- Articles & preprints
- Submission
- Policies
- Peer review
- Editorial board
- About
- EGU publications
- Manuscript tracking

Journal cover
Journal topic
**Nonlinear Processes in Geophysics**
An interactive open-access journal of the European Geosciences Union

Journal topic

- Articles & preprints
- Submission
- Policies
- Peer review
- Editorial board
- About
- EGU publications
- Manuscript tracking

- Articles & preprints
- Submission
- Policies
- Peer review
- Editorial board
- About
- EGU publications
- Manuscript tracking

**Research article**
17 Sep 2020

**Research article** | 17 Sep 2020

Applications of matrix factorization methods to climate data

- CSIRO Oceans and Atmosphere, Hobart, Australia

- CSIRO Oceans and Atmosphere, Hobart, Australia

**Correspondence**: Dylan Harries (dylan.harries@csiro.au)

**Correspondence**: Dylan Harries (dylan.harries@csiro.au)

Abstract

Back to toptop
An initial dimension reduction forms an integral part of many analyses in
climate science. Different methods yield low-dimensional representations
that are based on differing aspects of the data. Depending on the features of
the data that are relevant for a given study, certain methods may be more
suitable than others, for instance yielding bases that can be more easily
identified with physically meaningful modes.
To illustrate the distinction between particular methods and identify
circumstances in which a given method might be preferred, in this paper we
present a set of case studies comparing the results obtained using the
traditional approaches of empirical orthogonal function analysis and *k*-means
clustering with the more recently introduced methods such as archetypal analysis
and convex coding.
For data such as global sea surface temperature anomalies, in which there is
a clear, dominant mode of variability, all of the methods considered
yield rather similar bases with which to represent the data while differing in reconstruction accuracy for a given basis size. However, in the absence of such
a clear scale separation, as in the case of daily geopotential height anomalies,
the extracted bases differ much more significantly between the methods.
We highlight the importance in such cases of carefully considering the
relevant features of interest and of choosing the method that best targets precisely those features so as to obtain more easily interpretable results.

Download & links

How to cite

Back to top
top
How to cite.

Harries, D. and O'Kane, T. J.: Applications of matrix factorization methods to climate data, Nonlin. Processes Geophys., 27, 453–471, https://doi.org/10.5194/npg-27-453-2020, 2020.

1 Introduction

Back to toptop
A ubiquitous step in climate analyses is the application of an initial dimension reduction method to obtain a low-dimensional representation of the data under study. This is, in part, driven by the purely practical fact that large, high-dimensional datasets are common, and to make analysis feasible, some initial reduction in dimension is required. Often, however, we would like to associate some degree of physical significance with the elements of the reduced basis, for instance by identifying separate modes of variability. Given the wide variety of possible dimension reduction methods to choose from, it is important to understand the strengths and limitations associated with each for the purposes of a given analysis.

Perhaps the most familiar example in climate science is provided by an empirical orthogonal function (EOF; Lorenz, 1956; Hannachi et al., 2007) or principal component analysis (PCA; Jolliffe, 1986), which identifies directions of maximum variance in the data, or, more generally, the directions maximizing a chosen norm. The difficulties inherent in interpreting EOF modes physically have been thoroughly documented (Dommenget and Latif, 2002; Monahan et al., 2009) and partly motivate various modifications of the basic EOF analysis (Kaiser, 1958; Richman, 1986; Jolliffe et al., 2003; Lee et al., 2007; Mairal et al., 2009; Witten et al., 2009; Jenatton et al., 2010).

Another approach to constructing interpretable representations is based on
cluster analysis, which, in its simplest variants, identifies regions
of phase space that are repeatedly visited (MacQueen, 1967; Ruspini, 1969; Dunn, 1973; Bezdek et al., 1984). The utility of clustering-based methods is founded
on the apparent existence of recurrent flow patterns over a range of
timescales (Michelangeli et al., 1995). As estimating the multidimensional probability density function (PDF) associated with the
distribution of states is generally difficult, clustering methods attempt to
detect groupings or regions of higher point density, which may in some cases
be approximations to peaks in the underlying distribution, and otherwise
detect preferred weather patterns or types (Legras et al., 1987; Mo and Ghil, 1988).
Extensions to standard clustering algorithms may take into account the fact
that such patterns usually exhibit some degree of persistence or
quasi-stationarity (Dole and Gordon, 1983; Renwick, 2005). Hierarchical or partitioning-based clustering techniques, of which *k*-means is a popular example,
have been widely used to identify spatial patterns
associated with regimes or to classify circulation types (see, e.g.,
Mo and Ghil, 1988; Stone, 1989; Molteni et al., 1990; Cheng and Wallace, 1993; Hannachi and Legras, 1995; Kidson, 2000; Straus et al., 2007; Fereday et al., 2008; Huth et al., 2008; Pohl and Fauchereau, 2012; Neal et al., 2016). Despite their widespread
use, there can be difficulties in interpreting the resulting patterns
(Kidson, 1997), while ambiguities in selecting the number of clusters
can lead to conflicting characterizations of regimes (Christiansen, 2007).
In particular, while *k*-means is widely used due to its simplicity,
if the data do not fall into well-defined, approximately spherical clusters, the method need not provide a particularly useful classification
of each sample, in which case an alternative clustering algorithm may be
more appropriate.

The output of clustering algorithms such as *k*-means is an assignment of
each data point to a cluster and a collection of cluster centroids corresponding to the mean within each cluster. The result is a partition of
the phase space in which the elements of each partition are taken to be
well represented by the cluster mean, as in vector quantization applications
(Lloyd, 1982; Forgey, 1965). While this can yield a useful decomposition
of the data, the ideal case occurring when the data do indeed form well-defined
clusters, a representation in terms of cluster means is not always effective
or suitable for all applications. For instance, the *k*-means centroids need
not be realized as observed points within the dataset^{1}, and, by virtue of their
definition, will generally not afford a good representation of edges or
extrema of the observed data. When such features are relevant, a possible
alternative is to employ methods that construct a basis from points lying on
(or outside of) the convex hull of the data. An example of this approach is
given by archetypal analysis (AA; Cutler and Breiman, 1994; Stone and Cutler, 1996; Seth and Eugster, 2016), in
which the basis vectors are required to be convex linear combinations of the
data points that minimize a suitable measure of the reconstruction error. This
has the effect of identifying points corresponding to extrema of features
within the data. Unlike partitioning-based cluster algorithms such as *k*-means, AA does not yield a partition of the data into a set of clusters,
so that it is not so straightforward to assign observations to a set of
regimes. Instead, each observation is represented as a linear combination of
the reduced basis of archetypes with appropriate convex weights and is in this sense similar to PCA. Compared to PCA, however, the archetype basis may (in
some cases) be more easily interpreted, as the basis vectors correspond to
extreme points of the observed data, or convex combinations thereof, rather
than abstract directions in the data space.

Unlike PCA and standard clustering methods, AA has only relatively recently
found use in climate studies (Steinschneider and Lall, 2015; Hannachi and Trendafilov, 2017). As a
result, there has been little comparison of the results of using AA for
dimension reduction versus more traditional approaches. Dimension reduction
methods such as PCA, *k*-means, and AA can all be expressed generically as
approximately representing the data in terms of some set of basis vectors
(Mørup and Hansen, 2012), possibly with an additional stochastic error
term^{2},

$$\begin{array}{}\text{(1)}& \mathit{x}\left(t\right)\approx \widehat{\mathit{x}}\left(t\right)=\sum _{i=\mathrm{1}}^{k}{z}_{i}\left(t\right){\mathit{w}}_{i}+\mathit{\u03f5}\left(t\right),\end{array}$$

where ** x**(

Ultimately though, a fuller understanding might be best obtained by using
generalizations of the above methods or combinations of multiple methods. AA can be regarded as a constrained convex encoding (Lee and Seung, 1997) of the
dataset, where the basis vectors are restricted to being linear combinations of observed data points. Relaxing this constraint allows for a more flexible reduction in which the archetypes may not be representable as convex
combinations of the data (Mørup and Hansen, 2012), although the effectiveness of doing
so will depend somewhat on the underlying data-generating process. The problem of finding a convex coding of a given dataset can be unified with PCA and
*k*-means clustering by phrasing each as a generic matrix factorization
problem (Singh and Gordon, 2008); we review the basic formulation in
Sect. 2. In addition to providing a consistent
framework for defining each method, it is straightforward to incorporate
additional constraints or penalties, e.g., for the purposes of feature
selection or to induce sparsity (Jolliffe et al., 2003; Lee et al., 2007; Mairal et al., 2009; Witten et al., 2009; Jenatton et al., 2010; Gerber et al., 2020). Solving the resulting
(usually constrained) optimization problem amounts to learning a dictionary
with which to represent the data, with different methods producing different
dictionaries. By carefully defining the optimization problem, the learned
dictionary can be tuned to target particular features in the data. Below, we
demonstrate this process by utilizing a recently introduced regularized convex coding (Gerber et al., 2020), which allows for feature selection to
be performed by varying a regularization parameter. By tuning the imposed
regularization to optimize the reconstruction or prediction error, the
relative performance of selecting a basis lying on or outside the convex hull
can be compared to one that preferentially extracts cluster means.

The purpose of this paper is to explore some of the above issues in the
context of climate applications, in the hope that this may provide a useful
aid for researchers in constructing their own analyses. In particular, we aim
to illustrate some of the strengths and weaknesses of PCA and other dimension
reduction methods, using as examples *k*-means, AA, and general convex coding,
by applying each method in a set of case studies. We first apply the methods to
an analysis of global sea surface temperature (SST) anomalies, as in
Hannachi and Trendafilov (2017). Interannual SST variability is in this case dominated by
El Niño–Southern Oscillation (ENSO) activity (Wang et al., 2004), for which there is a large-scale separation between this mode and the sub-leading modes of variability, and consequently all methods are effective in detecting this
feature of the data. Differences between the methods do arise at smaller
scales, however. We then consider a similar analysis of daily 500 hPa geopotential height anomalies. Unlike SST, the time series of height anomalies generally does not exhibit noticeably large excursions corresponding to
weather extremes; in this case, extremes in the data are characterized not by
the amplitude of the anomalies, but by their temporal persistence. This demonstrates a limitation of all of the considered methods, namely, that (at
least in their basic formulation) persistence or quasi-stationarity is not
taken into account. Thus, a direct application of AA or convex codings may not
be effective in detecting extremes, while clustering methods may be more
informative in detecting recurrent weather patterns (to the extent that any
such regimes are present in the analyzed fields).

The remainder of this paper is structured as follows. In Sect. 2 we review the dimension reduction methods used. In Sect. 3, we compare the results obtained using each method in a set of case studies to illustrate the distinctions between the methods. Finally, in Sect. 4 we summarize our observations and discuss possible future extensions.

2 Matrix factorizations

Back to toptop
In this section, we first describe the dimension reduction methods that we
use in our case studies. As noted above, PCA, *k*-means,
and convex coding applied to multidimensional data can all be phrased as
matrix factorization problems. Given a collection^{3} of *d*-dimensional data points ** x**(

$$\begin{array}{}\text{(2)}& \mathbf{X}\approx \widehat{\mathbf{X}}={\mathbf{ZW}}^{T},\end{array}$$

where $\mathbf{Z}\in {\mathbb{R}}^{T\times k}$ is a *T*×*k*-dimensional matrix with rows giving the weights *z*_{i}(*t*), and $\mathbf{W}\in {\mathbb{R}}^{d\times k}$
contains the basis vectors *w*_{i} as columns. Note that here and
below we assume that the columns of **X** have zero mean; if not, this can always
be arranged by first centering the data,

$$\mathbf{X}=\left({\mathbf{I}}_{T\times T}-{\displaystyle \frac{\mathrm{1}}{T}}{\mathbf{1}}_{T}{\mathbf{1}}_{T}^{T}\right)\stackrel{\mathrm{\u0303}}{\mathbf{X}},$$

where $\stackrel{\mathrm{\u0303}}{\mathbf{X}}$ denotes the original data, **I**_{T×T} the *T*×*T*
identity matrix, and **1**_{T} the *T*-dimensional vector of ones.

The factors **W** and **Z** are calculated as the minimizers of a suitably
chosen cost function $F(\mathbf{W},\mathbf{Z};\mathbf{X})$, measuring (in some application-dependent sense) the quality of the reconstruction $\widehat{\mathbf{X}}$, subject to the constraint that
they are within certain feasible regions Ω_{Z} and Ω_{W},

$$\begin{array}{}\text{(3)}& (\mathbf{W},\mathbf{Z})\equiv \underset{\mathbf{Z}\in {\mathrm{\Omega}}_{Z},\mathbf{W}\in {\mathrm{\Omega}}_{W}}{\mathrm{arg}\phantom{\rule{0.125em}{0ex}}min}F(\mathbf{W},\mathbf{Z};\mathbf{X}).\end{array}$$

A typical cost function takes the form of a decomposable loss function, measuring the reconstruction error, together with a set of penalty terms imposing any desired regularization, i.e.,

$$\begin{array}{}\text{(4)}& \begin{array}{rl}F(\mathbf{W},\mathbf{Z};\mathbf{X})& ={\displaystyle \frac{\mathrm{1}}{T}}\sum _{t=\mathrm{1}}^{T}\mathrm{\ell}(\mathbf{W},\mathit{z}(t);\mathit{x}(t\left)\right)+{\mathit{\lambda}}_{W}{\mathrm{\Phi}}_{W}\left(\mathbf{W}\right)\\ & +{\mathit{\lambda}}_{Z}{\mathrm{\Phi}}_{Z}\left(\mathbf{Z}\right).\end{array}\end{array}$$

In Eq. (4) we have, for simplicity, supposed that
**W** and **Z** are independently regularized, with the tunable parameters
*λ*_{W} and *λ*_{Z} governing the amount of regularization. Common
choices for the loss function include the ℓ_{2}-norm^{4},

$$\begin{array}{}\text{(5)}& \mathrm{\ell}(\mathbf{W},\mathit{z}(t);\mathit{x}(t\left)\right)\equiv {\displaystyle \frac{\mathrm{1}}{\mathrm{2}}}{\u2225\mathit{x}\left(t\right)-\mathbf{W}\mathit{z}\left(t\right)\u2225}_{\mathrm{2}}^{\mathrm{2}},\end{array}$$

in which case the unregularized cost is proportional to the residual sum of squared errors (RSS), the Kullback–Leibler divergence for non-negative data,

$$\begin{array}{}\text{(6)}& \begin{array}{rl}{\mathrm{\ell}}_{\mathrm{KL}}(\mathbf{W},\mathit{z}(t);\mathit{x}(t\left)\right)& =\sum _{i=\mathrm{1}}^{d}\left({x}_{i}\right(t)\mathrm{ln}{\displaystyle \frac{{x}_{i}\left(t\right)}{\left(\mathbf{W}\mathit{z}\right(t){)}_{i}}}-{x}_{i}(t)\\ & +\left(\mathbf{W}\mathit{z}\right(t\left){)}_{i}\right),\end{array}\end{array}$$

or the wider class of Bregman divergences (Bregman, 1967; Banerjee et al., 2005; Singh and Gordon, 2008).
In addition to the soft constraints imposed by the penalty terms Φ_{W}(**W**)
and Φ_{Z}(**Z**), the feasible regions ${\mathrm{\Omega}}_{Z}\subset {\mathbb{R}}^{T\times k}$ and ${\mathrm{\Omega}}_{W}\subset {\mathbb{R}}^{d\times k}$
define a set of hard constraints that must be obeyed by the optimal solutions.
The definition of a given method thus comes down to a small number of
modeling choices regarding the cost function and feasible regions.

PCA, in its synthesis formulation, is equivalent to minimizing an ℓ_{2}-loss
function with ${\mathit{\lambda}}_{W}={\mathit{\lambda}}_{Z}=\mathrm{0}$, i.e.,

$$\begin{array}{}\text{(7)}& {F}_{\mathrm{PCA}}(\mathbf{W},\mathbf{Z};\mathbf{X})={\displaystyle \frac{\mathrm{1}}{\mathrm{2}T}}{\u2225\mathbf{X}-{\mathbf{ZW}}^{T}\u2225}_{F}^{\mathrm{2}},\end{array}$$

where ∥A∥_{F} denotes the Frobenius norm of a matrix,
subject to the constraint that the reconstruction $\widehat{\mathbf{X}}$ has rank at most
*k*. The problem in this case has a global minimum (Eckart and Young, 1936) given by
the singular value decomposition (SVD), and the retained basis vectors
*w*_{i} correspond to the directions of maximum variance. In our
numerical case studies, we adopt the convention $\mathbf{Z}=\mathbf{U}\mathbf{\Sigma}/\sqrt{T-\mathrm{1}}$,
**W**=**V** for the PCA factors **W** and **Z**, with the SVD of **X**=**U****Σ****V**^{T} for real **X**. While the existence of a direct numerical solution for the optimal
factorization makes PCA very flexible and easy to apply, the basis vectors may
be difficult to interpret, for instance if they have many non-zero components.
Sparse variants of PCA that attempt to improve on this may be arrived at by
introducing a sparsity-inducing regularization Φ_{W}(**W**), of which a common choice is the ℓ_{1}-penalty

$$\begin{array}{}\text{(8)}& {\mathrm{\Phi}}_{W}\left(\mathbf{W}\right)=\sum _{i=\mathrm{1}}^{k}{\u2225{\mathit{w}}_{i}\u2225}_{\mathrm{1}}.\end{array}$$

The partitioning that results from *k*-means clustering may also be written
in terms of a factorization $\widehat{\mathbf{X}}={\mathbf{ZW}}^{T}$. Whereas PCA by construction
yields an orthonormal basis with which to represent the data, the
*k*-means decomposition is in terms of the cluster centroids and
iteratively attempts to minimize the within-cluster variance (Hastie et al., 2005). This is equivalent to minimizing an ℓ_{2}-cost function,
as in Eq. (7), subject to the constraint that the weights
matrix **Z** has binary elements, ${Z}_{ti}\in \mathit{\{}\mathrm{0},\mathrm{1}\mathit{\}}$, and rows with unit
ℓ_{1}-norm, ∥** z**(

$$\begin{array}{}\text{(9)}& {\mathrm{\Omega}}_{Z}=\mathit{\{}\mathbf{Z}\in {\mathbb{R}}^{T\times k}|{Z}_{ti}\in \mathit{\{}\mathrm{0},\mathrm{1}\mathit{\}},\phantom{\rule{1em}{0ex}}\mathbf{Z}{\mathbf{1}}_{k}={\mathbf{1}}_{T}\mathit{\}}.\end{array}$$

The corresponding basis matrix **W** then contains the cluster centroids
*w*_{i} as columns. Although both PCA and *k*-means can be seen to
minimize the same objective function^{5}, the additional
constraints in *k*-means clustering mean that finding the exact clustering
is NP-hard (Aloise et al., 2009; Mahajan et al., 2012), and heuristic methods must be used
instead (e.g., Lloyd, 1982; Hartigan and Wong, 1979). These iterative methods are not
guaranteed to find a globally optimal clustering and must be either run multiple times with different initial guesses or combined with more
sophisticated global optimization strategies to reduce the chance of finding
only a local minimum.

The convex codings that we employ below, like PCA and *k*-means, are based
on minimizing the least squares loss Eq. (7). The least
restricted version (Lee and Seung, 1997; Gerber et al., 2020) requires only that the
reconstruction lies in the convex hull of the basis, or, in other words,
that the weights **Z** satisfy the constraints

$$\begin{array}{}\text{(10)}& {\mathrm{\Omega}}_{Z}=\mathit{\{}\mathbf{Z}\in {\mathbb{R}}^{T\times k}|\mathbf{Z}\u2ab0\mathrm{0},\phantom{\rule{1em}{0ex}}\mathbf{Z}{\mathbf{1}}_{k}={\mathbf{1}}_{T}\mathit{\}},\end{array}$$

where the condition **A**⪰0 indicates an element-wise inequality. In the absence of any hard constraints on the basis **W**, the optimization
problem to be solved reads

$$\begin{array}{}\text{(11)}& (\mathbf{W},\mathbf{Z})=\underset{\mathbf{Z}\in {\mathrm{\Omega}}_{Z}}{\mathrm{arg}\phantom{\rule{0.125em}{0ex}}min}\left[{\displaystyle \frac{\mathrm{1}}{\mathrm{2}T}}{\u2225\mathbf{X}-{\mathbf{ZW}}^{T}\u2225}_{F}^{\mathrm{2}}+{\mathit{\lambda}}_{W}{\mathrm{\Phi}}_{W}\left(\mathbf{W}\right)\right],\end{array}$$

with Ω_{Z} as in Eq. (10); note
that, if these constraints are further tightened to require that
the columns of **Z** be non-negative and orthonormal, ${\mathbf{Z}}^{T}\mathbf{Z}={\mathbf{I}}_{k\times k}$,
solving this optimization problem (in general,
for *λ*_{W}=0) would yield the hard *k*-means clustering of the data
(Li and Ding, 2006). In the absence of any regularization (*λ*_{W}=0), it
follows from the Eckart–Young theorem that for a given basis size *k* the reconstruction error achieved by PCA is always no larger than that achieved by
this convex coding, which in turn achieves a residual error that is smaller than
that for *k*-means. Thus, were the goal to simply achieve the optimal
least squares reconstruction error for a given basis size, PCA remains the
preferred choice. The more constrained decomposition implied by performing a
convex coding may yield advantages in terms of interpretability or feature
selection. The choice of regularization Φ_{W} provides further flexibility
in this respect; for instance, an ℓ_{1}-penalty as in Eq. (8) can be used to induce sparsity, while
Gerber et al. (2020) suggest a penalty term of the form

$$\begin{array}{}\text{(12)}& {\mathrm{\Phi}}_{W}\left(\mathbf{W}\right)={\displaystyle \frac{\mathrm{1}}{dk(k-\mathrm{1})}}\sum _{i,j=\mathrm{1}}^{k}{\u2225{\mathit{w}}_{i}-{\mathit{w}}_{j}\u2225}_{\mathrm{2}}^{\mathrm{2}},\end{array}$$

observing that this places an upper bound on the sensitivity of the
reconstruction to changes in the data. It is worth noting that, when combined
with an ℓ_{2}-loss function, as the regularization *λ*_{W}→∞
all of the basis vectors *w*_{i} reduce to the global mean of the
dataset.

For *λ*_{W}=0, the optimal basis vectors *w*_{i} produced by
solving Eq. (11) will tend to lie
on or outside of the convex hull of the data. Standard AA adds a further
more conservative constraint to force the *w*_{i} to only lie on the
convex hull, thus requiring that the archetypes are realizable in terms of convex
combinations of the observed data. Under the more relaxed constraints, while
the overall residual error might be reduced, one may extract basis vectors
that, in reality, never occur and so are difficult to make sense of. In AA,
the additional constraint amounts to requiring that **W**=**X**^{T}**C**^{T} for some
non-negative $\mathbf{C}\in {\mathbb{R}}^{k\times T}$ with unit ℓ_{1}-norm rows,
i.e., **C**∈Ω_{C} with

$$\begin{array}{}\text{(13)}& {\mathrm{\Omega}}_{C}=\mathit{\{}\mathbf{C}\in {\mathbb{R}}^{k\times T}|\mathbf{C}\u2ab0\mathrm{0},\mathbf{C}{\mathbf{1}}_{T}={\mathbf{1}}_{k}\mathit{\}}.\end{array}$$

The corresponding optimization problem reads

$$\begin{array}{}\text{(14)}& (\mathbf{C},\mathbf{Z})=\underset{\mathbf{C}\in {\mathrm{\Omega}}_{C},\mathbf{Z}\in {\mathrm{\Omega}}_{Z}}{\mathrm{arg}\phantom{\rule{0.125em}{0ex}}min}{\displaystyle \frac{\mathrm{1}}{\mathrm{2}T}}{\u2225\mathbf{X}-\mathbf{ZCX}\u2225}_{F}^{\mathrm{2}},\end{array}$$

with Ω_{Z} as in Eq. (10). As for *k*-means,
the optimization problem to be solved in performing either a convex coding or
archetypal analysis cannot be solved exactly, and numerical methods must
be employed to find the corresponding factorization. We describe one such algorithm
for doing so in Appendix A.

The various choices of cost function and constraints defining the above
methods are summarized in Table 1.
While all of the methods fit within the broader class of matrix factorizations,
the different choices of cost functions and constraints lead to important
differences in the low-dimensional representation of the data produced by each
method. For instance, in contrast to PCA, the basis vectors produced by *k*-means,
convex coding, or AA are in general not orthogonal. In some circumstances, this
non-orthogonality may be advantageous
when the structure necessary to ensure orthogonal basis vectors (e.g., via
appropriate cancellations) obscures important features or makes interpretation of
the full PCA basis vectors difficult. A *k*-means clustering may, for example,
provide a much more natural reduction of the data when multiple distinct,
well-defined clusters are present. The cost function and choice of constraints
that define convex coding and archetypal analysis imply that the optimal basis
vectors produced by these methods are such that their convex hull (i.e., the
set of all linear combinations of the basis vectors with weights summing to one)
best fits the data. Consequently, both are well suited for describing data where points can be usefully characterized in terms of their relationship
with a set of extreme values, be they spatial patterns of large positive or negative anomalies in a geophysical field or particular combinations of spectral
components in a frequency domain representation of a signal.
PCA and *k*-means may be less useful in such cases, as neither yields a
decomposition of the data in terms of points at or outside of the boundaries of
the observations. AA differs from more general convex encodings in imposing
the stricter requirement that the dictionary elements, i.e., the archetypes,
lie on the boundary of the data. It is, in this sense, conservative, in that
the features extracted by AA lie on the convex hull of the data and so correspond
to a set of extremes that are nevertheless consistent with the observed data.
In the absence of any regularization (*λ*_{W}→0), the general convex codings
that we consider admit basis vectors that lie well outside of the observed data.
By doing so, the method finds a set of basis vectors whose convex hull better
reconstructs the data than in AA, at the cost of representing it in terms of points
that may not be physically realistic. This behavior and the impact of the different choices of cost function and constraints are sketched in
Fig. 1.

3 Case studies

Back to toptop
We now turn to a set of case studies that demonstrate some of the implications of the various differences noted above in realistic applications.
We consider two particular examples that highlight the importance of considering
the particular physical features of interest when choosing among possible
dimension reduction methods. The first example that we consider, an analysis of SST
anomalies, is characterized by a large separation of scales between modes of
variability together with key physical modes, particularly ENSO, that can be
directly related to extreme values of SST anomalies. This means that the basis
vectors or spatial patterns extracted by PCA, *k*-means, and convex coding are
for the most part rather similar in structure, and so the choice of method may
be guided by other considerations, such as the level of reconstruction error.
We then contrast this scenario with the example of an analysis of mid-latitude
geopotential height anomalies, where there is neither scale separation nor coincidence of the physical modes solely with extreme anomaly values.
As a result, methods that are based on constructing a convex coding of the data,
without targeting features that capture the dynamical characteristics of the relevant
modes, produce representations that are, arguably, more difficult to interpret and hence may be less suitable than clustering-based methods.

Following Hannachi and Trendafilov (2017), we first apply the methods described in
Sect. 2 to monthly SST data. The source of the data
is the Hadley Centre Sea Surface Temperature dataset (HadISST), version 1.1
(Rayner et al., 2003), consisting of monthly SST values on a $\mathrm{1}{}^{\circ}\times \mathrm{1}{}^{\circ}$ global grid spanning the time period from January 1870
to December 2018. Monthly anomalies are calculated by removing from
the full time series a linear warming trend and additive seasonal component, where the annual cycle is estimated based on the 1981 to 2010
base period. The analysis region is restricted to the region between
45.5^{∘} N and 45.5^{∘} S.

As the standard and most familiar method, we first perform PCA on
the SST anomalies over the time period January 1870 to December 2018
to establish a baseline set of modes. The anomalies at each grid cell are
area weighted by the square root of the cosine of the point's latitude.
To provide a rough measure of the out-of-sample reconstruction
error^{6}, the EOFs and PCs are evaluated on the first 90 % of the anomaly
time series (i.e., January 1870 to February 2004), and the root mean-square error (RMSE) defined by

$$\begin{array}{}\text{(15)}& {\text{RMSE}}_{\text{train}/\text{test}}\left(k\right)=\sqrt{{\displaystyle \frac{\mathrm{1}}{dT}}\sum _{i=\mathrm{1}}^{d}\sum _{t=\mathrm{1}}^{T}{\left[{x}_{i}\left(t\right)-{\widehat{x}}_{i}\left(t\right)\right]}^{\mathrm{2}}}\end{array}$$

is computed for the training set and for the test set consisting of the
remaining 10 % of the data. In Eq. (15),
${\widehat{x}}_{i}\left(t\right)$ is the *i*th dimension of the reconstruction from
*k* EOFs, and *T* refers to the size of the training or test set, as
appropriate. We consider the results of retaining the first *k* modes for
$k=\mathrm{1},\mathrm{\dots},\mathrm{40}$; for reference, the first 40 modes account for
approximately 85 % of the total variance. The fraction of the total variance
in the training set associated with the leading 10 modes is shown in
Fig. 2. The most obvious feature of this
variance spectrum is the well-known separation between the first and subsequent
modes, with the first mode associated with ENSO variability on interannual
timescales and large spatial scales. This is evident in the spatial patterns of the EOFs shown in Fig. 3, in which the first mode
shows the canonical ENSO pattern of SST anomalies, while the higher-order modes correspond to spatially smaller-scale variability.

With the above EOF patterns as a point of reference, we now turn to comparing
the representation of the dataset produced by each of *k*-means, archetypal
analysis, and convex coding. In each case, the same dataset (i.e., latitude
weighted, detrended anomalies) is used as input to the dimension reduction
method. The individual data points consist of anomaly maps at each time step; in other words, the dimension reduction is performed in the state space rather
than the sample space (Efimov et al., 1995), such that each dictionary vector
corresponds to a particular spatial pattern of SST anomalies.
We fit each method using dictionaries of size $k=\mathrm{1},\mathrm{\dots},\mathrm{20}$ on
the first 90 % of the dataset, using the last 10 % of samples to provide
a simple picture of out-of-sample performance.

We consider the results of a *k*-means clustering analysis of the data first.
For each value of *k*, the algorithm is restarted 100 times with different
initial conditions, and the partitioning found to have the lowest reconstruction error is chosen. As is typical of clustering procedures where the number of
clusters must be specified a priori, determining the most appropriate choice of
*k* is non-trivial. One commonly used heuristic is to inspect a “scree”
plot of the within-cluster sum of squares (Tibshirani et al., 2001),

$$\begin{array}{}\text{(16)}& {W}_{k}=\sum _{i=\mathrm{1}}^{k}{\displaystyle \frac{\mathrm{1}}{\left|{C}_{i}\right|}}\sum _{{t}_{\mathrm{1}},{t}_{\mathrm{2}}\in {C}_{i}}{\u2225\mathit{x}\left({t}_{\mathrm{1}}\right)-\mathit{x}\left({t}_{\mathrm{2}}\right)\u2225}_{\mathrm{2}}^{\mathrm{2}},\end{array}$$

where $\left|{C}_{i}\right|$ is the size of cluster *C*_{i} as a function of the number of clusters. The preferred number of clusters *k*^{∗} is identified as the location
of an elbow or kink in this curve if any such feature is present. Alternatively, various indices (see, e.g., Arbelaitz et al., 2013, for a review) or Monte Carlo procedures, such as the gap statistic (Tibshirani et al., 2001), have been
proposed for assessing whether a given *k* is suitable. For the present
application, plots of the normalized within-cluster sum of squares and the
gap statistic computed using 100 Monte Carlo experiments for each *k* with
a null model generated by PCA are shown in Fig. 4.
The plot of *W*_{k} as a function of *k* does not show an obvious elbow at any
particular *k*≤20; similarly, using the gap statistic curve, one would conclude that the *k*=1 cluster is preferred^{7}. This simply reflects
the fact that the anomaly data do not form well-separated clusters (with respect to the assumed Euclidean distance) in the full state space,
making interpreting the different clusters as dynamically distinct
regimes difficult, although this does not preclude using the clustering model
as a discrete representation of the full dynamics (e.g., Kaiser et al., 2014).

The partitioning provided by a simple clustering method such as *k*-means may
also still be useful in classifying samples so long as
proximity in state space, or similarity more generally, carries meaningful
information for an application. In the case of detrended SST anomalies, the
magnitudes and spatial distribution of the anomalies are of themselves
informative, and key drivers such as ENSO manifest as relatively large
excursions from anomalies associated with higher-frequency variability.
Consequently, a *k*-means clustering of the anomalies does result in a
partition in which the cluster centroids, or a subset thereof, can be
identified with physical modes, as the algorithm finds several
clusters consisting of these extremes of the point cloud, while the remainder partition the bulk of the data. The simplest, if somewhat trivial, case for
which this can be seen is for *k*=3 clusters,
shown in Fig. 5.
Two of the plotted centroids are recognizable as canonical El Niño and
La Niña patterns, while the last corresponds to a near-climatological state.
The former two clusters are dominated by months at the ends of the
first principal axis (i.e., along the leading EOF).
The fact that two clusters are required to
represent both phases of the leading EOF is due to the fact that the overall
sign of the cluster centroids, which correspond to points in the data space,
is meaningful, unlike in PCA; note that the same is true for the convex
coding methods to be considered below. To obtain some sense of
the relative distributions of the points within each of the clusters, in
Fig. 6 we show the projection of the data into
two dimensions generated by metric multidimensional scaling (MDS). The points assigned to the first and second clusters are those
furthest from the mean state, while the remaining cluster accounts for the
bulk region. The difference between the first and second centroids is closely
aligned with the leading EOF. Similar behavior results when a larger number of
clusters is specified.

The fact that, due to the dominating role of ENSO variability, a *k*-means
clustering results in several clusters corresponding to large-magnitude anomalies in turn implies that the corresponding centroids will be
relatively close to the convex hull of the anomaly point cloud. Consequently,
these centroids will also closely resemble a subset of the archetypes derived
from an archetypal analysis of the same data, at least qualitatively. Relaxing
the requirement that the dictionary vectors lie on the convex hull, one may
still expect that a convex coding applied to the data will identify features
along the same directions, albeit with inflated magnitude so as to reduce the
resulting reconstruction error. To verify this, we perform a standard
archetypal analysis and a regularized convex coding of the detrended anomalies,
in each case restarting the optimization algorithm 100 times and choosing the
best encoding from the multiple starts. The regularization in
Eq. (12) is used; to
illustrate the effect of this regularization, we show results for *λ*_{W}=0
(i.e., no regularization) and *λ*_{W}=10^{3}. The fitted archetypes for
*k*=3 are shown in Fig. 7, and the corresponding
dictionary vectors for the regularized convex coding are shown in
Fig. 8.
The patterns obtained using the two methods in Figs. 7
and 8 are very similar; in both cases,
El Niño and La Niña patterns are found together with a third
configuration containing a cold tongue in the equatorial Pacific with positive
anomalies to the north and south, as noted by Hannachi and Trendafilov (2017). However,
the magnitude of the anomalies is substantially larger when using an
unregularized convex coding as the fitted dictionary vectors are chosen to sit
well outside the observed point cloud. While the states correspond to very
similar relative signs of anomalies, it is arguably more difficult to interpret
the individual dictionary vectors found by the unregularized convex coding
physically, as the large anomalies represent far more extreme states than are
observed in the data. On the other hand, this also permits a much smaller
reconstruction error for a given basis size and so may be preferable when a higher-fidelity reconstruction is required^{8}. The effect of
the regularization Eq. (12) is to penalize over-dispersion
of the dictionary vectors and so select features that are insensitive to small variations in the input data. For sufficiently large *λ*_{W},
the resulting states lie on or within the convex hull of the data and as a result are more comparable to the states found by AA or *k*-means. This is
illustrated in Fig. 9, from which it can be seen that the
basis vectors produced by the regularized convex coding are closer (in
Euclidean distance) to those produced by AA than the unregularized basis
vectors.

This trade-off between reproducing the data with small errors and
constraining the basis vectors to be close to the observed data is also
evident in Fig. 10, where the reconstruction RMSEs within the training and held-out test datasets for each of the methods are plotted as
a function of the dimension of the reduced-order representation. In the absence of any regularization, the RMSE produced by a convex coding
of the data is close to the globally optimal result obtained from PCA.
For *λ*_{W}=10^{3}, the obtained RMSE is larger and similar to that for AA,
while *k*-means leads to the largest errors due to the very coarse
representation of each data point by the centroid of its assigned cluster. It follows that, for a given number of basis vectors, an unregularized convex
coding yields performance approaching that of PCA in terms of reconstruction
errors, while AA and *k*-means in turn are more constrained and hence reproduce
the data with somewhat larger errors. The ordering of the methods in terms
of reconstruction error observed in Fig. 10 is expected to
be the case more generally. As noted in
Sect. 2, PCA provides the globally optimal
reconstruction of the data matrix with a given rank, in the absence of any
constraints, and so amounts to a lower bound on the achievable reconstruction
error. Of the remaining methods, the additional freedom to locate the basis
elements outside of the convex hull of the data when performing an unregularized
convex coding allows for a lower reconstruction error than is achievable using
archetypes. For a given basis size the hard clustering resulting from
*k*-means generally results in the largest RMSE. For larger values of the
regularization *λ*_{W}, the optimal basis elements sit within the convex hull
of the data and provide a fuzzy representation of the data with a progressively
increasing RMSE. In this particular analysis of SST data,
the performance of the different methods with respect to reconstruction error
is one distinguishing factor that may guide the choice of method; while all four
produce similar large-scale spatial patterns, for a given basis size PCA provides
the lowest reconstruction error and might be preferred if information loss
is a significant concern.

SST anomalies are an example of a dataset in which a dominant mode is well separated in scale from subleading modes of variability. As a result, all of the dimension reduction methods that we consider extract similar bases (patterns) with which to represent the data, and these can be identified with well-known physical modes. Moreover, physically interesting events such as extremes correspond to large-magnitude anomalies relative to the mean state, i.e., at the boundary of the point cloud, and so can be directly extracted by those methods that look for dictionary vectors in the convex hull of the data. This is not true for many variables of interest, however, and so we now compare the behavior of the methods when applied to data that do not exhibit these features.

We consider Northern Hemisphere (NH) daily mean anomalies of 500 hPa
geopotential height, ${Z}_{\mathrm{g}\mathrm{500}\phantom{\rule{0.125em}{0ex}}\mathrm{hPa}}^{\prime}$, between 1 January 1958 and
31 December 2018. The geopotential height data used are obtained from the
Japanese 55-year Reanalysis (JRA-55, Kobayashi et al., 2015; Harada et al., 2016).
Anomalies are formed by subtracting the climatological daily mean based on the
1 January 1981 to 31 December 2010 reference period. Unlike monthly SST, there
is no clear scale separation in the resulting time series; in
Fig. 11 we show the leading EOFs and the percentage of the
total variance explained by each.
Additionally, the characterization of regimes is more complicated. In
particular, physically significant features in the height field are expected
to be quasi-stationary or persistent (Michelangeli et al., 1995; Dole and Gordon, 1983; Mo and Ghil, 1988) and are not solely distinguished by their location in the state space. Such
features may therefore be better identified as modes of the state space PDF,
arising either due to longer residency times or frequent recurrence of
particular states. Where this gives rise to regions of higher density in state
space, clustering algorithms such as *k*-means might be expected to reasonably
well represent the corresponding regimes. Convex reduction methods, on the
other hand, default to being less sensitive to recurrence; as in the preceding
SST example, the fitted dictionary vectors will correspond to points near the
boundary of observed data. In doing so, AA and methods based on convex coding
preferentially represent the data in terms of deep, but potentially
infrequent or highly transient, lows or highs. While this
remains an adequate representation simply for reconstructing the
full observations, the resulting basis may be more difficult to relate to
traditionally identified metastable atmospheric regimes, for instance.

In the case of NH geopotential height anomalies, the cluster centroids,
archetypes, and dictionary basis vectors that result from applying *k*-means,
AA, and convex coding to the leading 167 PCs^{9} of ${Z}_{\mathrm{g}\mathrm{500}\phantom{\rule{0.125em}{0ex}}\mathrm{hPa}}^{\prime}$ are shown for *k*=4 clusters in
Fig. 12. Note that inspection of the scree plots
(not shown) does not indicate a strong preference for a given number of
clusters or states for *k*≤20, and we choose *k*=4 as a simple example.
The relative dispositions of each of the states with respect to each other,
as measured by Euclidean distance, are visualized in Fig. 13
on the basis of a two-dimensional metric MDS. In some respects, the performance
of the different methods is similar to that seen in the SST example; for example,
as expected on general grounds, the ordering of the methods with respect to achieved RMSE (not shown) is the same as in Fig. 10.
Similarly, AA and the unregularized convex coding select, by design, basis vectors
that lie on or outside of the convex hull of the data, with the latter having
the freedom to choose a basis corresponding to much larger departures from the
mean so as to reduce the reconstruction error; note that the precise
degree to which the basis vectors lie outside the convex hull will depend on
the particular data at hand, but the behavior is otherwise generic.
However, unlike the SST case, the
${Z}_{\mathrm{g}\mathrm{500}\phantom{\rule{0.125em}{0ex}}\mathrm{hPa}}^{\prime}$ data do not exhibit a dominant axis of variability; i.e., there is no clear scale separation between modes. In
Fig. 13 this manifests as the absence of a clearly preferred
axis along which the basis vectors are distributed^{10}; cf. Fig. 9. As the variance in the
data is not dominated by a single principal axis, the basis vectors extracted
by each of the methods are more evenly distributed around the point cloud,
in turn leading to greater differences between the fitted bases.

While all the methods identify a feature that is strongly reminiscent of the North Atlantic Oscillation, there is somewhat more variation in the remaining representative states, in contrast to the case of SST anomalies. In particular, the centroids
identified by *k*-means are no longer in close correspondence to the representative states constructed by either AA or an unregularized convex coding.
The *k*-means centroids are characterized by relatively small magnitude
anomalies with positive amplitude at longitudes associated with blocking
(Pelly and Hoskins, 2003); in particular, the location of the anticyclonic anomaly
in cluster 3 in Fig. 12 closely coincides with the
center of action of the EU1 (Barnston and Livezey, 1987) or SCA (Bueh and Nakamura, 2007)
pattern associated with Scandinavian blocking. The archetypes, in comparison,
appear to exhibit a more pronounced wave-train structure, in addition to
corresponding to generally larger magnitude anomalies. As expected, the basis
obtained via an unregularized convex coding represents far larger anomalies again,
defining representative states that are much more extreme than the bulk of the
observed daily anomaly fields. The role of the regularization in feature
selection is clearly illustrated by comparing the two right-most columns of
Fig. 12. By increasing the regularization parameter,
the method can be tuned to place less emphasis on capturing large-amplitude, noisy variations and target less sensitive features in the data. In this case,
for *λ*_{W}≈1 the convex coding basis vectors are in close
agreement with the archetypes, while for *λ*_{W}≈10 they essentially
coincide with the *k*-means centroids. In this sense, by appropriate choice of
regularization it is possible to interpolate between a representation of the
data in terms of points on the convex hull and mean features, depending on how
much weight is placed on minimizing the reconstruction error.
While here we consider only a few levels of regularization for illustrative
purposes, in general the degree to which this is done, i.e., the choice of
*λ*_{W}, can be guided by standard model selection methods,
such as cross-validation.

In the absence of any regularization, the patterns obtained by a convex
coding and by AA correspond to extreme departures from the mean state but cannot necessarily be directly interpreted as individually
representing particular physical extremes. As noted above, this arises due to
the fact that such extremes are not necessarily associated with boundaries in
state space but may instead be due to extended residence or persistence of a
given (non-extreme) state. This difficulty in directly relating atmospheric
extremes to a basis produced by AA or similar methods can be clearly demonstrated
by considering the representation of a given event in terms of this basis.
A dramatic example is provided by the 2010 summer heatwave in western Russia
that saw an extended period of well-above-average daily temperatures and poor air quality and was associated with substantial excess mortality and economic
losses (Barriopedro et al., 2011; Shaposhnikov et al., 2014). The upper-level circulation
during July 2010 was characterized by persistent blocking over eastern
Europe (Dole et al., 2011; Matsueda, 2011). The associated monthly mean pattern of
height anomalies for that month (see, e.g., Fig. 2 of Dole et al., 2011) most
closely resembles the pattern for cluster 3 obtained with *k*-means in
Fig. 12. Consistent with this is the fact that most
days during July 2010 are assigned to this cluster, as shown in
Fig. 14. Consequently, cluster 3 might be reasonably
well interpreted as representing (a class of) European blocking events.

In contrast, despite the severity of this event,
individual daily height anomalies during July 2010 are not unambiguously
identified as extremes by AA or the less constrained convex coding.
In Fig. 14 we show the time series of weights associated with the basis vectors found by AA and by convex coding either without
regularization, *λ*_{W}=0, or with a regularization parameter, *λ*_{W}=1, which for these data yields dictionary vectors very similar to those found by
AA. For all three cases, roughly equal weights are assigned to each basis
vector during July 2010. The complete anomaly is therefore represented as a
mixture of all of the dictionary vectors rather than being assigned to a single characteristic type of extreme. Evidently, any single basis pattern
extracted by these methods does not directly correspond to a particular
indicator of extreme weather conditions; extreme events of practical concern
may be spread across multiple basis vectors, making identifying such events in
this representation more difficult. The hard clustering obtained using *k*-means
is perhaps more easily interpreted in this case, as the resulting cluster
affiliations show frequent occurrences of the European blocking cluster
(i.e., cluster 3 in Figs. 12 and
14) during the peak heatwave period, with a smaller number of days assigned to clusters 2 and 4, suggesting a clearer picture of
residence in a single, persistent blocking state. On the other hand,
the coarse representation of the actual anomalies by the single cluster 3
centroid is poor compared to the reconstruction provided by AA and convex
coding but could be improved by, for example, making use of a soft clustering algorithm instead. To summarize, in the geopotential height case where extreme
events are defined not just by large anomalies but also by persistent structures, methods such as *k*-means or the regularized convex coding applied here provide a more
direct starting point for analyses of such events. These methods better identify
the relevant structures and so provide bases that are more amenable to
interpretation in terms of physical extremes. Unregularized convex coding and
archetypal analysis, on the other hand, are less appropriate in this respect,
as they do not yield a direct assignment of extreme events to individual
states.

Finally, it is worth noting that, as in the SST case study, the ambiguous classification of events provided by the convex-hull-based methods is less of a problem when state space location alone (e.g., temperature anomalies) is in itself a relevant feature. When this is not the case, as here, methods that take advantage of state space density, either due to recurrence or persistence, may be more easily interpreted, or alternatively hybrid approaches could be used in order to partition the state space.

4 Conclusions

Back to toptop
Representing a high-dimensional dataset in terms of a highly reduced basis
or dictionary is an essential step in many climate analyses. Beyond the
practical necessity of doing so, it is usually also desirable for the
individual elements of the representation to be identifiable with physically
relevant features for the sake of interpretation. A wide range of popular
dimension reduction methods, including PCA, *k*-means clustering, AA, and
convex coding, can be written down as particular forms of a basic matrix
factorization problem. These methods differ in the details of the measure of
cost that is optimized and the feasible solution regions, with the result that
different methods yield representations suitable for targeting different
features of the data. Of the methods considered here, PCA extracts directions in
state space corresponding to maximal variability, *k*-means locates central
points, and AA and similar convex codings identify points on or outside
the convex hull of the observed data and so find a representation in terms of
extreme points in state space. As different features may be relevant for
different applications, it is important to consider these distinctions between
these factorization methods and carefully choose a method that is effective
for extracting the features of interest.

In some cases, the representations obtained using different dimension reduction
methods are very similar, and one can identify more or less easily interpretable
features using any given method. This is exemplified by our first case study
of SST anomalies, in which the presence of the dominant ENSO mode ensures that
PCA, *k*-means clustering, AA, and convex coding all identify similar bases
corresponding to well-known physical modes. In this case, the main distinction
between the methods arises in the nature of the classification of the data
(e.g., hard versus soft clustering) and the accuracy with which the original
data can be reconstructed for a given level of compression.

As our second case study demonstrates, neglecting the important role played by temporal persistence in dynamically relevant features can lead to representations that are difficult to interpret and may not be as effective for studying persistent states. Clustering-based approaches, or more generally methods that attempt to approximate modes in the PDF rather than targeting the tails of the distribution, are likely to be a better choice in these circumstances. This can also be achieved by appropriate regularization so as to reduce sensitivity to transient features or outliers, which otherwise drive the definition of the basis in methods such as AA. In all of the methods that we have considered, a lack of independence in time is not explicitly modeled. Extensions to the simple methods that we have considered to account for non-independence are also possible, albeit usually at the cost of increased complexity. Singular spectrum analysis (see, e.g., Jolliffe, 1986) is a familiar example of one such extension for PCA. Similar generalizations can also be constructed for convex decompositions of the data. For instance, by virtue of the decomposability of the least squares cost function, it is possible to construct a joint convex discretization of instantaneous and lagged values of the variables of interest, which forms the basis of the scalable probabilistic approximation method (Gerber et al., 2020). In this approach, the decomposition may be parameterized in terms of a transition matrix relating the weights at different times, thus naturally incorporating a temporal constraint into the discretization. An associated regularization parameter allows the amount of temporal regularization to be appropriately tuned. Moreover, because the optimization problem remains separable in this case, the individual optimization steps can be parallelized and the method remains scalable. More sophisticated regularization strategies (Horenko, 2020) can further improve upon the performance of the method, even in the case of relatively small sample sizes and feature degeneracies.

The idea of imposing temporal regularization via assumed dynamics
for the latent weights suggests that another approach to better target
particular features is to start from an appropriately defined generative model or otherwise explicitly incorporate appropriate time dependence when constructing a reduction method. An underlying probabilistic model is already suggested by the
stochastic constraints that are imposed on the weights in AA and in convex coding,
a feature that is already taken advantage of in the case of the scalable
probabilistic approximation.
Corresponding latent variable models can naturally be constructed for
PCA (Roweis, 1998; Tipping and Bishop, 1999) and AA (Seth and Eugster, 2016). From the point of
view of incorporating temporal dependence between samples, starting from a
probabilistic model is fruitful as it is conceptually straightforward to
incorporate a model for the dynamics of the latent weights *z*_{t}
(Lawrence, 2005; Wang et al., 2006; Damianou et al., 2011).
Taken together with a choice of conditional distribution for the observations,
point estimation in the latent variable model is once again achieved by
optimization of a suitable loss function. Regularization is in this context
provided by suitable choice of prior distributions for the dictionary and
weights. An additional advantage of starting from a generative model is the
possibility of applying the full machinery of Bayesian inference rather than
obtaining only point estimates (Mnih and Salakhutdinov, 2008; Salakhutdinov and Mnih, 2008; Virtanen et al., 2008; Shan and Banerjee, 2010; Gönen et al., 2013), although scaling such analyses remains a challenging
issue. While this approach appears to be natural for constructing dimension
reduction methods that may flexibly take into account more complicated
dependence structures, it is by no means guaranteed to provide improved
low-dimensional representations of the data and likely depends heavily on choices such as the assumed generative process for the weights. For example,
in simple constructions with weights drawn according to a Gaussian process,
in the absence of strong prior information the fitted bases are driven
to be very similar to those found by ordinary PCA in order to maximize the
likelihood of the observed data, while at the same time being substantially more expensive to fit. Thus, the development of temporally regularized
methods derived from underlying stochastic models remains a topic for further
investigation.

Appendix A: Numerical solution for convex coding dictionary and archetypes

Back to toptop
As noted in Sect. 2, the optimization problem
posed in either the convex coding case or by AA cannot
be solved analytically. Moreover, the combined
cost function is not convex in the full set of variables **W** and **Z**
(or **C** and **Z** for AA). It is, though, convex in either **W** or **Z** separately,
when the other is held fixed, and local stationary points may be
straightforwardly found by alternating updates of the basis and weights. For
instance, using the penalty function Eq. (12), we may
update **W** with fixed **Z** via

$$\begin{array}{}\text{(A1)}& \mathbf{W}\leftarrow {\mathbf{X}}^{T}\mathbf{Z}{\left[{\mathbf{Z}}^{T}\mathbf{Z}+{\displaystyle \frac{\mathrm{4}T{\mathit{\lambda}}_{W}}{dk(k-\mathrm{1})}}\left(k{\mathbf{I}}_{k\times k}-{\mathbf{1}}_{k\times k}\right)\right]}^{-\mathrm{1}}.\end{array}$$

For fixed **W**, **Z** may then be updated by projected gradient descent.
An alternative that avoids having to perform direct projection onto the
simplex (Mørup and Hansen, 2012) is to reparameterize the latent weights **Z** as

$$\begin{array}{}\text{(A2)}& {Z}_{ti}={\displaystyle \frac{{H}_{ti}}{{\sum}_{i=\mathrm{1}}^{k}{H}_{ti}}},\phantom{\rule{1em}{0ex}}{H}_{ti}\ge \mathrm{0},\end{array}$$

which automatically satisfies the stochastic constraint on ** z**(

$$\begin{array}{}\text{(A3a)}& {\displaystyle}\begin{array}{rl}{\mathrm{\nabla}}_{W}F(\mathbf{W},\mathbf{Z};\mathbf{X})& \leftarrow -{\displaystyle \frac{\mathrm{1}}{T}}\left({\mathbf{X}}^{T}\mathbf{Z}\right)\\ & +{\displaystyle \frac{\mathrm{1}}{T}}\mathbf{W}\left[{\mathbf{Z}}^{T}\mathbf{Z}+{\mathit{\lambda}}_{W}{\mathbf{G}}_{W}\right],\end{array}\text{(A3b)}& {\displaystyle}\mathbf{W}\leftarrow \mathbf{W}-{\mathit{\eta}}_{W}{\mathrm{\nabla}}_{W}F(\mathbf{W},\mathbf{Z};\mathbf{X}),\text{(A3c)}& {\displaystyle}{\mathrm{\nabla}}_{Z}F(\mathbf{W},\mathbf{Z};\mathbf{X})\leftarrow -{\displaystyle \frac{\mathrm{1}}{T}}X\mathbf{W}+{\displaystyle \frac{\mathrm{1}}{T}}{\mathbf{ZW}}^{T}\mathbf{W},\text{(A3d)}& {\displaystyle}\begin{array}{rl}{H}_{ti}& \leftarrow max\mathit{\{}{Z}_{ti}-{\mathit{\eta}}_{H}[\left({\mathrm{\nabla}}_{Z}F\right(\mathbf{W},\mathbf{Z};\mathbf{X}){)}_{ti}\\ & -\sum _{j=\mathrm{1}}^{k}{Z}_{tj}\left({\mathrm{\nabla}}_{Z}F\right(\mathbf{W},\mathbf{Z};\mathbf{X}\left){)}_{tj}\right],\mathrm{0}\mathit{\}},\end{array}\text{(A3e)}& {\displaystyle}{Z}_{ti}\leftarrow {\displaystyle \frac{{H}_{ti}}{{\sum}_{i=\mathrm{1}}^{k}{H}_{ti}}},\end{array}$$

where, for simplicity, we have assumed that the derivative of the
penalty term Φ_{W} may be written in the form

$$\frac{\partial {\mathrm{\Phi}}_{W}}{\partial \mathbf{W}}}={\displaystyle \frac{\mathrm{1}}{T}}{\mathbf{WG}}_{W},$$

which is the case if, for instance, Φ_{W}∝Tr[**WG**_{W}**W**^{T}] for some symmetric matrix **G**_{W}. The step-size parameters
*η*_{W} and *η*_{H} may either be fixed or determined by performing
a line search. This procedure may be iterated until the successive changes in the total cost fall below a given tolerance. As convergence to a
global minimum is not guaranteed, in practice this procedure is repeated
for multiple initial guesses for **W** and **Z** to try to improve the
likelihood of locating the optimal solution. Inspecting the update
equations Eq. (A3), the cost per iteration is seen
to scale as $O\left(dTk\right)+O\left({k}^{\mathrm{2}}T\right)+O\left({k}^{\mathrm{2}}d\right)+O\left(kd\right)+O\left(kT\right)$. In particular,
in the usual case of interest with $d,T\gg k$, the leading contribution
to the cost is linear in each of *d*, *T*, and *k*, i.e., *O*(*d**T**k*), making
the method suitable for large datasets and comparable with *k*-means
and similar decompositions (Mørup and Hansen, 2012; Gerber et al., 2020).

Code and data availability

Back to toptop
Code and data availability.

The HadISST SST dataset used in this study is provided by the
UK Met Office Hadley Centre and may be accessed at
https://www.metoffice.gov.uk/hadobs/hadisst/ (Rayner et al., 2003).
The JRA-55 geopotential height data used are made available through the JRA-55 project and may be accessed following the procedures described at
https://jra.kishou.go.jp/JRA-55/index_en.html (Kobayashi et al., 2015).
All source code used to perform the analyses presented in the main text may
be found at https://doi.org/10.5281/zenodo.3723948 (Harries and O'Kane, 2020). The PCA and *k*-means results and the visualizations using metric MDS were obtained using the routines provided by the scikit-learn Python package
(Pedregosa et al., 2011). Plots were generated using Python package Matplotlib (Hunter, 2007).

Author contributions

Back to toptop
Author contributions.

All of the authors designed the study. TJO'K proposed the specific case studies, and DH implemented the code to perform the numerical comparisons and generated all plots and figures. All of the authors contributed to the direction of the study, discussion of results, and the writing and approval of the manuscript.

Competing interests

Back to toptop
Competing interests.

The authors declare that they have no conflict of interest.

Acknowledgements

Back to toptop
Acknowledgements.

The authors would like to thank Illia Horenko for continual guidance and for his valuable inputs over the course of this project and Vassili Kitsios and Didier Monselesan for encouragement and helpful discussions about the various methods.

Financial support

Back to toptop
Financial support.

This research has been supported by the Australian Commonwealth Scientific and Industrial Research Organisation (CSIRO) through a ResearchPlus postdoctoral fellowship and by the CSIRO Decadal Climate Forecasting Project.

Review statement

Back to toptop
Review statement.

This paper was edited by Zoltan Toth and reviewed by two anonymous referees.

References

Back to toptop
Aloise, D., Deshpande, A., Hansen, P., and Popat, P.: NP-hardness of Euclidean sum-of-squares clustering, Mach. Learn., 75, 245–248, https://doi.org/10.1007/s10994-009-5103-0, 2009. a

Arbelaitz, O., Gurrutxaga, I., Muguerza, J., Pérez, J. M., and Perona, I.: An extensive comparative study of cluster validity indices, Pattern Recognition, 46, 243–256, 2013. a

Banerjee, A., Merugu, S., Dhillon, I. S., and Ghosh, J.: Clustering with Bregman Divergences, J. Mach. Learn. Res., 6, 1705–1749, https://doi.org/10.1007/s10994-005-5825-6, 2005. a

Barnston, A. G. and Livezey, R. E.: Classification, Seasonality and Persistence of Low-Frequency Atmospheric Circulation Patterns, Mon. Weather Rev., 115, 1083–1126, https://doi.org/10.1175/1520-0493(1987)115<1083:csapol>2.0.co;2, 1987. a

Barriopedro, D., Fischer, E. M., Luterbacher, J., Trigo, R. M., and García-Herrera, R.: The Hot Summer of 2010: Redrawing the Temperature Record Map of Europe, Science, 332, 220–224, https://doi.org/10.1126/science.1201224, 2011. a

Bezdek, J. C., Ehrlich, R., and Full, W.: FCM: The Fuzzy c-Means Clustering Algorithm, Comput. Geosci., 10, 191–203, https://doi.org/10.1109/igarss.1988.569600, 1984. a

Bregman, L.: The relaxation method of finding the common point of convex sets and its application to the solution of problems in convex programming, USSR Comp. Math. Math+, 7, 200–217, https://doi.org/10.1016/0041-5553(67)90040-7, 1967. a

Bueh, C. and Nakamura, H.: Scandinavian pattern and its climatic impact, Q. J. Roy. Meteor. Soc., 133, 2117–2131, https://doi.org/10.1002/qj.173, 2007. a

Cheng, X. and Wallace, J. M.: Cluster Analysis of the Northern Hemisphere Wintertime 500-hPa Height Field: Spatial Patterns, J. Atmos. Sci., 50, 2674–2696, https://doi.org/10.1175/1520-0469(1993)050<2674:CAOTNH>2.0.CO;2, 1993. a

Christiansen, B.: Atmospheric Circulation Regimes: Can Cluster Analysis Provide the Number?, J. Climate, 20, 2229–2250, https://doi.org/10.1175/JCLI4107.1, 2007. a

Cutler, A. and Breiman, L.: Archetypal Analysis, Technometrics, 36, 338–347, 1994. a

Damianou, A. C., Titsias, M. K., and Lawrence, N. D.: Variational Gaussian Process Dynamical Systems, in: Advances in Neural Information Processing Systems 24 (NIPS 2011), 12–17 December 2011, Granada, Spain, 2510–2518, 2011. a

Ding, C. and He, X.: K-means clustering via principal component analysis, in: Proceedings of the twenty-first international conference on Machine learning (ICML 2004), 4–8 July 2004, Banff, Canada, 29–37, 2004. a

Dole, R. M., Hoerling, M., Perlwitz, J., Eischeid, J., Pegion, P., Zhang, T., Quan, X.-W., Xu, T., and Murray, D.: Was there a basis for anticipating the 2010 Russian heat wave?, Geophys. Res. Lett., 38, L06702, https://doi.org/10.1029/2010GL046582, 2011. a, b

Dole, R. M. and Gordon, N. D.: Persistent Anomalies of the Extratropical Northern Hemisphere Wintertime Circulation: Geographical Distribution and Regional Persistence Characteristics, Mon. Weather Rev., 111, 1567–1586, https://doi.org/10.1175/1520-0493(1983)111<1567:PAOTEN>2.0.CO;2, 1983. a, b

Dommenget, D. and Latif, M.: A cautionary note on the interpretation of EOFs, J. Climate, 15, 216–225, https://doi.org/10.1175/1520-0442(2002)015<0216:ACNOTI>2.0.CO;2, 2002. a

Dunn, J. C.: A fuzzy relative of the ISODATA process and its use in detecting compact well-separated clusters, J. Cybernetics, 3, 32–57, https://doi.org/10.1080/01969727308546046, 1973. a

Eckart, C. and Young, G.: The approximation of one matrix by another of lower rank, Psychometrika, 1, 211–218, https://doi.org/10.1007/BF02288367, 1936. a

Efimov, V., Prusov, A., and Shokurov, M.: Patterns of interannual variability defined by a cluster analysis and their relation with ENSO, Q. J. Roy. Meteor. Soc., 121, 1651–1679, 1995. a

Fereday, D. R., Knight, J. R., Scaife, A. A., Folland, C. K., and Philipp, A.: Cluster Analysis of North Atlantic–European Circulation Types and Links with Tropical Pacific Sea Surface Temperatures, J. Climate, 21, 3687–3703, https://doi.org/10.1175/2007JCLI1875.1, 2008. a

Forgey, E.: Cluster analysis of multivariate data: Efficiency vs. interpretability of classification, Biometrics, 21, 768–769, 1965. a

Gerber, S., Pospisil, L., Navandar, M., and Horenko, I.: Low-cost scalable discretization, prediction, and feature selection for complex systems, Sci. Adv., 6, eaaw0961, https://doi.org/10.1126/sciadv.aaw0961, 2020. a, b, c, d, e, f

Gönen, M., Khan, S., and Kaski, S.: Kernelized Bayesian matrix factorization, in: Proceedings of the 30th International Conference on Machine Learning (ICML 2013), 17–19 June 2013, Atlanta, USA, 864–872, 2013. a

Hannachi, A. and Legras, B.: Simulated annealing and weather regimes classification, Tellus A, 47, 955–973, https://doi.org/10.1034/j.1600-0870.1995.00203.x, 1995. a

Hannachi, A. and Trendafilov, N.: Archetypal analysis: Mining weather and climate extremes, J. Climate, 30, 6927–6944, https://doi.org/10.1175/JCLI-D-16-0798.1, 2017. a, b, c, d

Hannachi, A., Jolliffe, I. T., and Stephenson, D. B.: Empirical orthogonal functions and related techniques in atmospheric science: A review, Int. J. Climatol., 27, 1119–1152, https://doi.org/10.1002/joc.1499, 2007. a

Harada, Y., Kamahori, H., Kobayashi, C., Endo, H., Kobayashi, S., Ota, Y., Onoda, H., Onogi, K., Miyaoka, K., and Takahashi, K.: The JRA-55 Reanalysis: Representation of Atmospheric Circulation and Climate Variability, J. Meteorol. Soc. Jpn. Ser. II, 94, 269–302, https://doi.org/10.2151/jmsj.2016-015, 2016. a

Harries, D. and O'Kane, T. J.: Matrix factorization case studies code, Zenodo, https://doi.org/10.5281/zenodo.3723948, 2020. a

Hartigan, J. A. and Wong, M. A.: Algorithm AS 136: A K-Means Clustering Algorithm, J. Roy. Stat. Soc. C-App., 28, 100–108, https://doi.org/10.2307/2346830, 1979. a

Hastie, T., Tibshirani, R., and Friedman, J.: The Elements of Statistical Learning: Data Mining, Inference and Prediction, Springer, New York, USA, 2005. a

Horenko, I.: On a scalable entropic breaching of the overfitting barrier in machine learning, Neural Computation, arXiv [preprint], arXiv:2002.03176, 8 February 2020. a

Hunter, J. D.: Matplotlib: A 2D graphics environment, Comput. Sci. Eng., 9, 90–95, https://doi.org/10.1109/MCSE.2007.55, 2007. a

Huth, R., Beck, C., Philipp, A., Demuzere, M., Ustrnul, Z., Cahynová, M., Kyselý, J., and Tveito, O. E.: Classifications of Atmospheric Circulation Patterns, Ann. NY Acad. Sci., 1146, 105–152, https://doi.org/10.1196/annals.1446.019, 2008. a

Jenatton, R., Obozinski, G., and Bach, F.: Structured Sparse Principal Component Analysis, in: Proceedings of the Thirteenth International Conference on Artificial Intelligence and Statistics, 13–15 May 2010, Sardinia, Italy, 366–373, 2010. a, b

Jolliffe, I.: Principal component analysis, Springer Verlag, New York, USA, 1986. a, b

Jolliffe, I. T., Trendafilov, N. T., and Uddin, M.: A Modified Principal Component Technique Based on the LASSO, J. Comput. Graph. Stat., 12, 531–547, https://doi.org/10.1198/1061860032148, 2003. a, b

Kaiser, E., Noack, B. R., Cordier, L., Spohn, A., Segond, M., Abel, M., Daviller, G., Östh, J., Krajnović, S., and Niven, R. K.: Cluster-based reduced-order modelling of a mixing layer, J. Fluid Mech., 754, 365–414, 2014. a

Kaiser, H. F.: The varimax criterion for analytic rotation in factor analysis, Psychometrika, 23, 187–200, https://doi.org/10.1007/BF02289233, 1958. a

Kidson, J. W.: The Utility Of Surface And Upper Air Data In Synoptic Climatological Specification Of Surface Climatic Variables, Int. J. Climatol., 17, 399–413, https://doi.org/10.1002/(SICI)1097-0088(19970330)17:4<399::AID-JOC108>3.0.CO;2-M, 1997. a

Kidson, J. W.: An analysis of New Zealand synoptic types and their use in defining weather regimes, Int. J. Climatol., 20, 299–316, https://doi.org/10.1002/(SICI)1097-0088(20000315)20:3<299::AID-JOC474>3.0.CO;2-B, 2000. a

Kobayashi, S., Ota, Y., Harada, Y., Ebita, A., Moriya, M., Onoda, H., Onogi, K., Kamahori, H., Kobayashi, C., Endo, H., Miyaoka, K., and Takahashi, K.: The JRA-55 Reanalysis: General Specifications and Basic Characteristics, J. Meteorol. Soc. Jpn. Ser. II, 93, 5–48, https://doi.org/10.2151/jmsj.2015-001, 2015 (data available at: https://jra.kishou.go.jp/JRA-55/index_en.html, last access: 12 April 2019). a, b

Lau, K.-M., Sheu, P.-J., and Kang, I.-S.: Multiscale Low-Frequency Circulation Modes in the Global Atmosphere, J. Atmos. Sci., 51, 1169–1193, 1994. a

Lawrence, N.: Probabilistic Non-linear Principal Component Analysis with Gaussian Process Latent Variable Models, J. Mach. Learn. Res., 6, 1783–1816, 2005. a

Lee, D. D. and Seung, H. S.: Unsupervised Learning by Convex and Conic Coding, in: Advances in Neural Information Processing Systems 9 (NIPS 1996), 2–5 December 1996, Denver, USA, 515–521, 1997. a, b

Lee, H., Battle, A., Raina, R., and Ng, A. Y.: Efficient Sparse Coding Algorithms, in: Advances in Neural Information Processing Systems 19 (NIPS 2006), 4–7 December 2006, Vancouver, Canada, 801–808, 2007. a, b

Legras, B., Desponts, T., and Piguet, B.: Cluster analysis and weather regimes, in: Seminar on the Nature and Prediction of Extra Tropical Weather Systems, 7–11 September 1987, Reading, UK, 123–150, 1987. a

Li, T. and Ding, C.: The Relationships Among Various Nonnegative Matrix Factorization Methods for Clustering, in: Proceedings of the Sixth International Conference on Data Mining (ICDM'06), 18–22 December 2006, Hong Kong, China, 362–371, 2006. a

Lloyd, S.: Least squares quantization in PCM, IEEE T. Inform. Theory, 28, 129–137, https://doi.org/10.1109/TIT.1982.1056489, 1982. a, b

Lorenz, E. N.: Empirical Orthogonal Functions and Statistical Weather Prediction, Tech. rep., Massachusetts Institute of Technology, Cambridge, UK, 1956. a

MacKay, D. J. C.: Information theory, inference and learning algorithms, Cambridge University Press, Cambridge, UK, 2003. a

MacQueen, J.: Some methods for classification and analysis of multivariate observations, in: Proceedings of the Fifth Berkeley Symposium on Mathematical Statistics and Probability, Volume 1: Statistics, 21 June–18 July 1965 and 27 December 1965–7 January 1966, Berkeley, USA, 281–297, 1967. a

Mahajan, M., Nimbhorkar, P., and Varadarajan, K.: The planar k-means problem is NP-hard, Theor. Comput. Sci., 442, 13–21, https://doi.org/10.1016/j.tcs.2010.05.034, 2012. a

Mairal, J., Bach, F., Ponce, J., and Sapiro, G.: Online Dictionary Learning for Sparse Coding, in: Proceedings of the Twenty-Sixth International Conference on Machine Learning, 14–18 June 2009, Montreal, Canada, 689–696, 2009. a, b

Matsueda, M.: Predictability of Euro-Russian blocking in summer of 2010, Geophys. Res. Lett., 38, L06801, https://doi.org/10.1029/2010GL046557, 2011. a

Michelangeli, P.-A., Vautard, R., and Legras, B.: Weather Regimes: Recurrence and Quasi Stationarity, J. Atmos. Sci., 52, 1237–1256, https://doi.org/10.1175/1520-0469(1995)052<1237:WRRAQS>2.0.CO;2, 1995. a, b

Mnih, A. and Salakhutdinov, R. R.: Probabilistic Matrix Factorization, in: Advances in Neural Information Processing Systems 20 (NIPS 2007), 3–6 December 2007, Vancouver, Canada, 1257–1264, 2008. a

Mo, K. and Ghil, M.: Cluster analysis of multiple planetary flow regimes, J. Geophys. Res.-Atmos., 93, 10927–10952, https://doi.org/10.1029/JD093iD09p10927, 1988. a, b, c

Molteni, F., Tibaldi, S., and Palmer, T. N.: Regimes in the wintertime circulation over northern extratropics. I: Observational evidence, Q. J. Roy. Meteor. Soc., 116, 31–67, https://doi.org/10.1002/qj.49711649103, 1990. a

Monahan, A. H., Fyfe, J. C., Ambaum, M. H. P., Stephenson, D. B., and North, G. R.: Empirical Orthogonal Functions: The Medium is the Message, J. Climate, 22, 6501–6514, https://doi.org/10.1175/2009JCLI3062.1, 2009. a

Mørup, M. and Hansen, L. K.: Archetypal analysis for machine learning and data mining, Neurocomputing, 80, 54–63, https://doi.org/10.1016/j.neucom.2011.06.033, 2012. a, b, c, d, e

Neal, R., Fereday, D., Crocker, R., and Comer, R. E.: A flexible approach to defining weather patterns and their application in weather forecasting over Europe, Meteorol. Appl., 23, 389–400, https://doi.org/10.1002/met.1563, 2016. a

Pedregosa, F., Varoquaux, G., Gramfort, A., Michel, V., Thirion, B., Grisel, O., Blondel, M., Prettenhofer, P., Weiss, R., Dubourg, V., Vanderplas, J., Passos, A., Cournapeau, D., Brucher, M., Perrot, M., and Duchesnay, E.: Scikit-learn: Machine Learning in Python, J. Mach. Learn. Res., 12, 2825–2830, 2011. a

Pelly, J. L. and Hoskins, B. J.: A New Perspective on Blocking, J. Atmos. Sci., 60, 743–755, https://doi.org/10.1175/1520-0469(2003)060<0743:ANPOB>2.0.CO;2, 2003. a

Pohl, B. and Fauchereau, N.: The Southern Annular Mode Seen through Weather Regimes, J. Climate, 25, 3336–3354, https://doi.org/10.1175/JCLI-D-11-00160.1, 2012. a

Rayner, N. A., Parker, D. E., Horton, E. B., Folland, C. K., Alexander, L. V., Rowell, D. P., Kent, E. C., and Kaplan, A.: Global analyses of sea surface temperature, sea ice, and night marine air temperature since the late nineteenth century, J. Geophys. Res.-Atmos., 108, 4407, https://doi.org/10.1029/2002JD002670, 2003 (data available at: https://www.metoffice.gov.uk/hadobs/hadisst/, last access: 29 April 2019). a, b

Renwick, J. A.: Persistent Positive Anomalies in the Southern Hemisphere Circulation, Mon. Weather Rev., 133, 977–988, https://doi.org/10.1175/MWR2900.1, 2005. a

Richman, M. B.: Rotation of Principal Components, J. Climatol., 6, 293–335, https://doi.org/10.1177/1746847713485834, 1986. a

Roweis, S. T.: EM algorithms for PCA and SPCA, in: Advances in Neural Information Processing Systems 10 (NIPS 1997), 1–6 December 1997, Denver, USA, 626–632, 1998. a, b

Ruspini, E. H.: A New Approach to Clustering, Inform. Control, 15, 22–32, 1969. a

Salakhutdinov, R. and Mnih, A.: Bayesian Probabilistic Matrix Factorization using Markov Chain Monte Carlo, in: Proceedings of the Twenty-Fifth International Conference on Machine Learning (ICML'08), 5–9 July 2008, Helsinki, Finland, 880–887, 2008. a

Seth, S. and Eugster, M. J.: Probabilistic archetypal analysis, Mach. Learn., 102, 85–113, https://doi.org/10.1007/s10994-015-5498-8, 2016. a, b, c

Shan, H. and Banerjee, A.: Generalized Probabilistic Matrix Factorizations for Collaborative Filtering, in: Proceedings of the Tenth IEEE International Conference on Data Mining, 14–17 December 2010, Sydney, Australia, 1025–1030, 2010. a

Shaposhnikov, D., Revich, B., Bellander, T., Bedada, G. B., Bottai, M., Kharkova, T., Kvasha, E., Lezina, E., Lind, T., Semutnikova, E., and Pershagen, G.: Mortality related to air pollution with the Moscow heat wave and wildfire of 2010, Epidemiology, 25, 359–364, https://doi.org/10.1097/EDE.0000000000000090, 2014. a

Singh, A. P. and Gordon, G. J.: A Unified View of Matrix Factorization Models, in: Proceedings of the Joint European Conference on Machine Learning and Knowledge Discovery in Databases, Part II, 15–19 September 2008, Antwerp, Belgium, 358–373, 2008. a, b

Steinschneider, S. and Lall, U.: Daily Precipitation and Tropical Moisture Exports across the Eastern United States: An Application of Archetypal Analysis to Identify Spatiotemporal Structure, J. Climate, 28, 8585–8602, https://doi.org/10.1175/JCLI-D-15-0340.1, 2015. a

Stone, E. and Cutler, A.: Introduction to archetypal analysis of spatio-temporal dynamics, Physica D, 96, 110–131, https://doi.org/10.1016/0167-2789(96)00016-4, 1996. a

Stone, R. C.: Weather types at Brisbane, Queensland: An example of the use of principal components and cluster analysis, Int. J. Climatol., 9, 3–32, https://doi.org/10.1002/joc.3370090103, 1989. a

Straus, D. M., Corti, S., and Molteni, F.: Circulation Regimes: Chaotic Variability versus SST-Forced Predictability, J. Climate, 20, 2251–2272, https://doi.org/10.1175/JCLI4070.1, 2007. a

Tibshirani, R., Walther, G., and Hastie, T.: Estimating the number of clusters in a data set via the gap statistic, J. Roy. Stat. Soc. B, 63, 411–423, https://doi.org/10.1111/1467-9868.00293, 2001. a, b

Tipping, M. E. and Bishop, C. M.: Probabilistic Principal Component Analysis, J. Roy. Stat. Soc. B, 61, 611–622, https://doi.org/10.1111/1467-9868.00196, 1999. a, b

Virtanen, T., Cemgil, A. T., and Godsill, S.: Bayesian extensions to non-negative matrix factorisation for audio signal modelling, in: Proceedings of the 2008 IEEE International Conference on Acoustics, Speech, and Signal Processing, 30 March–4 April 2008, Las Vegas, USA, 1825–1828, 2008. a

Wang, C., Xie, S.-P., and Carton, J. A.: A Global Survey of Ocean–Atmosphere Interaction and Climate Variability, American Geophysical Union (AGU), 1–19, https://doi.org/10.1029/147GM01, 2004. a

Wang, J., Hertzmann, A., and Fleet, D. J.: Gaussian Process Dynamical Models, in: Advances in Neural Information Processing Systems 18 (NIPS 2005), 5–8 December 2005, Vancouver, Canada, 1441–1448, 2006. a

Witten, D. M., Tibshirani, R., and Hastie, T.: A penalized matrix decomposition, with applications to sparse principal components and canonical correlation analysis, Biostatistics, 10, 515–534, https://doi.org/10.1093/biostatistics/kxp008, 2009. a, b

A simple example
would be a naïve analysis of noisy annular data, for which *k*-means may
give rise to centroids lying within the annulus; as noted above,
in general the success of all of the methods considered in the following will
depend on the underlying topology of the data.

Probabilistic formulations of PCA and AA have also been
proposed (Roweis, 1998; Tipping and Bishop, 1999; Seth and Eugster, 2016), in which the problem is
formulated as inference under an appropriate latent variable model; similarly,
soft *k*-means is well known to be closely related to Gaussian mixture modeling, e.g., MacKay (2003). For simplicity, in the following
comparisons we will only consider the deterministic formulations of the
methods.

In the following,
we use notation appropriate for a time series of observations with separate samples indexed by time *t*, but of course the discussion is not limited to
this case.

The
ℓ_{p}-norm of a vector ** x**∈ℝ

See Ding and He (2004) for additional
discussion of the relationship between PCA and *k*-means.

Of course, in practice proper estimates of the out-of-sample performance would be obtained by an appropriate cross-validation procedure or similar. However, as here we are primarily interested in the qualitative differences between the different methods in terms of extracting recognizable states, we do not focus on the technical details of model selection or optimizing predictive performance and simply present these out-of-sample estimates to show the general features of each method.

We note that, when using a null model generated by PCA, the gap curve in Fig. 4 might be considered to indicate possible clustering into five clusters within the single, large cluster. However, this conclusion depends strongly on the precise null model used to generate the reference data.

Similar remarks can be made
for the AA/PCH-*δ* model proposed in Mørup and Hansen (2012).

Clustering on the PCs was done so as to reduce the overall cost of the methods; we have checked that, for small numbers of clusters, the spatial patterns that result are very similar.

Similar behavior is
evident for different numbers of states; e.g., for *k*=3 the MDS projection results in a triangular arrangement of points with the climatological point
located close to the centroid of the resulting shape.

Short summary

Different dimension reduction methods may produce profoundly different low-dimensional representations of multiscale systems. We perform a set of case studies to investigate these differences. When a clear scale separation is present, similar bases are obtained using all methods, but when this is not the case some methods may produce representations that are poorly suited for describing features of interest, highlighting the importance of a careful choice of method when designing analyses.

Different dimension reduction methods may produce profoundly different low-dimensional...

Nonlinear Processes in Geophysics

An interactive open-access journal of the European Geosciences Union