through cross-validation for computational efficiency. Specifically, given the optimal pa- rameters, the model is conjugate, and hence sequential updating and prediction is analytic and therefore nearly instant. It is very interesting to note that the optimal prediction error for the Bayesian model is less than for the least squares model, indicating that prediction is improved with regularization (provided by the zero- 61 Figure 4.9: Cross-validation errors using Bayesian posterior mean. For com- parison, the least squares error is 5.69. From this, we observe a ridge of excellent prediction. Hence there is some tradeoff between the two parameters to be tuned. mean prior and/or correlation). The area of lowest prediction error occurs along a diagonal of g and φ and has value approximately 5.35. This is likely due to the fact that an increase in g results in a more diffuse posterior which regularizes less, while increases in φ result in increased correlation between regions and hence more reg- ularization. Hence the optimal prediction seems to occur for moderate amounts of regularization. The cross-validation RMSE can also be calculated for each region by summing only over years. In this way we can gain an improved perspective on the model’s prediction performance. However, while cross-validation RMSE gives an idea of the prediction performance of a model, it does little to tell of a model’s bias. To 62 do this we decompose the RMSE into the model’s prediction bias and variance. Doing this for each region, we obtain Figure 4.10 detailing the prediction RMSE, bias, and variance of the Bayesian and LS models in each region. From this figure we observe that, with the exception of one or two individual regions, the Bayesian model improves RMSE in all areas except for southern Manitoba. Digging deeper, we see a negative bias in this area. Thus the regularization of the model is perhaps not useful in this region due to some systematic differences in this area. Specifi- cally, this section of southern Manitoba is known to use significant irrigation ([34]). As a result, further model development might be explored in this area to account for irrigation. 4.4 Conclusion In this chapter we have examined the role of SI in predicting crop yields, empha- sizing the need to create a judicious low-dimensional summary in order to improve prediction. Simply averaging SI over the entire season is inefficient, as yield may be insensitive to stress in certain parts of the summer. The traditional solution to this problem is to average over a reduced window of data, hence cutting out those areas lacking in sensitivity from the analysis. However, this one dimensional fea- ture is not particularly sensitive to changes in stress indices within that window. For example, a region which has low SI in June but high SI in July might ultimately have the same averaged value as another region which had just the opposite trend. To address this issue, we have implemented principal components analysis to cre- ate a set of flexible summary statistics which better describe the variations in SI, and as a result improve prediction considerably. We also demonstrated principal components’ improved performance over wavelet bases. 63 Figure 4.10: Cross-Validation results by region. The Bayesian model im- proves prediction by all standards in the majority of regions. We plot root mean squared error (RMSE) as well as the breakdown to bias and variance for the models. (a) (b) (c) (d) (e) (f) 64 We have also shown the importance of incorporating spatial correlation into crop yield models; ignoring this information can lead to bias both in model identi- fication and prediction. Specifically, we observed that a common least squares fit of crop yield on some explanatory variables over the entire region resulted in biased residual errors, and hence violated the assumptions of the model. To avoid this problem, we could fit each agricultural region with its own model. The problem, however, is that this ignores information between crop regions, and as such we ob- served reduced prediction power and model identifiability. We addressed this issue through the use of a Bayesian model which modelled all regions together, yet ac- counted for spatial correlation. This model smooths and stabilizes prediction and also allows for analytic and therefore efficient updating and prediction. In addi- tion, we created a non-stationary prior distribution to address the issue of province to province variability resulting from provincial differences in policy, management, and other factors affecting yield. Through cross-validation, we demonstrated this model to achieve improved prediction performance in modelling Canadian wheat yield over the least squares model which ignores spatial dependence, and hope that others will attempt to replicate our findings in other contexts based on the promise seen in this application. 65 Chapter 5 Nonstationary Modeling Through Dimension Expansion1 5.1 Introduction Recently there has been great interest in using spatial statistical methods to model environmental processes, with the aim of both gaining an improved understanding of underlying processes and making predictions at locations where measurements of a process are not available. The majority of such methods make the assump- tion that the underlying process is stationary ([20]) which, for many environmental systems, may be untenable. In this chapter, we focus on accurately explicating the nonstationary structure that often arises in measurements of atmospheric, agricultural, and other environ- mental systems. If these systems share one underlying theme it is complex spatial 1A version of Chapter 5 has been published. Bornn, L., Shaddick, G., Zidek, J. (2012) Mod- elling Nonstationary Processes Through Dimension Expansion. Journal of the American Statistical Association. Vol. 107, Pages 281-289. [15] 66 structures, being influenced by such features as topography, weather, and other en- vironmental factors. For example, the air quality characteristics of cities are likely to be more similar than that of rural areas irrespective of their geographic prox- imity. Ideally we might model these effects directly; however, information on the underlying causes is often not routinely available. Hence when modeling environ- mental systems there exists a need for a class of models that are more complex than those which rely on the assumption of stationarity. In the field of atmospheric science, empirical orthogonal functions have been used to model a nonstationary process as the sum of a stationary isotropic pro- cess and a set of basis functions with random coefficients representing departures from nonstationarity ([66, 67]). Current approaches to modeling nonstationary pro- cesses in the statistical literature broadly comprise those that (i) combine locally stationary processes to create an overall nonstationary process and (ii) build upon the framework of ‘image warping’. A number of approaches for handling nonstationarity assume that over small enough spatial domains, the effects of nonstationarity are negligible, and hence locally stationary models may be used. This concept is the basis of kernel ap- proaches, early examples of which can be found in [39, 40]. The process–convolution approach ([45, 46]) relies on the notion that a wide range of stationary Gaussian processes may be expressed as a kernel convolved with a Gaussian white noise process, with the kernel being allowed to vary spatially to account for nonstation- arity. The form of the kernel allows for a broad expression of potential covariance functions, with a Gaussian kernel corresponding to a Gaussian covariance function and other choices of kernel resulting in other correlation structures. Similarly, [32] suggested modeling the field as a weighted average of local stationary processes 67 within a set of regions, an idea which was later extended to include a continuous convolution of stationary processes ([33]). Various difficulties still remain in this class of models, including the lack of a complete and easily interpretable global model and the choice of local regions and details of the weight structures. An al- ternative approach proposed by [78] is that of “image warping”, the central idea of which is that a nonstationary process may be stationary in a deformed, or warped, version of geographic space. Multi-dimensional scaling (or related methods) can be used to find the deformed locations with a mapping between the original and deformed space found using, for example, a thin plate spline. The principal idea underlying the proposed method is that of embedding the original field in a space of higher dimension where it can be more straightfor- wardly described and modelled. Specifically, we shift the dimensionality of the problem from 2 or 3 dimensions to 4, 5, or more in order to recover stationarity in the process; we term our methodology “dimension expansion.” Our starting point is that nonstationary systems may be represented as low-dimensional projections of high-dimensional stationary systems (see, for example, [68]). The method is superficially similar to that of image warping; however, it differs fundamentally in that here the locations in the geographic space are retained, with added flexibility obtained through the extra dimensions. Additionally, it addresses one of the major issues with the image warping approach, namely folding of the space. This occurs in image warping if the estimate of the function that transforms from geographical to deformed space is not injective. As a result of folding, two geographically dis- tinct locations may be mapped to the same location, meaning the variation between them will be incorrectly treated as measurement error and small scale variation (i.e. the nugget), which is expressly appropriate only for collocated and other proximal 68 monitoring sites. In such cases, mapping quantities such as prediction intervals be- comes particularly challenging both in terms of implementation and interpretation. The remainder of the chapter is organised as follows: Section 2 introduces the dimension expansion framework proposed here, including an illustrative example to demonstrate the fundamental concepts behind the approach. This example is then used to draw comparisons to image warping. In Section 3, dimension expan- sion is applied to two real life examples. First, the solar radiation dataset originally used in [78] and used as a test-bed in various more recent image warping papers is studied. Second, we study air pollution from seventy-seven monitoring locations in the United Kingdom which show clear signs of nonstationarity. We highlight the ability of dimension expansion to accurately model such data as measured through cross-validated prediction error. Finally, Section 4 provides a discussion and sug- gestions for future developments. 5.2 Dimension Expansion While early work ([20]) dealt primarily with stationary models, it is now gener- ally recognized that many spatial processes {Y (x) : x ∈ S }, (S ∈ Rd) fail to satisfy this assumption. Environmental systems might exhibit behaviour that looks locally stationary, yet when considered over large and heterogenous domains they very often exhibit nonstationarity. For ease of notation, we consider Y (x) to be a (potentially nonstationary) mean-zero Gaussian process and place our emphasis on representing the nonstationary structure. A principal task in spatial statistics is estimating a variogram model (or correla- tion function) to explain spatial dependencies. The theoretical variogram, defined 69 as 2γ(xi,x j) = E (|Y (xi)−Y (x j)|2) is typically modeled using a parametric stationary variogram γφ (h) depending only on h = xi− x j , the difference vector between locations, and the parameter(s) φ . If the field is nonstationary, such a model will be a misspecification. In response, we transform the set of observed spatial locations S ∈Rd into one of higher dimen- sion S ′ ∈ Rd+p, where p > 0 and S is a subset of the dimensions of S ′. Put plainly, such an approach amounts to allowing extra dimensions for the observed locations x1, . . . ,xs , notated as z1, . . . ,zs such that the field Y ([x,z]) is stationary with a variogram model γφ ([xi,zi]− [x j ,z j ]). Here [x,z] is the concatenation of the dimensions x and z. Perrin and Meiring [68] explore this idea in the particular case where both the covariance function and the expansion from x to [x, z] are known. In their motivat- ing example, they consider the following stationary covariance on the plane: cov(Y ([xi,zi]),Y ([x j ,z j ])) = exp(−|xi− x j |− |zi− z j |). By restricting to the set z = x2 and defining Y ′(x) =Y ([x,x2]), the resulting covari- ance function on this reduced-dimension field is nonstationary, namely cov(Y ′(xi),Y ′(x j)) = exp(−|xi− x j |)exp(1+ |xi+ x j |). Perrin and Meiring [68] then consider the reverse problem, proving that a non- stationary random field indexed by Rd (with moments of order greater than 2) 70 can always be represented as second-order stationary in R2d . It is not, however, necessary to move from Rd to R2d to obtain the existence of a stationary field. Consider a recent result of Perrin and Schlather [69], which states that a Gaus- sian random vector can always be interpreted as a realization of a stationary field in R p, p ≥ 2, subject to moment constraints on the vector, namely that all com- ponents have equal expectation with the covariance matrix having identical com- ponents on the diagonal. From this it is straightforward to state that, similarly, a realization of a Gaussian process in Rd may be interpreted as a realization of a stationary field inRd+p, p≥ 2 (similarly, subject to moment constraints), with the covariance function ignoring the initial d dimensions. The above results show the existence of higher-dimensional stationary repre- sentations for nonstationary fields, yet in the vast majority of situations neither a nonstationary variogram, nor an analytic mapping to higher dimensions, is known. Here we construct a framework for using higher-dimensional representations to model nonstationary systems, with the goal of learning the latent dimensions non- parametrically from information contained within the data. To learn the expanded, or latent, dimensions z1, . . . ,zs we propose φ̂ ,Z = argmin φ ,Z ′ ∑ i< j (v∗i, j− γφ (di, j( [ X ,Z ′ ] )))2 (5.1) where v∗i j estimates the spatial dispersion between sites i and j, for example v∗i j = 1 |τ|∑τ |Y (xi)−Y (x j)|2, with τ > 1 indexing multiple observations of the system, the handling of which 71 is considered in the discussion, and di, j([X ,Z ]) is the i, jth element of the distance matrix of the (augmented) locations [X ,Z ]. Once the matrix Z ∈Rs×R p is found, a function f is built such that f (X ) ≈ Z . While a wide range of options exist, we follow Sampson and Guttorp [78] in using thin plate splines, here one for each dimension of Z . Thin plate splines are a mapping from one space to another such that the integral of the squared second order derivatives of the mapping function is minimized. For d = 2, this corresponds to minimizing s ∑ i=1 ||zi− f (xi)||2+λ2 ∫ R2 [( ∂ 2 f ∂ 2x1 )2 +2 ( ∂ 2 f ∂x1∂x2 )2 + ( ∂ 2 f ∂ 2x2 )2] dx1dx2, and therefore the smoothing parameter λ2 is analogous to λIW , the thin plate spline parameter in the image warping framework. Setting λ2 = 0 results in an interpo- lating spline, whereas λ2→ ∞ results in the linear least squares fit. The nonlinear functions f are therefore linear combinations of basis functions centered at the lo- cations S ∈ Rd . Once a model is built in the expanded space, f−1 will bring us from the manifold in Rd+p defined by (X , f (X )),X ∈Rd back to the original space. Due to our unique formulation, we have f−1(Z) = X , and we need not be con- cerned with the difficulties associated with ensuring that f is bijective as in earlier approaches. Thus we may view the originally observed locations X as a projection from a manifold within a higher dimensional space, [X ,Z ], in which the process is stationary. As an obvious (and direct) example, a process which is stationary given both geographical location and elevation may result in a nonstationary field given only longitude and latitude. Learning the latent dimensions (whether or not they have a physical meaning, such as elevation) means that a stationary model may be 72 used in the expanded space. In many situations, it is unclear how many additional dimensions are needed to accurately model the spatial field. One could use cross-validation or a model selection technique to determine the dimensionality of Z ; however, recognizing that (5.1) might result in overfitting the spatial dispersions, we would also like to regularize the estimation of Z . As a result, we modify (5.1) by including a group lasso penalty term on Z , where the groups are the dimensions of Z ([87]). The resulting objective function is φ̂ ,Z = argmin φ ,Z ′ ∑ i< j (v∗i, j− γφ (di, j( [ X ,Z ′ ] )))2+λ1 p ∑ k=1 ||Z ′·,k||1. (5.2) where Z ′·,k is the k-th column (dimension) of Z ′ . As a consequence of this revised objection function, one need only determine a maximum number of dimensions p and the shrinkage parameter λ1, whereupon the learned augmented dimensions Z will be both regularized towards zero and sparse in dimension. Hence λ1 can be viewed as regularizing the estimation of Z and determining the dimension of the problem, whereas λ2 controls the smoothness of the augmented dimensions; we suggest learning both through cross-validation, although other model fit diagnos- tics or prior information may be used as well. It is relatively straightforward to solve (5.2) using the gradient projection method of Kim et al. [52], which conducts block-wise updates for group lasso with general loss functions. Here the blocks are the dimensions of Z , and hence the optimization is efficient even for a large number of spatial locations. Optimization details are 73 given in the appendix. For ease of exposition we use an exponential variogram, γφ (x1,x2) = φ1(1− exp(−||x1− x2||/φ2))+φ3, which works well in the examples that follow, although the method applies analo- gously to other variograms. 5.2.1 Illustrative example We now present an illustrative example to help explain the concepts behind this proposed dimension expansion approach, as well as demonstrate the inability of image warping to handle complex nonstationarity. Specifically, we simulate a Gaussian process with s= 100 locations on a 3-dimensional half-ellipsoid centered at (0,0,0) such that the projection to the first 2 dimensions is a disk centered at the origin. Here, as throughout, distances are Euclidean. Figure 5.1 plots the empirical variograms for the original 3-dimensional space as well as the 2-dimensional pro- jection, the latter of which results in a highly noisy empirical variogram cloud. Our goal is to recover the lost dimension through dimension expansion by optimizing (5.2) with λ1 = 0.5, chosen to induce Z to have one dimension. Here we calculate the matrix of empirical dispersions v∗i j using 1000 realizations of the Gaussian pro- cess. Figure 5.2 shows the resulting learned locations as well as the corresponding empirical variogram, where we see that dimension expansion is capable of recover- ing the lost dimension, resulting in a variogram that closely reproduces the original. 74 Figure 5.1: Empirical variograms from the original process (left) as well as a 2-D projection (right) on the illustrative ellipsoid example. A fitted exponential variogram is shown by the solid line. Figure 5.2: Learned latent locations (left) using λ2 = 10−4 as well as corre- sponding empirical variogram (right) after dimension expansion is ap- plied. The fitted exponential variogram is shown by the solid line. 75 5.2.2 Image warping and folding In the image warping approach, Sampson and Guttorp [78] employ non-metric multidimensional scaling to move the locations along the geographic space, fol- lowed by fitting of the variogram γφ using traditional variogram fitting methods. From this, a function f is found to go from the original to the warped locations, and back via f−1. A number of adaptations of this approach have been proposed. [82] proposed modeling the covariance function as a linear combination of radial basis functions using maximum likelihood (as suggested by [59]). [63] and [64] noted that the multi-stage algorithm of Sampson and Guttorp does not correspond to a unified optimization problem and instead propose finding the locations and fitting the variogram using a single objective function, an approach also pursued by [62]. It is worth noting that [63] also explore mappings from R2 to R3 in the context of analyzing acid rain data, as the same-dimension mapping was inca- pable of describing the nonstationarity arising in the observed field. In a similar vein, [48] propose using simulated annealing to fit the spatial deformation model. Rather than imposing smoothness on the deformation through thin plate splines, they use Delauney triangulation to constrain the mapping f from folding on itself. In order to acknowledge the uncertainty associated with the deformed locations, [22], [79], and recently [80] have proposed Bayesian implementations of this ap- proach, the latter additionally using observed covariate information to warp into higher dimensions. As described in the introduction, the image warping framework can suffer from problems of folding, namely of f not being bijective (See Zidek et al. [90] for a particularly extreme example of folding). Considering the illustrative example of 76 Figure 5.3: Warped grid of locations (top) and corresponding variograms (bottom) for various settings of the thin plate spline parameter λIW using the image warping technique of Sampson and Guttorp [78]. Section 2.1, admittedly designed to be illustrative of such folding, we apply the image warping technique ([78]) with f modeled as a thin plate spline. Because the image warping framework contains no term similar to λ1 to regularize the warped locations, smoothing must be done through the thin plate spline parameter λIW (analogous to λ2 in the proposed dimension expansion framework). Figure 5.3 shows the warped grids and resulting empirical variograms for various settings of λIW applied to the ellipsoid example introduced in Section 2.1. We observe imme- diately that for a highly penalized f (corresponding to large λIW ) the space does not fold; however, the variogram fit is very poor. As λIW is relaxed to improve the fit, the space begins to fold, highlighting a potentially serious problem with the im- age warping framework – an issue which is addressed in the dimension expansion paradigm proposed here. Also related to the proposed dimension expansion method are latent space 77 models such as that proposed by Hoff et al. [47]. Here, latent dimensions are used to help learn a network of relationships between individuals. Recent work in the field of spatial statistics has also exploited latent dimensions to ensure valid cross- covariance functions in multivariate fields. Specifically, Apanasovich and Genton [2] use latent dimensions for the different variables in order to build a class of valid cross-covariance functions. 5.3 Applications We now present two applications of dimension expansion applied to the modeling of nonstationary processes using two real data sets. The first uses the solar radiation data ([44]) studied in the original image warping paper of Sampson and Guttorp [78]. The second consists of measurements from a network of air pollution (black smoke) monitoring sites in the UK, further details of which can be found in [29]. 5.3.1 Solar radiation The data of Hay [44] includes measurements of solar radiation from 12 stations in the area surrounding Vancouver. Due to the location and elevation of station 1 (Grouse mountain), the field is inherently nonstationary, as exhibited by the sam- ple variogram (Figure 5.4). This figure shows the original and warped locations using Sampson and Guttorp’s image warping approach with corresponding vari- ogram plot. Image warping moves the locations (in particular the station at Grouse mountain, which is the northernmost location) to achieve something closer to sta- tionarity. It is worth noting that overfitting may be controlled through the parameter λIW of the thin plate spline. Figures 5.4 and 5.5 show the results of applying the dimension expansion ap- 78 Figure 5.4: Original locations and empirical variogram for the solar ra- diation data (left); warped locations and associated empirical vari- ogram using image warping with λIW = 0.1 (centre); learned loca- tions with associated empirical variogram using dimension expansion with λ1 = 0.5,λ2 = 10−4 (right). The units for the semi-variance are (MJm−2day−1)2, and for distance are km (UTM coordinates, divided by 1000). The fitted variogram is shown by a solid line, and points as- sociated with Grouse mountain (station 1) are highlighted using an “X”. proach using λ1 = 0.5 and λ1 = 0.2, respectively, using a maximum number of dimensions of p = 5. The original locations are shown, as well as the added di- mensions (Z). With λ1 = 0.5 (Figure 5.4, right), dimension expansion adds one additional dimension which primarily serves to push Grouse mountain out of the plane, reflecting the a priori suggestion that it is elevation that leads to the station’s spurious correlation pattern. Interestingly, the contours of the learned dimension closely resemble the elevation contours of the mountains surrounding the Vancou- ver area. With λ1 = 0.2 (Figure 5.5), 2 extra dimensions are used, and the fit of the parametric variogram improves marginally. We can therefore see how λ1 in- 79 Figure 5.5: Dimension expansion of the solar radiation surface using λ1 = 0.2,λ2 = 10−4. Z here is 5 dimensions, with Z3,Z4, and Z5 set to zero as a result of the sparsity-inducing penalization. The first two panels show the learned locations, and the right-most panel shows the associ- ated empirical variogram (fitted variogram shown in red). The units for the semi-variance are (MJm−2day−1)2, and for distance are km. Points associated with Grouse mountain (station 1) highlighted using an “X”. fluences the number of extra dimensions used, as well as the shrinkage in each dimension, in order to control the level of fit. 5.3.2 Air pollution The data consists of annual average concentrations of black smoke (µgm−3) over a period of sixteen years from 77 locations within the UK operating between April 1978 and March 1993 (inclusive) and was obtained from the Great Britain air qual- ity archive ( Sites were selected in areas defined wholly or partially residential and measurements were aggregated to ward level (based on the 1991 census) using a geographical information system ([29]). The majority of wards contained a single site, but where there were more than one, records were either joined together if the time periods did not overlap or averaged if time peri- ods of operation were simultaneous. Due to similarities in levels of air pollution in urban locations, even if they are not geographically close, the field is known 80 to be nonstationary. Specifically, we see in Figure 5.8 reduced empirical disper- sions for distances around 280−290km (the distance between London and Liver- pool/Manchester), indicating that these urban centers are more similar than their distances would suggest. Our goal is to uncover and explore this nonstationarity through the dimension expansion framework. We begin with cross-validation to learn the optimal setting of the parameters λ1,λ2 using (5.2) as described in Section 2. Figure 5.6 shows the resulting cross- validation RMSE for various parameter settings. We can see that moderate values of both λ1 and λ2 result in the best prediction performance. As λ1 increases to its highest value (104.5), no dimension expansion occurs, and hence λ2 has no impact. From this it is straightforward to see that the use of the original geographic space is a special case of the dimension expansion framework. Using these parameter values, the dimensionally sparse optimization (5.2) used by dimension expansion leaves all but one dimension of Z set to zero. This dimen- sion is shown in Figure 5.7, where we see a strong ridge connecting London, Birm- ingham, Liverpool, and Manchester. Hence in the extra dimension major cities are moved closer together while rural areas are pushed further away. The variograms before and after the dimension expansion are shown in Figure 5.8, where we see no indications of nonstationarity after dimension expansion is applied. 5.4 Discussion By augmenting the dimensionality of the underlying geographic space, we have developed a highly flexible approach for handling the nonstationarity that often arises in environmental systems. While ostensibly similar to image warping, the proposed method avoids the issue of folding and allows one to model much more 81 2.5 3.0 3.5 4.0 4.5 8. 8 9. 0 9. 2 9. 4 9. 6 9. 8 CV Error from UK Black Smoke Data log10(λ1) C V  R M S E λ2 = 10 −5 λ2 = 10 −4 λ2 = 10 −3 λ2 = 10 −2 λ2 = 10 −1 Figure 5.6: Leave-10-out cross-validated prediction error of dimension ex- pansion applied to the UK black smoke data. Here we see optimal pre- diction for moderate values of both λ1 and λ2. complex nonstationarity patterns through interdimensional expansions, allowing the user to perform nonparametric learning of the mapping function. In addition, through the use of a group lasso penalty, we are able to estimate the number of augmented dimensions, as well as regularize the optimization problem. Lastly, we have highlighted the practical application of the dimension expansion approach in three examples, two of which use data from observed environmental processes. It 82 Figure 5.7: Coordinate surface of the learned dimension following dimension expansion. A strong ridge is visible connecting major cities indicat- ing closer correlation between these locations than would be suggested in geographic space. The locations of a selection of major cities are shown; (1) London, (2) Birmingham, (3) Manchester, (4) Liverpool and (5) Bristol. is worth noting that while we have developed the spatial model in terms of vari- ograms, it could alternatively be expressed in terms of covariances; see, for exam- ple, Gneiting et al. [38] for a thorough comparison of the two approaches. In general, models will comprise a spatial mean or trend term together with spa- tial covariance for deviations from this trend. It is desirable to maximally reflect the 83     0   100   200   300   400 50 10 0 15 0 20 0 Variogram (Observed Locations) Distance S em i-v ar ia nc e      0    500    1000   1500    2000 50 10 0 15 0 20 0 Variogram (Learned Locations) Distance S em i-v ar ia nc e Figure 5.8: Binned empirical and fitted (solid line) variograms on the UK black smoke data, following dimension expansion. In the original ge- ographic space, we see a dip in the empirical variogram at roughly 280km, corresponding to the distance between London and Liverpool/- Manchester. After dimension expansion is applied, the ridge between London and Liverpool/Manchester removes this effect of nonstationar- ity. The units for the semi-variance are (µgm−3)2 and for distance are km. variation in the response using the mean function and thus known covariates, but inevitably the mean function will not be able to capture all of the spatial variation and thus residual spatial variation must be modeled specifically. When all relevant covariates are included in the mean term, it is commonly assumed that the resulting spatial term is stationary. However, as the Karhunen-Loeve expansion shows, the modeling of spatial trend and covariance are inseparable and misspecification of the former will induce a second order distortion in the latter, thus violating any as- sumptions of stationarity in many cases. Due to the complexity of environmentally processes, misspecification is inevitable because all relevant covariates can never be known or, even if known, observed. In the air pollution example presented here, concentrations in cities appeared to be more similar than that of rural areas irrespective of their geographic proximity. If available, it would be possible to in- 84 corporate a measure of rurality in the mean function, possibly produced using a geographical information system based on population density data. However, even if such information were available, stationarity would still not be guaranteed and so there is a need for methods such as that proposed here to allow nonstationary models to be constructed for the spatial process. 5.5 Optimization of Equation (5.2) As with traditional multi-dimensional scaling, penalization functions of the form (5.1) do not have a unique maximum. However, the learned locations are unique up to rotation, scaling, and sign. The optimization problem (5.2) is more regular- ized, however, due to the presence of the l1-norm. Specifically, not all rotations and scalings of the learned locations will have the same l1-norm, and hence the resulting optimization is unique only up to sign and indices of zero/non-zero di- mensions. For example, consider finding the location of a single latent location in 2 latent dimensions. The locations (1,0) and ( 1/ √ 2,1 √ 2 ) fit the objection func- tion (1) equally well, yet their l1-norms are 1 and √ 2, respectively. Note that since the end goal is not to learn the dimensions, but rather to find an expanded space in which the process is stationary, the existence of multiple possible expansions is not important, so long as one of the projections in the equivalence class is found. In our experience, traditional optimization procedures such as Nelder-Mead or the BroydenFletcherGoldfarbShanno method ([65]) work well for a moderate amount of locations (s< 100) and dimensions (p< 3). For larger problems, it may be necessary to use purpose-built optimization routines intended for generalized group lasso. Let Ω(U ) be the first term in equation 5.2, where U = [X ,Z ]. Then 85 column k of the gradient matrix is ∇kΩ(U ) = 2 p Γ ◦ (U ·,k1p×p−1p×pU T·,k)1p×1 where Γi, j = ( γφ (di j(U ))−ν∗i j ) ∂γφ ∂di j . Using this gradient information, the gradient projection algorithm of Kim et al. [52] may be used to optimize (5.2). The algorithm proceeds as follows: Initialize : U 0 = 0, α : sufficiently small positive constant for t = 1, . . . ,T do Set u =U t−1−α∇Ω(U t−1) and η = {1, . . . , p} while M j > 0 ∀ j do For j = 1, . . . , p M j = I( j ∈ η)× ( ||u j||+ M−∑ j∈η ||u j|| |η | ) Set η = { j : M j > 0} end Set U t−1·, j = u j M j ||u j|| for j = 1, . . . , p end Return U T From this, one can alternate between optimizing the parameters of the vari- ogram and the latent locations. Further algorithmic details, such as the tuning of M and the setting of the algorithmic parameter α , can be found in Kim et al. [52]. 86 Chapter 6 Conclusion 6.1 Summary In this thesis, we have expanded the scientific base on latent correlation structures. Firstly, the proposed product graphical model prior improves flexibility in model- ing decomposable graphical models, borrowing strength from the immense litera- ture on product partition and related models. The product graphical model prior allows one to encourage (or discourage) clustering of the graphs, and therefore can induce sparsity in the correlation matrix through clique separation; consequently, the product graphical model empowers practitioners to encapsulate their true prior beliefs to build a model more attuned to the problem at hand. Secondly, we have shown the importance of incorporating spatial correlation into crop yield models; ignoring this information can lead to bias both in model identification and prediction. Specifically, we observed that a common least squares fit of crop yield on some explanatory variables over the entire region resulted in bi- ased residual errors, and hence violated the assumptions of the model. To avoid this 87 problem, we could fit each agricultural region with its own model. The problem, however, is that this ignores information between crop regions, and as such we ob- served reduced prediction power and model identifiability. We addressed this issue through the use of a Bayesian model which modelled all regions together, yet ac- counted for spatial correlation. This model smooths and stabilizes prediction and also allows for analytic and therefore efficient updating and prediction. In addi- tion, we created a non-stationary prior distribution to address the issue of province to province variability resulting from provincial differences in policy, management, and other factors affecting yield. Through cross-validation, we demonstrated this model to achieve improved prediction performance in modelling Canadian wheat yield over the least squares model which ignores spatial dependence, and hope that others will attempt to replicate our findings in other contexts based on the promise seen in this application. Lastly, by augmenting the dimensionality of the underlying geographic space, we have developed a highly flexible approach for handling the nonstationarity that often arises in environmental systems. While ostensibly similar to image warping, the proposed method avoids the issue of folding and allows one to model much more complex nonstationarity patterns through interdimensional expansions, al- lowing the user to perform nonparametric learning of the mapping function. In addition, through the use of a group lasso penalty, we are able to estimate the number of augmented dimensions, as well as regularize the optimization problem. Lastly, we have highlighted the practical application of the dimension expansion approach in three examples, two of which use data from observed environmental processes. 88 6.2 Future Work 6.2.1 Nonstationarity A Bayesian image warping approach which allows covariate effects to be included in the correlation structure has recently been suggested by Schmidt et al. [80]. By treating covariates as analogous to geographic coordinates, they warp the combined location-covariate space into a deformed space of the same dimension. To achieve computational efficiency, they consider a special case which restricts the form of the possible mapping function and assumes the spatial process to be a 2D manifold with covariates treated as separate values at each location. In practice, environmental data will often take the form of a number of mea- surements made over time at each location rather than true spatial replications per se. In order to try and isolate the purely spatial part of the process, the mean func- tion may incorporate a temporal component, modelling underlying temporal pat- terns and allowing the possibility of time-varying covariates, or even space-time interactions. In the absence of such covariate information, it would be possible to consider the notion of time-varying nonstationarity structure. For instance, if one wants to study the changing impact of cities and industrial areas on air pollution levels, examining changes in stationarity over time would be a valuable way to un- derstand these changes. The dimension expansion framework is also amenable to multivariate extensions. We are currently exploring a scenario whereby the dimen- sion expansion functions and locations have a hierarchical structure, allowing the dimension expansion to vary for different variables, yet be tied together through the hierarchy. We have demonstrated how dimension expansion can be used to perform pre- 89 dictions in the transformed, stationary space and mapped back to the original space. At present the choice of the mapping, learning of latent locations, and prediction are performed in isolation. As the Sampson and Guttorp (1992) approach was set within a Bayesian framework by Damian et al. [22] and Schmidt and O’Hagan [79], setting the current algorithmic approach within such an inferential frame- work would allow the inherent uncertainty to be accurately reflected in resulting inferences and this is the goal of current research. 6.2.2 Sparsity and group correlation In addition to further studies of nonstationarity, we are also developing new work for modeling sparse multivariate time-series, where dependence is present in the correlation between successive values, between neighboring variables, and in the shared sparsity patterns. We propose an original class of flexible Bayesian linear models for dynamic sparsity modelling. The proposed class of models expands upon the existing Bayesian literature on sparse regression using generalized multi- variate hyperbolic distributions. The properties of the models are explored through both analytic results and simulation studies. We demonstrate the model on a fi- nancial application where it is shown that it accurately represents the patterns seen in the analysis of stock and derivative data, and is able to detect major events by filtering an artificial portfolio of assets. Early work has been released as a technical report ([17]). 6.2.3 Structural health monitoring The use of latent structure in modeling correlation is important in a wide variety of fields, including structural health monitoring (SHM). One of the principal goals 90 of SHM is to monitor structures of various sorts for damage. By placing vibration sensors on a structure and inducing vibration, we can learn the resonant frequen- cies of the structure in its damaged and undamaged states. We have developed a set of tools based on support vector regression ([4, 11, 13]), and are currently us- ing Bayesian graphical models to address the same problem while simultaneously incorporating knowledge of the correlation structure between vibration sensors. 6.2.4 Monte Carlo The use of latent structure in modeling often results in partially or fully noniden- tifiable models, for which Monte Carlo becomes very difficult due to ridges and multiple modes in the posterior distribution of interest. In addition to developing Monte Carlo methods for prior sensitivity and cross-validation ([12]), we are also developing a suite of tools for automated Monte Carlo analysis in complex models, such as those which attempt to model latent structure. 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