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Sequence analysis is an increasingly popular approach for the analysis of life-courses represented by categorical sequences, i.e. as the ordered collection of activities experienced by subjects over a given time period. Several criteria have been introduced in the literature to measure pairwise dissimilarities among sequences. Typically, dissimilarity matrices are employed as the input to heuristic clustering algorithms, with the aim of identifying the most relevant patterns in the data. Here, we propose a model-based clustering approach for categorical sequence data. The technique is applied to a popular survey data set containing information on the career trajectories, in terms of monthly labour market activities, of a cohort of Northern Irish youths tracked from the age of 16 to the age of 22. Specifically, we develop a family of methods for clustering sequence data directly based on mixtures of exponential-distance models, which we call MEDseq. The Hamming distance and weighted variants thereof are employed as the distance metric. The existence of a closed-form expression for the normalising constant using these metrics facilitates the development of an ECM algorithm for model fitting. We allow the probability of component membership to depend on fixed covariates. The MEDseq models can also accommodate sampling weights, which are typically associated with life-course data. Including the weights and covariates in the clustering process in a holistic manner allows new insights to be gleaned from the Northern Irish data.
arXiv:1908.07963v3 [stat.ME] 13 Jul 2021
This is a preprint. The revised version of this paper is published as
K. Murphy, T. B. Murphy, R. Piccarreta, and I. C. Gormley (2021) “Clustering longitudinal
life-course sequences using mixtures of exponential-distance models”. Journal of the Royal
Statistical Society: Series A (Statistics in Society). [doi: 10.1111/rssa.12712].
Clustering Longitudinal Life-Course Sequences
Using Mixtures of Exponential-Distance Models
Keefe Murphy1, T. Brendan Murphy2,3,
Raffaella Piccarreta4, I. Claire Gormley2,3
1Department of Mathematics and Statistics, Maynooth University, Ireland
2School of Mathematics and Statistics, University College Dublin, Ireland
3Insight Centre for Data Analytics, University College Dublin, Ireland
4Department of Decision Sciences, Università Bocconi, Milano, Italy
Sequence analysis is an increasingly popular approach for analysing life courses rep-
resented by ordered collections of activities experienced by subjects over time. Here,
we analyse a survey data set containing information on the career trajectories of a
cohort of Northern Irish youths tracked between the ages of 16 and 22. We propose
a novel, model-based clustering approach suited to the analysis of such data from a
holistic perspective, with the aims of estimating the number of typical career trajec-
tories, identifying the relevant features of these patterns, and assessing the extent to
which such patterns are shaped by background characteristics.
Several criteria exist for measuring pairwise dissimilarities among categorical se-
quences. Typically, dissimilarity matrices are employed as input to heuristic clustering
algorithms. The family of methods we develop instead clusters sequences directly using
mixtures of exponential-distance models. Basing the models on weighted variants of
the Hamming distance metric permits closed-form expressions for parameter estima-
tion. Simultaneously allowing the component membership probabilities to depend on
fixed covariates and accommodating sampling weights in the clustering process yields
new insights on the Northern Irish data. In particular, we find that school examination
performance is the single most important predictor of cluster membership.
Keywords: exponential-distance models, gating covariates, life-course sequences, model-
based clustering, survey sampling weights, weighted Hamming distance.
1 Introduction
Sequence analysis (SA) is an umbrella term for tools defined to explore and describe categor-
ical life-course data. Specifically, attention is focused on the ordered sequence of states (or
activities) experienced by individuals over a given time-span (usually at Tequally spaced
discrete time periods). Here we focus on the transition from school to work for a cohort
of Northern Irish youths, using survey data obtained from the 1999 sweep of the Status
Zero Survey (McVicar,2000;McVicar and Anyadike-Danes,2002) — henceforth referred to
as the MVAD data — in which, for each individual, a sequence of monthly labour market
activities experienced between the ages of 16 and 22 is recorded.
Typically, the goal of sequence analysis is to identify the most relevant patterns in the
data. To this end, pairwise dissimilarities among sequences in their entirety are first assessed.
Dissimilarity matrices are then employed to identify the most typical trajectories using, in
the vast majority of applications, cluster analysis. These problems are receiving increasing
attention in the demographic and social literature, also due to the increasing number of
retrospective as well as prospective longitudinal studies, such as the British Household
Panel Survey (BHPS)1and the subsequent larger and more wide-ranging UK Household
Longitudinal Study (Understanding Society)2, or the Socio-Economic Panel for Germany
(SOEP)3, the National Longitudinal Surveys for the USA (NLS)4, and the Generations &
Gender Programme for selected European countries (GGP)5. All of these surveys, much like
the MVAD data considered in this paper, collect information about labour market activities,
as well as other significant life events.
Quantifying the distance between such categorical sequences is not a trivial task. Optimal
matching (OM), developed by Abbott and Forrest (1986) and extended to sociology by Ab-
bott and Hrycak (1990), is popular among the SA community. OM is derived from the
edit distance originally proposed in the field of information theory and computer science
by Levenshtein (1966). The OM metric assigns costs to the different types of edits, namely
insertion, deletion, and substitution. Typically, insertion and deletion are assigned a cost of
1while substitution costs are allowed to vary. However, specifying these costs often involves
subjective choices, which may lead to violations of the triangle inequality if not done carefully.
Several proposals in the literature introduced criteria to improve or guide the choice of costs
in OM; Muñoz-Bullón and Malo (2003), for instance, estimate the substitution-cost matrix
in a data-driven fashion using the between-states transition rates. Alternative dissimilarity
criteria have also been introduced to allow control over the importance assigned to the char-
acteristics of the sequences (namely, the collection of experienced states, their timing, or
their duration) in the assessment of their differences: see Studer and Ritschard (2016) for an
excellent discussion. Even so, there are no results proving that one procedure is superior to
others and the choice of dissimilarity measure remains a fundamental question for researchers.
Given a dissimilarity matrix Dobtained from a set of sequences S= (s1,...,sn), where n
is the number of subjects, cluster analysis is usually applied to group sequences and identify
the most typical trajectories experienced by the sampled individuals. Heuristic clustering
algorithms, either hierarchical or partitional, are typically used. In many applications, it
is also of interest to relate sequences to a set of baseline covariates. Within the described
framework, this is solely done by relating the uncovered hard clustering partition to covari-
ates using, for example, multinomial logistic regression (MLR). This approach was adopted
in McVicar and Anyadike-Danes (2002), after applying Ward’s agglomerative clustering al-
gorithm (Ward,1963) to an OM dissimilarity matrix to obtain G= 5 clusters of the MVAD
sequences, without performing model selection. Such an approach is questionable from a
few points of view. Firstly, the original sequences are substituted by a categorical variable
indicating cluster membership, thus disregarding the heterogeneity within clusters. This
is clearly only sensible when the clusters are sufficiently homogeneous otherwise sequences
which are weakly related to clusters would be regarded as similar to those in their cluster.
However, a clear clustering structure can often be obtained only by increasing the number
of clusters (often with some clusters possibly small in size). More importantly, suitable par-
titions do not necessarily lead to suitable response variables as input for the MLR. It thus
seems desirable to cluster sequences and relate the clusters to the covariates simultaneously.
Thus, the aim of our analysis is three-fold; to estimate the number of typical trajectories
in the MVAD data, to identify the relevant features of these patterns, and to establish to
what extent such patterns are shaped by the individuals’ background characteristics, as cap-
tured by a set of baseline covariates measured at age 16. To address these issues, we propose
to cluster the MVAD sequences in a model-based fashion, allowing the covariates to affect the
soft cluster membership probabilities, rather than leaving them exogenous to the clustering
model. This permits us to better understand if and to what extent the typical sequence pat-
terns characterising each cluster are affected by specific covariates. Model-based clustering
methods typically assume that the data arise from a finite mixture of Gdistributions;
Bouveyron et al. (2019) provide an excellent overview. In principle, any distribution(s)
can be used, though the term ‘model-based clustering’ was popularised by Banfield and
Raftery (1993), in which the component distributions are assumed to be parsimoniously
parameterised multivariate Gaussians with component-specific parameters. Such models
have been recently extended to the mixture of experts setting (Gormley and Frühwirth-
Schnatter,2019) to facilitate dependence on fixed covariates (Murphy and Murphy,2020).
However, these models can be problematic when applied to dissimilarity matrices, either due
to non-identifiability or because the input data are usually far from Gaussian. This prob-
lem cannot be addressed by applying multidimensional scaling to Dbecause the resulting
low-dimensional configuration is also typically far from Gaussian. Notably, our attempts to
fit non-Gaussian mixtures in these settings did not yield useful results.
Another popular framework for clustering categorical data is latent class analysis (LCA;
Lazarsfeld and Henry 1968)). Agresti (2002) shows the connection between model-based
clustering and LCA. Such models are finite mixtures in which the component distributions
are assumed to be multi-way cross-classification tables with all variables mutually indepen-
dent. Latent class regression models (LCR; Dayton and Macready 1988) are particularly
interesting, because their connection to the mixture of experts framework permits the in-
clusion of covariates to predict the latent class memberships. However, fitting such models
is challenging when the sequence length, the number of categories, or the number of latent
classes are even moderately large, due to the explosion in the number of parameters.
Evidently, there is a conflict of perspectives between the model-based and the heuristic,
distance-based approaches to clustering in the SA community. For this reason, and the others
mentioned above, we model the sequences directly (in the sense that the sequences them-
selves are treated as inputs, rather than D) with the implicit substitution costs which define
the distance metric being estimable parameters of a generative probability model rather than
inputs (either estimated or subjectively specified), via D, to a heuristic clustering algorithm.
This is achieved using parsimonious mixtures of exponential-distance models, which typi-
cally depend on a central sequence and a precision parameter in a way that relates to the
chosen distance metric. Our framework for analysing the MVAD data, as a model-based
approach which nonetheless relies on distances, thus reconciles the aforementioned conflict.
Mostly for reasons of computational convenience, we use dissimilarities based on sim-
ple matching, in particular the Hamming distance (Hamming,1950). Although the focus
on substitution operations has the sociological advantage of targeting trajectories with con-
temporaneous similarities — in contrast to the prohibited insertion and deletion operations,
which focus on matching states irrespective of their timing — this distance is liable to suffer
from temporal rigidity, since anticipations and/or postponements of the same choices in life
courses are not accounted for. Hence, similar sequences shifted by one time period may be
maximally distant from one another. While misalignment is less of a concern for sequences
exhibiting long durations in the same state, we address the issue using weighted variants of
the Hamming distance, characterised by a range of constraints on the precision parameters
in the mixture setting. This leads to the novel MEDseq model family, which can be seen
as similar to a version of the k-medoids/PAM algorithm (Kaufman and Rousseeuw,1990,
Chapter 2) based on the Hamming distance with some restrictions relaxed. We defer the
comparison to Section 4.2 as the parallels relate to technical issues of model estimation.
Importantly, information is also available with the MVAD data on the survey sampling
weights, which are only incorporated in the MLR stage of the analysis in McVicar and
Anyadike-Danes (2002). While sampling weights can be incorporated into heuristic cluster-
ing algorithms, such as Ward’s hierarchical clustering (by weighting the linkages between
clusters) or k-medoids, and subsequently in the MLR, one of the advantages of our ap-
proach is that both the covariates and the weights are incorporated simultaneously. This is
achieved by leveraging the model-based paradigm; the weights are incorporated by appro-
priately weighting the likelihood function and the covariates are incorporated by assuming
they influence the soft component membership probabilities.
MEDseq models, like standard SA heuristic clustering algorithms and LCA models, ap-
proach the clustering task from the holistic perspective of treating trajectories as whole units
of analysis, in order to uncover groups of similar sequences. In contrast, a number of mul-
tistate models employing finite mixtures with Markov components (e.g. Melnykov 2016a;
Pamminger and Frühwirth-Schnatter 2010) or with hidden Markov components (Helske
et al.,2016) have recently attained popularity for the analysis of categorical sequence data.
Such models focus on modelling instantaneous transitions within the life course and on fac-
tors that might explain the probability of experiencing them. As described by Wu (2000),
this amounts to a difference between considering sequences in their entirety under the MED-
seq framework or as time-to-event processes under the Markovian framework. Indeed, as
our aim is to establish sequence typologies for the MVAD data, a holistic approach is prefer-
able to Markovian approaches. The former concentrates on questions of global similarities
and considers the full richness of the trajectories without discarding the details of episode
ordering, duration, or transition (Muñoz-Bullón and Malo,2003), while the latter frame-
work makes the often unsuitable simplifying assumption that trajectories can be efficiently
summarised only by their recent past (Piccarreta and Studer,2019).
The remainder of the article is organised as follows. Section 2presents some ex-
ploratory analysis of the MVAD data. Section 3develops the MEDseq family of mixtures of
exponential-distance models. Section 4describes the model fitting procedure and discusses
factors affecting performance. Section 5presents results for the MVAD data, including appli-
cations of MEDseq models and comparisons to other methods. The insights gleaned from
the MVAD data under the optimal MEDseq model are summarised in Section 6. The paper
concludes with a discussion on the MEDseq methodology and potential future extensions in
Section 7. A software implementation of the full MEDseq model family is provided by the as-
sociated Rpackage MEDseq (Murphy et al.,2021). The package was developed specifically for
this application and is available from (RCore Team,2021).
2 Status Zero Survey: MVAD Data
The term ‘MVAD data’ refers throughout to a cohort of n= 712 Northern Irish youths
aged 16 and eligible to leave compulsory education as of July 1993 who were observed at
monthly intervals until June 1999 as part of the Status Zero Survey (Armstrong et al.,
1997;McVicar,2000;McVicar and Anyadike-Danes,2002). The subjects were interviewed
about the labour market activities they experienced, distinguishing between employment
(EM), further education (FE), higher education (HE), joblessness (JL), school (SC), or
training (TR). Each observation iis represented by an ordered categorical sequence of length
T= 72, with an alphabet Vof size v= 6 possible states, e.g. si= (si,1, si,2,...,si,72)=
(SC,SC, ...,TR,TR, ...,EM,EM). The sequences share a common length, the time periods
are equally spaced, and there are no missing data. In the context of the Northern Irish
education system at the time, SC refers to secondary school, which may be a grammar school
to which entrance is granted upon completion of an exam. At age 16, students take General
Certificate of Secondary Education (GCSE) examinations; students who do well are eligible
to continue in school for a further two years (to sit A-level exams) or to leave, e.g. to a
training/apprenticeship programme (TR). Further education (FE) is distinguished from
higher education (HE); FE typically refers to applied post-GCSE courses while HE refers
to third-level/university courses, typically pursued at age 18 after the successful completion
of A-level exams. Notably, the transitions HE SC and TR HE are never observed.
It is of interest to relate the MVAD sequences to covariates in order to understand
whether different characteristics (related to gender, community, geographic and social con-
ditions, and personal abilities) impact on the school-to-work trajectories. These covariates
are summarised in Table 1. All covariates were measured at the age of 16 (i.e. at the start
of the study period in July 1993), with the exception of ‘Funemp’ and ‘Livboth’, and are
thus static background characteristics. As achieving 5 or more grades at A–C in GCSE ex-
ams is the traditional cut-off point for progression to the additional two-years of secondary
school required for a transition to HE, we expect the ‘GCSE5eq’ covariate in particular
to be strongly associated with the clustering. The MVAD data also come with associated
observation-specific survey sampling weights, which depend on the ‘Grammar’ and ‘Loca-
tion’ covariates. Specifically, the sample was stratified in such a way that a predetermined
number of subjects were in each state, for each location and both school types, immediately
after the end of the compulsory education period (Armstrong et al.,1997).
Table 1: Available covariates for the MVAD data set. For binary covariates, the event denoted by 1is
indicated. Otherwise, the levels of the categorical covariate ‘Location’ are grouped in curly brackets.
Covariate Description
Catholic 1=yes
FMPR SOC code of father’s current or most recent job as of the beginning of the survey,
1=SOC1 (Standard Occupational Classification: professional, managerial, or related)
Funemp Father’s employment status as of June 1999, 1=employed
GCSE5eq Qualifications gained by the end of compulsory education, 1=5+ GCSE grades at A–C, or equivalent
Gender 1=male
Grammar Type of secondary education, 1=grammar school
Livboth Living arrangements as of June 1995, 1=living with both parents
Location Location of school, one of five Education and Library Board areas in Northern Ireland,
{Belfast, North Eastern, South Eastern, Southern, Western}
The MVAD data are available in the Rpackages MEDseq and TraMineR (Gabadinho et al.,
2011). As the data have been used to illustrate some of the functionalities of the TraMineR
package in its associated vignette6, interesting features of an exploratory analysis of the data
can be found therein. However, we reproduce plots of the transversal state distributions in
Figure 1and the transversal entropies in Figure 2, i.e. the Shannon entropies of the state
distributions at each time point (Billari,2001), with the sampling weights accounted for in
both cases. Notably, fewer than vstates are observed in certain months.
Figure 1shows that the number of subjects who found employment increased over time.
Conversely, fewer students were in training or further education by the end of the obser-
vation period. Most students appear to have entirely left school within 2/3 years of the
commencement of the survey. Finally, while students only reached the age of 18 and began
to pursue higher education from July 1995 onwards, a number of students had already pur-
sued further education during the two preceding years. Figure 2confirms that the level of
heterogeneity in the state distribution varies over time. In particular, the entropy declines
after Sep 1995, by which point most students had left school.
Weighted Proportions
Jul.93 Jul.94 Jul.95 Jul.96 Jul.97 Jul.98
0.0 0.2 0.4 0.6 0.8 1.0
Further Education
Higher Education
Figure 1: Overall state distribution for the weighted
MVAD data, coloured by state.
Weighted Entropy Index
Jul.93 Jul.94 Jul.95 Jul.96 Jul.97 Jul.98
0.0 0.2 0.4 0.6 0.8 1.0
Figure 2: Transversal entropy plot for the weighted
MVAD data.
Interestingly, many students were jobless during the first two months of observation. As
the vast majority of cases notably remained in the same state in this period, which coincided
with the summer break from school, all subsequent analyses are conducted on a version of
the data with the first two time points removed. Hence, we work hereafter with sequences of
length T= 70, commencing with the return to school in September 1993. As the sampling
design depends on ‘Grammar’ and ‘Location’, the term ‘all covariates’ henceforth refers
to all other covariates in Table 1. While Murphy and Murphy (2020) show that the same
covariate can affect more than one part of a mixture of experts model, and in different ways,
removing the quantities used to define the weights eases the interpretability of the results.
3 Modelling
In this section, we introduce the novel family of MEDseq models. The exponential-distance
model is described in Section 3.1, extended to account for the sampling weights in Section
3.2, expanded into a family of mixtures in Section 3.3, and finally embedded within the mix-
ture of experts framework in Section 3.4 in order to accommodate the available covariates.
3.1 Exponential-Distance Models
For an arbitrary distance metric d(·,·), location parameter θ, and precision parameter λ, the
probability mass function (PMF) of an exponential-distance model (EDM) for sequences is
f(si|θ, λ, d) = exp(λd(si,θ))
vexp(λd(σ,θ)) = Ψ(λ, θ|T, v)1exp(λd(si,θ)) ,(1)
with the corresponding log-likelihood function given by
(θ, λ |S,d) =
log f(si|θ, λ, d) = λ
d(si,θ)nlog Ψ(λ, θ|T, v).(2)
Such a model is analogous to the Gaussian distribution (characterised by the squared
Euclidean distance from the mean) and similar to the Mallows model for permutations (Mal-
lows,1957). Indeed, mixtures of Mallows models have been used to cluster rankings (Murphy
and Martin,2003). We only consider models with λ0. When λ= 0, the distribution
of sequences is uniform. For λ > 0, the central sequence θ= (θ1,...,θT)is the mode, i.e.
the sequence with highest probability, and the probability of any other sequence decays
exponentially as its distance from θincreases. The precision parameter λcontrols the speed
of this decay. Larger λvalues cause sequences to concentrate around θ, tending toward a
point-mass as λ→ ∞. Notably, λis not identifiable when all sequences are identical.
The log-likelihood in (2) is generally intractable, as the normalising constant Ψ(λ, θ|T , v)
depends on the parameter λ(under OM and other, more complicated distances, it can
also depend on θ), as well as the fixed constants T > 1and v > 1, and requires a sum
over all possible sequences. With reference to the MVAD data, for example, computing
Ψ(λ, θ|T, v)is practically infeasible as there are cardST
v=vT= 670 possible sequences.
Fortunately, however, the normalising constant exists in closed form under the Hamming
distance, dH(si,sj) = PT
t=1 1(si,t 6=sj,t), in a manner which facilitates direct enumeration
and crucially does not depend on θ, as a sum with only T+ 1 terms. Consider, for example,
the Hamming distances between all ternary (v= 3) sequences of length T= 4. From the
arbitrary reference sequence (0,0,0,0), there is 1count of a distance of 0,8counts of a
distance of 1,24 counts of a distance of 2,32 counts of a distance of 3, and 16 counts of a
distance of 4. Thus, ΨH(λ|T= 4, v = 3) = e0+ 8eλ+ 24e2λ+ 32e3λ+ 16e4λ. Hence,
the normalising constant under the Hamming distance metric depends on the parameter λ,
the sequence length T, and the number of categories v, and simplifies greatly:
ΨH(λ|T, v) =
p=0 T
p(v1)pexp(λp) = (v1) eλ+ 1T.(3)
Inspired by the generalised Mallows model (Irurozki et al.,2019), the EDM in (1) based
on the Hamming distance can be extended to one based on the weighted Hamming distance.
By introducing Tprecision parameters λ1,...,λT, one for each time point (i.e. sequence
position), and expressing the exponent in (1) as dWH(si,θ|λ1,...,λT) = PT
t=1 λt1(si,t 6=θt)
rather than λdH(si,θ) = λPT
t=1 1(si,t 6=θt), different time periods can contribute differently
to the overall distance, weighted according to the period-specific precision parameters. Thus,
the distance from θto siunder dWH(·,·|·)becomes a sum of the λtvalues associated with
each time point which differs from the corresponding θt, across the whole sequence. This acts
as implicit variable selection and allows modelling situations in which there is high consensus
regarding the state values of some time periods, with large uncertainty about the values of
others. Accounting for the alignment of contemporaneous matchings in this way helps to pre-
vent sequences with the same (Hamming) distance from θfrom having the same probability.
Given that sequences equidistant from θcan nevertheless exhibit element-wise mismatches
between themselves, this may help later, in the mixture setting, to induce stronger between-
cluster separation and within-cluster homogeneity. The non-constant transversal entropies
in Figure 2suggest that this extension may also be fruitful in terms of capturing different
degrees of dispersion in the state distributions of the MVAD data over time. Crucially, the
various benefits outlined above can be achieved without any tractability sacrifices. The
log-likelihood in (2) is merely rewritten with the weighted Hamming distance decomposed
into its Tcomponents and the normalising constant in (3) also modified accordingly:
(θ, λ1,...,λT|S,dWH) =
i=1 "T
t=1 λt1(si,t 6=θt) + log(v1) eλt+ 1#.
Though other dissimilarity measures are available for sequences, we henceforth consider
measures based on the Hamming distance only, chiefly for the computational reasons outlined
above. In our setting, λdH(·,·)can be seen as a special case of OM with all substitution costs
set to λand no insertions or deletions. As it has time-varying substitution costs, dWH (·,·|·)
is similar to the dynamic Hamming distance (Lesnard,2010), a prominent alternative to
OM. However, such costs in our models are always assumed to be common with respect
to each pair of states. Hence, dWH (·,·|·)equates to the Gower distance between nominal
variables (Gower,1971) with equally weighted states and unequally weighted time points.
3.2 Incorporating Sampling Weights
Sampling weights are often associated with life-course data, as the data typically arise from
surveys where the weights are used to correct for representivity bias under stratified sampling
designs. Following Chambers and Skinner (2003), the sampling weights w= (w1,...,wn)
are incorporated into the EDM by exponentiating the likelihood of each sampled unit by the
attached weight wi, which is akin to unit ibeing observed witimes. The resultant pseudo
likelihood Lw(·|·)reweights the likelihood contribution for each unit in order to rebalance
the information in the observed sample to approximate the balance of information in the
target finite population. The sampling weights ware thus interpretable as being inversely
proportional to the unit inclusion probabilities, remain fixed, and are confined to those
included in the sample. Notably, f(si|θ, λ, d)wif(si|θ, wiλ, d), such that the weights
induce a unit-specific rescaling of the precision parameter; it follows that the observed data
are independent but not identically distributed.
A secondary benefit of incorporating weights is that it facilitates computational gains
in the presence of duplicate cases. Such duplicates are likely when dealing with discrete
life-course data. This non-uniqueness can be exploited using likelihood weights for compu-
tational efficiency, by fitting models to the subset of unique sequences only, weighted by the
sum of the sampling weights (if available, otherwise wi= 1 i) across each corresponding set
of duplicates. In modifying win this way, cases with different sampling weights which are
otherwise duplicates are also treated as duplicates, in such a way that the (pseudo) likelihood
is unaltered. The number of duplicates clearly lowers when considering both the sequences
themselves and their associated covariate patterns. In particular, all cases are unique when
there are continuous covariates. Nonetheless, in the MVAD data, and in many applications,
the covariates are all categorical. Hence, exploiting non-uniqueness in this manner can be
extremely computationally convenient. For instance, only 490 of the n= 712 sequences in
the MVAD data set are distinct. However, to avoid notational confusion, all subsequent
expressions are written as though duplicate cases have not been discarded.
Though the weights for the MVAD data sum to 711.52, we henceforth follow Xu et al.
(2013) in always assuming that the weights have been normalised to sum to the sample size
n. In doing so, subsequent expressions are simplified further and the use of model selection
criteria (see Section 4.3) relying on the pseudo likelihood is facilitated. While the resultant
rescaling of the MVAD weights is negligible, we note that multiplying wby a scalar does
not affect parameter estimation.
3.3 A Family of Mixtures of Exponential-Distance Models
Extending the EDM based on the Hamming distance with sampling weights to the model-
based clustering setting yields a pseudo likelihood function of the form
Lw(θ1,...,θG, λ |S,w,dH) =
i=1 "G
((v1) eλ+ 1)T#wi
where the mixing proportions τ1,...,τGare positive and sum to 1. Thus, the clustering
approach is both model-based and distance-based, thereby bridging the gap between these
two ‘cultures’ in the SA community.
The mixture setting naturally suggests a further extension, whereby the precision param-
eter λcan be constrained or unconstrained across clusters, in addition to the aforementioned
allowance for the precision parameters to vary (or not) across time points. Within a family of
models we term ‘MEDseq’, we thus define the CC,UC,CU, and UU models, where the first
letter denotes whether precision parameters are constrained (C) or unconstrained (U) across
clusters and the second denotes the same across time points. Notably, all models deviate
from the simple matching distance on which they are based, as even the most constrained CC
model could be said to employ a weighted variant thereof, by virtue of allowing for λ6= 1.
The model family allows moving between more parsimonious models and more heavily pa-
rameterised, flexible models which may provide a better fit to the data. As the precision
parameters relate to the substitution costs characterising variants of the Hamming distance,
quantities used to define the overall distance measure are allowed to vary in different ways,
while still being treated as model parameters rather than inputs. In particular, models with
names beginning with Ureflect scenarios in which the implicit substitution costs differ across
clusters. Hence, the UU model is analogous to the hierarchical Wardpalgorithm (de Amorim,
2015), in the sense of having cluster-specific feature weights (albeit with no tuning required).
Given the role played by λwhen it takes the value 0, whereby the state distribution is uni-
form, it is convenient and natural to include a noise component (denoted by N), whose single
precision parameter is fixed to 0, to robustify inference by capturing deviant cases and min-
imising their deleterious effects on parameter estimation for the other, more defined clusters.
Adding this extension to each of the 4models above, regardless of how precision parameters
are otherwise specified, completes the MEDseq model family with the CCN,UCN,CUN, and
UUN models. When G= 1, the CC,CU, and CCN models can be fitted. When G= 2, the
UCN and UUN models are equivalent to the CCN and CUN models, respectively, as there
is only one non-noise component. As the noise component arises naturally from restricting
the parameter space, we consider the noise component as one of the Gcomponents, denoted
hereafter with the subscript 0. All 8model types are summarised further in Appendix A.
3.4 Incorporating Covariates
We now illustrate how to incorporate the available covariate information into the clustering
process, both to guide the construction of the clusters and to find the typical trajectories
which can be best predicted by covariates. As is typical for model-based clustering analyses,
the data are augmented in MEDseq models by introducing a latent cluster membership
indicator vector zi= (zi,1, . . . , zi,G), where zi,g = 1 if observation ibelongs to cluster gand
zi,g = 0 otherwise. The MEDseq approach can be easily extended to incorporate the possible
effects of covariates on the assignments of sequences to clusters by allowing the covariates
to influence the distribution of the latent variable zi. Thus, such covariates are interpreted
differently from those used to define the sampling weights, if any.
The inclusion of covariates is achieved under the mixture of experts framework (Jacobs
et al.,1991;Gormley and Frühwirth-Schnatter,2019), by extending the mixture model to
allow the mixing proportions for observation ito depend on covariates xi. This, rather than
having covariates enter through the component distributions, is particularly attractive, as
the interpretation of the remaining component-specific parameters is the same as it would
be under a model without covariates. For example, in the case of the CC MEDseq model
f(si|θ1,...,θG, λ, β1,...,βG,xi,dH)wi="G
((v1) eλ+ 1)T#wi
where the mixing proportions τgxi|βg(henceforth τg(xi), for simplicity) are referred to as
the ‘gating network’, with τg(xi)>0and PG
g=1 τg(xi) = 1, as usual, and β1,...,βGare the
gating network regression parameters. Such a model can be seen as a conditional mixture
model (Bishop,2006) because, given the covariates xi, the distribution of the sequences is
a finite mixture model under which zihas a multinomial distribution with a single trial
and probabilities equal to τg(xi). The distance-based k-medoids algorithm, though closely
related (see Section 4.2), does not accommodate the inclusion of gating covariates in this way.
Incorporating covariates in ‘hard’ clustering algorithms using MLR, as per McVicar
and Anyadike-Danes (2002), has been criticised because the hard assignment of extraneous
cases can negatively impact internal cluster cohesion and the MLR coefficient estimates
(Piccarreta and Studer,2019). An advantage of the noise component in MEDseq models
is that it captures uniformly distributed sequences that deviate from those in the other,
more defined clusters. Filtering outliers in this way lessens their impact on the non-noise
gating network coefficients, thereby enabling more accurate inference and improving the
interpretability of the effects of the covariates. Moreover, the ‘soft’ partition obtained under
the model-based paradigm allows the cluster membership probabilities for sequences lying
on the boundary between two neighbouring clusters to be quantified and the effect of such
sequences on the gating network coefficients to be mitigated.
As per Murphy and Murphy (2020), the CCN,UCN,CUN, and UUN models which include
an explicit noise component can be restricted to having covariates only influence the mixing
proportions for the non-noise components, with all observations therefore assumed to have
equal probability of belonging to the uniform noise component (i.e. by replacing τ0(xi)
with τ0). We refer to the former setting as the gated noise (GN) setting and to the latter
as the non-gated noise (NGN) setting. The NGN setting is the more parsimonious one,
makes more clear the distinction between EDM components and the uniform component,
and is particularly apt when τ0is expected to be small and/or the sequences are expected to
overwhelm the gating covariate(s) in determining which cases are noise. Gating covariates
can only be included when G2under the GN setting or when there are 2or more
non-noise components under the NGN setting.
4 Model Estimation
This section describes our model-fitting approach and some implementation issues that arise
in practice. Specifically, Section 4.1 outlines the ECM algorithm employed for parameter
estimation, Section 4.2 discusses the initialisation thereof with reference to the similarities
between MEDseq models and the k-medoids and k-modes (Huang,1997) algorithms, and the
issues of model selection, covariate selection, and model validation are treated in Section 4.3.
4.1 Model Fitting via ECM
Parameter estimation is greatly simplified by the existence of a closed-form expression for
the normalising constant for MEDseq models based on the Hamming or weighted Hamming
distances. We focus on maximum (pseudo) likelihood estimation using a simple variant
of the EM algorithm (Dempster et al.,1977). For simplicity, model fitting details are
described chiefly for the CC MEDseq model with sampling weights and gating covariates.
Additional details for other model types are deferred to Appendix B; so, too, are technical
details pertaining to estimation of the precision parameter(s). The complete data pseudo
likelihood for the CC model is given by
c(θ1,...,θG, λ, β1,...,βG|S,X,Z,w,dH) =
i=1 "G
g=1 τgxiexp(λdH(si,θg))
((v1) eλ+ 1)T!zi,g #wi
and the complete data pseudo log-likelihood hence has the form:
c(θ1,...,θG, λ, β1,...,βG|S,X,Z,w,dH) =
zi,gwi[log τgxiλdH(si,θg)
Tlog(v1) eλ+ 1.
Under this model, the distribution of sidepends on the latent cluster membership variable
zi, which in turn depends on covariates xi, while siis independent of xiconditional on zi.
The iterative algorithm for MEDseq models follows in a similar manner to that for
standard mixture models. It consists of an E-step (expectation) which replaces for each
observation the missing data ziwith their expected values b
zi, which sum to 1, followed by a
M-step (maximisation), which maximises the expected complete data pseudo log-likelihood.
The M-step consists of a series of conditional maximisation (CM) steps in which each pa-
rameter is maximised individually, conditional on the other parameters remaining fixed.
Hence, model fitting is in fact conducted using an expectation conditional maximisation
(ECM) algorithm (Meng and Rubin,1993). Aitken’s acceleration criterion is used to as-
sess convergence of the non-decreasing sequence of weighted pseudo log-likelihood estimates
(Böhning et al.,1994). Parameter estimates produced on convergence achieve at least a
local maximum of the pseudo likelihood function. Upon convergence, cluster memberships
are estimated via the maximum a posteriori (MAP) classification, i.e. cases are assigned to
the cluster gto which they most probably belong via MAP(b
zi) = arg maxg∈ {1,...,G}bzi,g.
The E-step (with similar expressions when λis unconstrained across clusters and/or time
points) involves computing expression (5), where (m+ 1) is the current iteration number:
i,g =Ezi,g si,xi,b
g, wi,dH
Note that the weights wiappear in neither the numerator nor the denominator, leaving the
E-step unchanged regardless of the inclusion or exclusion of weights.
Subsequent subsections describe the CM-steps for estimating the remaining parame-
ters in the model. These individual CM-steps rely on the current estimates b
Z(m+1) =
nto provide estimates of the gating network regression coefficients b
and hence the mixing proportion parameters bτ(m+1)
g(xi), as well as the central sequence(s)
gand component precision parameter(s) b
λ(m+1), though technical details for the latter,
as they are the element which distinguishes the various MEDseq model types, are deferred
to Appendix B. It is clear from (4) that the sampling weights can be accounted for by
simply multiplying every b
iby the corresponding weight wi. Conversely, in the CM-
steps which follow, corresponding formulas for unweighted MEDseq models can be recovered
by replacing bz(m+1)
i,g wiwith bz(m+1)
i,g .
4.1.1 Estimating the Gating Network Coefficients
The portion of (4) corresponding to the gating network, given by Pn
i=1 PG
g=1 zi,gwilog τg(xi),
is of the same form as a MLR model with weights given by wi, here written with component
1as the baseline reference level for identifiability reasons:
log τg(xi)
τ1(xi)= log Pr(zi,g = 1)
Pr(zi,1= 1) =e
xiβgg2,with β1= (0,...,0),
where e
xi= (1,xi). Thus, methods for fitting such models, with b
Z(m+1) as the response,
can be used to estimate the gating network regression parameters b
g. As closed-form
updates are unavailable for MLR coefficients, due to the nonlinear numerical optimisation
involved, this step merely increases (rather than maximises) the expectation of this term.
However, the monotonicity of the sequence of pseudo log-likelihood estimates is preserved
and convergence is still guaranteed. Subsequently, the mixing proportions are given by
g(xi) = expe
h=1 expe
Conversely, τis estimated exactly via bτ(m+1)
i=1 bz(m+1)
i,g wiwhen there are no gating
covariates. Since Pn
i=1 wi=n, this is simply the weighted mean of the g-th column of the
matrix b
Z(m+1). However, τcan also be constrained to be equal (i.e. τg=1
/Gg) across
clusters. Thus, situations where τi,g =τg(xi),τi,g =τg, or τi,g =1
/Gare accommodated.
The standard errors of the gating network’s MLR at convergence are not a valid means of
assessing the uncertainty of the coefficient estimates as the cluster membership probabilities
are estimated rather than fixed and known. Therefore, we adapt the weighted likelihood
bootstrap (WLBS) of O’Hagan et al. (2019) to the MEDseq setting. This is implemented
by multiplying the sampling weights wby draws from an n-dimensional symmetric uniform
Dirichlet distribution and refitting the MEDseq model. To ensure stable estimation of the
standard errors, B= 1000 such samples are used here. To ensure rapid convergence and
to circumvent label-switching problems, the estimated b
Zmatrix from the original model is
used to initialise the ECM algorithm for each sample with new likelihood weights. Finally,
the standard errors of the gating network coefficients across the Bsamples are obtained.
Although this approach does not produce fully valid variance estimates when there are
sampling weights which arise from stratified designs, we adopt the WLBS in what follows in
order to provide approximate standard errors. This issue is particularly pronounced when
the probability of being included in the sample depends on quantities being modelled. This
concern provides additional justification for the aforementioned removal of the Grammar
and Location covariates from our analysis.
4.1.2 Estimating the Central Sequences
The location parameter θis sometimes referred to as the Fréchet mean or the central se-
quence. The k-medoids/PAM algorithm, which is closely related to the MEDseq models with
certain restrictions imposed (see Section 4.2), fixes the estimate of b
θgto be the medoid of
cluster g(Kaufman and Rousseeuw,1990), i.e. the observed sequence siSwith minimum
weighted distance from the others currently assigned to the same cluster. This estimation
approach is especially quick as the Hamming distance matrix for the observed sequences is
pre-computed. Notably, this greedy search strategy may fail to find the optimum solution.
However, for a G= 1 unweighted EDM based on the Hamming distance, the maximum
likelihood estimate (MLE) of θis given simply by the modal sequence, meaning that each
θtis independently given by the most frequent state at the t-th time point. This is intuitive
when dH(si,θ)is expressed as TPT
t=1 1(si,t =θt), as b
θmaximises the number of element-
wise agreements. Thus, the parameter has a natural interpretation. For more complicated
distance metrics, the first-improvement algorithm (Hoos and Stützle,2004) or a genetic
algorithm could be used to estimate θ. Notably, the modal sequence need not be an observed
sequence in S. It is also notable that any b
θtmay be non-unique under any of the proposed
estimation strategies. Such ties, if any, are broken at random.
For G > 1, under the ECM framework, central sequence position estimates b
g,t are
given by arg maxϑVt
i=1 bz(m+1)
i,g wi1(si,t =ϑ), where Vtis the subset of vtvstates
observed at time point tacross all cases. As this expression is independent of the precision
parameter(s), it holds for all MEDseq model types, including those based on weighted Ham-
ming distance variants. Thus, b
gis similarly estimated easily and exactly via a weighted
mode (much like k-modes), whereby each b
g,t is given by the category corresponding to
the maximum of the sum of the weights bz(m+1)
i,g wiassociated with each of the vtobserved
state values. Similarly, the central sequence under a weighted G= 1 model is also estimated
via a weighted mode, with the weights given only by wi. Notably, to estimate the central
sequences for a MEDseq model of any type without sampling weights, one need only remove
wifrom these terms. Note also that θ0does not need to be estimated for models with an
explicit noise component as it does not contribute to the likelihood.
4.2 ECM Initialisation and Comparison to k-medoids
MEDseq models share relevant features with the PAM algorithm. Both consider sequences
from a holistic perspective and both rely on distances to a cluster centroid. However,
PAM treats the matrix of pairwise distances between sequences as a pre-computed input,
while under MEDseq models the distances to the centroids (and the costs which define the
distance metric) are recomputed at each iteration, with the sequences themselves as input.
Otherwise, compared to PAM based on the Hamming distance, MEDseq models differ only
in that i) θgis estimated by the modal sequence rather than the medoid, ii) τis estimated,
or dependent on covariates via τg(xi), rather than constrained to be equal, iii) λis free to
vary across clusters and/or time points, rather than being implicitly set to 1, iv) a noise
component can be included, and v) the ECM algorithm rather than the classification EM
algorithm (CEM; Celeux and Govaert,1992) is used. The CEM algorithm employed by PAM
uses hard assignments ezi,g, computed in its C-step, such that ez(m+1)
i,g = 1 if g=MAPb
and ez(m+1)
i,g = 0 otherwise, for which the denominator in (5) need not be evaluated.
Thus, a CC model fitted by CEM, with λ= 1, equal mixing proportions, and the
central sequences estimated by the medoid rather than the modal sequence, is equivalent
to k-medoids based on the Hamming distance. We leverage these similarities by applying
k-medoids to the Hamming distance matrix in order to initialise the ECM algorithm with
‘hard’ starting values for the allocation matrix Z. In particular, we rely on a weighted
version of PAM available in the Rpackage WeightedCluster (Studer,2013), itself initialised
using Ward’s hierarchical clustering. The more closely related k-modes algorithm (Huang,
1997) is not used, as case-weighted implementations are currently unavailable. In any case,
our strategy is less computationally onerous than using multiple random starts. Moreover,
our experience suggests that the ECM algorithm converges quickly when our initialisation
strategy is adopted and that a great many number of random starts are required in order to
achieve comparable performance. For models with an explicit noise component, an initial
guess of the prior probability τ0that observations are noise is required. Allocations are then
initialised, assuming the last component is the one associated with λg= 0, by multiplying
the initial (G1)-column Zmatrix by 1τ0and appending a column in which each entry
is τ0. We caution that the initial τ0should not be too large.
4.3 Model Selection and Validation
In contrast to heuristic clustering approaches like k-medoids and Ward’s hierarchical method,
the model-based paradigm facilitates principled model-selection using likelihood-based in-
formation criteria. In the MEDseq setting, the notion of model selection refers to identifying
the optimal number of components Gin the mixture and finding the best MEDseq model
type in terms of constraints on the precision parameters. Variable selection on the subset
of covariates included in the gating network can also improve the fit. For a given set of
covariates, one would typically evaluate all model types over a range of Gvalues and choose
simultaneously both the model type and Gvalue according to some criterion. Thereafter,
different fits with different covariates can be compared according to the same criterion.
The Bayesian Information Criterion (BIC; Schwarz 1978) includes a penalty term which
depends on the number of free parameters kin the model. The parameter counts can be
deceptive for MEDseq models. In particular, regarding the estimation of b
θg,t, we note that
identifying the modal state for a given time point implicitly involves estimating occurrence
probabilities for (vt1) states and then selecting the most common. This is accounted for
in Appendix A, wherein the number of free parameters in under each MEDseq model type
is summarised. We also note that the penalty klog nis applied to the maximum pseudo
log-likelihood estimate in the sample-weighted setting (Xu et al.,2013).
Beyond its use in identifying the optimal Gand precision parameter settings, the BIC is
also employed in greedy stepwise selection algorithms in order to guide the inclusion/exclusion
of relevant gating covariates. We propose a bi-directional search strategy in which each step
can potentially consist of adding or removing a non-noise component or adding or removing
a covariate. Interaction terms are not considered. Every potential action is evaluated over
all possible model types at each step, rather than considering changing the model type as an
action in itself. Changing the gating covariates or changing the number of components can
affect the model type, as observed by Murphy and Murphy (2020). While this makes the
stepwise search more computationally intensive, it is less likely to miss optimal models as it
explores the model space. For steps involving both gating covariates and a noise component,
models with both the GN and NGN settings can be evaluated and potentially selected.
A backward stepwise search starts from the model, with all covariates included, consid-
ered optimal in terms of the number of components Gand the MEDseq model type. On the
other hand, a forward stepwise search uses the optimal model with no covariates included
as its starting point. In both cases, the algorithm accepts the action yielding the highest
increase in the BIC at each step. The computational benefits of upweighting unique cases
and discarding redundant cases are stronger for the forward search, as early steps with fewer
covariates are likely to have fewer unique cases across sequence patterns and covariates.
As a means of validating the model chosen by BIC, we turn to silhouette analysis to assess
the quality of the clustering in terms of internal cohesion, where high cohesion indicates high
between-cluster distances and strong within-cluster homogeneity. Typically, the silhouette
width is defined for clustering methods which produce a ‘hard’ partition (Rousseeuw,1987),
and the average silhouette width (ASW) or weighted average silhouette width (wASW;
Studer 2013) is used as a model selection criterion. However, Menardi (2011) introduces the
density-based silhouette (DBS) for model-based clustering methods. This allows the ‘soft’
assignment information to be used, which is discarded when using the MAP assignments in
the computation of the wASW. The empirical DBS for observation iis given by
maxh∈ {1,...,n}logbz0
As observations are assigned to clusters via the MAP classification, c
dbsiis proportional to
the log-ratio of the posterior probability associated with the MAP assignment of observation
i(denoted by bz0
i) to the maximum posterior probability that the observation belongs to
another cluster (denoted by bz1
i). Use of the MAP classification implies 0c
with high values indicating a well-clustered data point. Ultimately, the mean or the median
dbs value can be used as a global quality measure, albeit with two modifications. Firstly, we
identify a set of crisply assigned observations having bz1
ilower than a tolerance parameter ǫ,
here set equal to 10100. These observations are given c
dbsivalues of 1and are excluded from
the computation of the maximum in the denominator of (6) for reasons of numerical stability.
Secondly, we account for the sampling weights by computing a weighted mean density-based
silhouette criterion (wDBS). While neither the wDBS nor wASW are defined for G= 1,
unlike the BIC, they are not employed here as model selection criteria. These silhouette
summary measures are used only to validate MEDseq clustering solutions and to facilitate
comparisons with other methods in Section 5.2. Higher values are preferred for both criteria.
5 Analysing the MVAD Data
Results of fitting MEDseq models to the weighted MVAD data are provided in Section
5.1. All results were obtained via our purpose-built Rpackage MEDseq (Murphy et al.,
2021). The impact of discarding the sampling weights is also studied. A comparison against
other approaches, including hierarchical, partitional, and model-based clustering methods,
is included in Section 5.2. A discussion of the insights gleaned from the solution obtained
by the optimal MEDseq model is deferred to Section 6.
5.1 Application of MEDseq
Weighted MEDseq models are fit for G= 1,...,25, across all 8model types (where al-
lowable), firstly with all covariates included in the gating network (again, where allowable).
The noise components, where applicable, are treated using the NGN setting. Figure 3shows
the behaviour of the BIC for these models. To better highlight the differences in BIC, lower
values for G < 5are not shown. Under these conditions, a G= 11 UUN model is identified
as optimal. The same model type and number of components are identified as optimal
when the noise components are treated with the GN setting, and when the same analysis
is repeated with no covariates at all.
−110000 −105000 −100000 −95000
Number of Components (G)
5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24 25
Figure 3: BIC values for all MEDseq model types, with weights and all covariates, for a range of Gvalues.
In refining the model further via greedy stepwise selection, both the forward search (see
Table 2) and backward search (see Table 3) thus begin with the same number of components
and the same model type. As previously stated, covariates used to define the sampling
weights are excluded in both cases. Notably, no step in either search elects to modify G
or the model type. Both searches converge to the same G= 11 UUN model with only the
single covariate ‘GCSE5eq’ in the NGN gating network, though the search in the forward
direction does so in fewer steps. Under this model, the probability of belonging to the noise
component is constant and does not depend on the included covariate.
Table 2: Summary of the steps taken to improve the BIC in the forward direction.
Optimal Step GModel Type Gating Covariates Gating Type BIC
11 UUN 93190.08
Add ‘GCSE5eq’ 11 UUN GCSE5eq NGN 92953.85
Stop 11 UUN GCSE5eq NGN 92953.85
Table 3: Summary of the steps taken to improve the BIC in the backward direction.
Optimal Step GModel Type Gating Covariates Gating Type BIC
11 UUN Catholic, FMPR, Funemp, GCSE5eq, Gender, Livboth NGN 93111.30
Remove ‘FMPR’ 11 UUN Catholic, Funemp, GCSE5eq, Gender, Livboth NGN 93068.09
Remove ‘Livboth’ 11 UUN Catholic, Funemp, GCSE5eq, Gender NGN 93025.73
Remove ‘Catholic’ 11 UUN Funemp, GCSE5eq, Gender NGN 92994.32
Remove ‘Funemp’ 11 UUN GCSE5eq, Gender NGN 92967.23
Remove ‘Gender’ 11 UUN GCSE5eq NGN 92953.85
Stop 11 UUN GCSE5eq NGN 92953.85
Notably, there is little difference between the respective clusterings produced by the
various models including no covariates, all covariates, and only GCSE5eq. Indeed, both the
soft b
Zmatrices and hard MAP assignments are almost identical between each pair of models;
relative to the optimal model after stepwise selection, there are only 1and 2cases assigned to
different clusters under equivalent models with no covariates and all covariates, respectively.
Thus, the sequences themselves overwhelm the covariates and there is little confounding
between the simultaneous roles of GCSE5eq under the optimal model in guiding both the
construction of the clusters and their interpretation. Moreover, the parsimony afforded by
discarding the other covariates simplifies the interpretation greatly. Thus, while adapting
the ‘two-step’ approach introduced for LCR (Bakk and Kuha,2018) to the MEDseq setting
may be of interest for other applications, the results for the MVAD data do not differ greatly
from those presented in Section 6, as shown in Appendix C.
For completeness, the analysis above is repeated with the sampling weights discarded
entirely and consideration given where appropriate to the two covariates used to define w.
In doing so, identical inference is obtained on the model type; however, the results differ in
terms of the optimal G(now 10), the uncovered partition, and the estimated model param-
eters. This is not surprising, as failure to account for win the clustering produces biased
estimates of the component-specific parameters and the cluster membership probabilities, as
well as the gating network coefficients. Additionally, an extra gating covariate (Grammar)
is included after stepwise selection in the unweighted analysis. However, the results are
reasonably robust to a coarsening of the sequences; in repeating all analyses with the data
subsetted into six-monthly intervals, similar inferences are again obtained. Notably, the
ECM algorithm’s runtime is not greatly reduced in doing so. Indeed, MEDseq models scale
more poorly with n(or, more specifically, the number of unique cases) rather than Tor v, as
the number of (pseudo) likelihood evaluations required for large nis more computationally
expensive than the number of simple matching evaluations required for long sequences.
5.2 Other Clustering Methods
To contrast the MEDseq results for the MVAD data with those obtained by other methods,
we present a non-exhaustive comparison against some distance-based and some Markovian
approaches. Regarding the former, we present only some common heuristic methods which
treat the distance matrix as the input using distance metrics which are commonly adopted
in the literature on life-course sequences, namely PAM and Ward’s method based on the
Hamming distance and OM. We note that fuzzy clustering offers an alternative distance-
based perspective which also allows for soft assignments (see D’Urso (2016) for an excellent
overview), with further, separate extensions for incorporating covariates and including a
noise cluster in Studer (2018) and D’Urso and Massari (2013), respectively. However, this
paradigm is not considered further, both for the sake of brevity and because case-weighted
implementations are currently unavailable. LCA and LCR, fit via the Rpackage poLCA
(Linzer and Lewis,2011), are also excluded, as they encounter computational difficulties due
to the explosion in the number of parameters for G3. Among the considered methods,
only MEDseq and the distance-based methods can accommodate the sampling weights.
Firstly, MEDseq models with no covariates and all covariates are compared against
weighted versions of k-medoids, using the Rpackage WeightedCluster (Studer,2013), and
Ward’s hierarchical clustering. Here, k-medoids is itself initialised using Ward’s method.
Neither method can be compared to MEDseq models in terms of BIC or wDBS values, as they
are not model-based and do not yield ‘soft’ cluster membership probabilities, respectively.
Thus, Figure 4shows a comparison of wASW values using MAP classifications where neces-
sary. Only the MEDseq model type (and gating network setting, for models with covariates)
with the highest wASW for each Gvalue is shown, for clarity. Note that the wASW is com-
puted using the observed Hamming distance matrix, which both comparators in Figure 4
utilise directly, while MEDseq models are only based on the Hamming metric. Nonetheless,
MEDseq models show superior or competitive performance across the majority of Gvalues.
In particular, the optimal model identified after stepwise selection achieves wASW=0.386.
The superior wASW values achieved by MEDseq models provide evidence that the proposed
methodology, which embeds features of the distance-based approaches into a model-based
setting, yields more compact and well-separated clusters. Notably, similar conclusions are
drawn when OM — with the same cost settings as used in McVicar and Anyadike-Danes
(2002) — is used in place of the Hamming distance for k-medoids and Ward’s method.
0.30 0.35 0.40 0.45
Number of Components (G)
Weighted ASW
2 4 6 8 10 12 14 16 18 20 22 24 26 28 30 32 34 36 38 40
MEDseq: no covariates
MEDseq: all covariates
Ward's Hierarchical Clustering
Figure 4: Values of the wASW measure, using Hamming distances, for the best MEDseq model type for
each Gvalue with no covariates and all covariates. Corresponding values for weighted versions of k-medoids
and Ward’s hierarchical clustering based on the Hamming distance are also shown.
Secondly, finite mixtures with first-order Markov components, fit via the Rpackage
ClickClust (Melnykov,2016b), are also included in the comparison. This package allows
the initial state probabilities to be either estimated or equal to 1
/vfor all categories; both
scenarios are considered and other function arguments are set to their default values. The
wASW values for the ClickClust models are not shown in Figure 4; they are much lower
than those of the other methods up to G= 5 and turn negative thereafter. Though this
implies inferior clustering behaviour for ClickClust models, the method also returns a
Zmatrix of cluster membership probabilities. Hence, these models are also compared to
MEDseq in terms of the wDBS measure in Figure 5. Again, only the best model of each type
is shown for each Gvalue; here, the MEDseq models again exhibit the best performance
over the entire range. Notably, the optimal G= 11 UUN MEDseq model with ‘GCSE5eq’
in the gating network achieves wDBS=0.455. An advantage of ClickClust is that it allows
sequences of unequal lengths, but this is not a concern for the MVAD data.
0.1 0.2 0.3 0.4 0.5 0.6
Number of Components (G)
Weighted Mean DBS
2 4 6 8 10 12 14 16 18 20 22 24 26 28 30 32 34 36 38 40
MEDseq: no covariates
MEDseq: all covariates
Figure 5: Values of the wDBS measure for the best MEDseq model type at each Gvalue with no covariates
and all covariates. Corresponding values for the best ClickClust model are also shown.
Thirdly, the Rpackage seqHMM (Helske and Helske,2019) provides tools for fitting mix-
tures of hidden Markov models, with gating covariates influencing cluster membership prob-
abilities. Such models allow cluster memberships to evolve over time, similar to mixed
membership models (Airoldi et al.,2014). They thus cannot be directly compared to MED-
seq models. However, we note that the seqHMM package provides a pre-fitted model for
the MVAD data, with the first two months also discarded and no covariates. The model
has 2clusters, with 3and 4hidden states, respectively, and achieves wDBS=0.50 and
wASW=0.23. Otherwise identical seqHMM models, including either all covariates or only
the GCSE5eq covariate chosen for the optimal MEDseq model via stepwise selection, both
achieve wDBS=0.47 and wASW=0.23. Notably, these wDBS and wASW values are much
worse than those for MEDseq models with G= 2. Overall, the ClickClust and seqHMM
results suggest that holistic approaches — MEDseq models, in particular — yield better
clusterings than Markovian ones for the MVAD data.
6 Discussion of the MVAD Results
To better inform a discussion of the results obtained by the optimal G= 11 UUN model for
the MVAD data, with the covariate GCSE5eq in the NGN gating network, its estimated
central sequences are first shown in Figure 6. Seriation has been applied, using the observed
Hamming distance matrix and the travelling salesperson combinatorial optimisation algo-
rithm (Hahsler et al.,2008), in order to give consecutive numbers to clusters with similar
estimated (weighted) modal sequences. Each cluster’s label is derived from the represen-
tation of b
θgin State-Permanence-Sequence format (SPS; Aassve et al.,2007). The same
ordering and labels are used in all subsequent graphical and tabular displays of results. The
uncovered clusters are shown in Figure 7, to which additional seriation has been applied in
order to also group the observations within clusters, for visual clarity. Finally, the average
time spent in each state by cluster — weighted by wiand the estimated cluster membership
probabilities — is shown in Table 4, along with the cluster sizes.
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Higher Education
Figure 6: Central sequences of the optimal G= 11 UUN model with the GCSE5eq gating covariate. The
SPS labels on the right characterise each non-noise cluster by the distinct successive states in b
θg, with
associated durations (in months).
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Figure 7: Clusters uncovered under the optimal G= 11 UUN model with the GCSE5eq gating covariate.
The rows correspond to the n= 712 observed sequences, including duplicate cases previously discarded
during model fitting, grouped according to the MAP classification and ordered according to the observed
Hamming distance matrix. Each cluster is named according to the SPS representation of b
Table 4: Average durations (in months) spent in each state by cluster, weighted by bzi,g wi, for the optimal
11-component UUN model, with the SPS labels derived from b
θg. Estimated cluster sizes bngcorrespond to
the MAP partition.
Cluster: g(SPS) bngEMployment Further Higher JobLessness SChool TRaining
Education Education
1 (SC,25)-(HE,45) 87 3.77 0.29 38.45 0.89 26.07 0.54
2 (FE,25)-(HE,45) 59 4.65 26.51 37.63 0.45 0.76 0.00
3 (SC,24)-(FE,36)-(HE,10) 18 3.40 30.58 8.56 4.07 21.84 1.56
4 (SC,25)-(EM,45) 32 35.60 1.68 3.63 2.85 25.60 0.63
5 (TR,37)-(EM,33) 60 28.29 1.24 0.00 3.38 1.35 35.74
6 (TR,22)-(EM,48) 67 45.84 1.47 0.00 3.15 1.51 18.03
7 (TR,5)-(EM,65) 165 57.50 2.11 0.00 5.16 1.62 3.62
8 (FE,22)-(EM,48) 95 41.30 22.65 0.99 3.04 1.30 0.72
9 (SC,10)-(FE,36)-(EM,24) 56 21.82 35.19 0.27 3.99 6.15 2.58
10 (TR,10)-(JL,2)-(TR,3)-(JL,55) 55 8.40 3.38 0.22 42.89 4.20 10.91
Noise — 18 21.50 11.42 1.19 14.51 2.20 19.18
This solution tends to group individuals who experience trajectories that are similar or
that differ only for relatively short periods. In particular, the dominating combinations of
states experienced over time are clearly identified, and differences in durations and/or age at
transition are quite limited in size. Within clusters, substantial reduction of misalignments
and/or differences in the durations of spells are evident. Ultimately, the partition is char-
acterised not only by the sequencing (i.e. the experienced, ordered combinations of states),
but also by the spell durations and the ages at transitions which appear mostly homoge-
neous within clusters. This can be explained by the fact that cases in the identified groups
tend to dedicate the same period of time (spells of 1, 2, or 3 years) to further/higher educa-
tion and/or training. This is interesting because one might expect the chosen dissimilarity
metric, as it based on the Hamming distance, to attach higher importance to the sequencing.
The 11-cluster solution for the MVAD data separates individuals who continued in school
(clusters 1, 3, and 4), or otherwise prolonged their studies after the end of compulsory
education (clusters 2, 8, and 9), from those who entered the labour market (clusters 5, 6,
and 7). The clear division visible for some clusters in Figure 7around Autumn 1995, when
new semesters of further and higher education commenced and the majority of those still
remaining in school had eventually left, corresponds to the time point in Figure 2after
which the entropies declined. Interestingly, individuals who experienced prolonged periods
of unemployment are mostly isolated in cluster 10; this is particularly important because
the Status Zero Survey aimed to identify such ‘at risk’ subjects. From this we conclude
that youth unemployment in Northern Ireland in this period was predominantly a problem
of small numbers experiencing long spells of non-participation in the labour market rather
than large numbers dipping into brief, frictional spells.
Clusters 1, 3, and 4 include subjects who continued school for about two years, presum-
ably to retake previously failed examinations or to pursue academic or vocational qualifi-
cations. These individuals are split into two groups depending on whether they continued
their studies (FE: cluster 3, or HE: cluster 1) or were employed directly (cluster 4). Clus-
ters 2, 8, and 9 group subjects who initially entered further education, for about two years
(clusters 2 and 8) or more (cluster 9). Most subjects in clusters 8 and 9 entered employment
directly after further education, whereas the vast majority of those in cluster 2 transitioned
to higher education, where they remained until the end of the observation period.
As for the clusters of individuals who moved quickly to the labour market after the
end of compulsory education, it is possible to distinguish between individuals who almost
immediately found a job and remained in employment for most of the observation period
(the large cluster 7) and individuals who entered government-supported training schemes
(clusters 5 and 6). A further separation is between subjects who were employed after about
2 years of training (cluster 6) and those who participated in training for a much longer
period (cluster 5). Importantly, most of the individuals in these two clusters were able to
find a job even if some respondents experienced some periods of unemployment.
It is interesting to observe that the cluster of careers dominated by persistent unem-
ployment (cluster 10) is characterised by different experiences at the end of the compulsory
education period. Indeed, some subjects entered employment directly after the end of com-
pulsory education but left or lost their job after some months, while some prolonged their
education before becoming unemployed. However, the majority entered a training period
that did not evolve into steady employment.
Notably, the optimal model identified is a UUN model, i.e. one whose precision param-
eters vary across both clusters and time points. Thus, model selection favours a flexible,
heavily-parameterised MEDseq variant which, while based on the simple Hamming dis-
tance, has cluster-specific and period-specific costs which allow element-wise mismatches
between sequences and the central sequences in different time periods in different clusters
to contribute differently to the overall distance measure. While a display of the estimated
precision parameters is omitted, for brevity, their values can be easily examined via the
MEDseq Rpackage. Nonetheless, it is already clear that the model captures different degrees
of heterogeneity in the cluster-specific state distributions of each month.
The coefficients of the gating network with associated WLBS standard errors are given
in Table 5, from which a number of interesting effects can be identified. The interpretation
of the effects of the covariates is made clearer by virtue of there being just one included
after stepwise selection. For completeness, gating network coefficients and associated WLBS
standard errors for the model with all covariates included are provided in Appendix C.
Table 5: Multinomial logistic regression coefficients and associated WLBS standard errors (in parentheses),
with SPS labels, for the NGN gating network of the optimal G= 11 UUN model with the GCSE5eq covariate.
Recall that GCSE5eq=1for subjects who achieved 5 or more grades at A–C (or equivalent) in GCSE exams.
Cluster: g(SPS) (Intercept) GCSE5eq
1(SC,25)-(HE,45) — —
2(FE,25)-(HE,45) 0.95 (0.44)0.47 (0.49)
3(SC,24)-(FE,36)-(HE,10) 0.46 (0.63)1.23 (0.73)
4(SC,25)-(EM,45) 0.58 (0.44)2.18 (0.58)
5(TR,37)-(EM,33) 1.03 (0.38)3.43 (0.55)
6(TR,22)-(EM,48) 1.19 (0.35)3.73 (0.50)
7(TR,5)-(EM,65) 1.70 (0.32)4.09 (0.47)
8(FE,22)-(EM,48) 0.60 (0.38)2.20 (0.42)
9(SC,10)-(FE,36)-(EM,24) 0.95 (0.39)3.20 (0.55)
10 (TR,10)-(JL,2)-(TR,3)-(JL,55) 0.90 (0.36)3.73 (0.72)
Relative to the reference cluster (cluster 1), characterised by those who prolonged their
schooling for two years to sit A-level exams before successfully transitioning to higher edu-
cation, all slope coefficients are notably negative. All students achieving 5 or more grades
at A–C in GCSE exams are therefore less likely to belong to all other clusters, relative
to cluster 1. Thus, the reference level for the effect of GCSE5eq is appropriate and the
interpretation is guided only by the magnitude of the slope coefficients and their associated
standard errors, as well as the intercepts. Firstly, the effects for cluster 2 and 3, capturing
other subjects who were in higher education by the end of the observation period, appear
slight (on the basis of the size of the standard errors of their slopes). Coupled with the
negative intercepts for these clusters, this suggests, as expected, that more academically
inclined students tend to prolong their education in order to improve their job prospects.
Conversely, all other intercepts are positive and all other slope coefficients appear to be
significantly different from 0. We can say, therefore, despite the two-year continuation in
school of subjects in cluster 4, that students who do well in GCSE exams are less likely to
belong to this cluster. Furthermore, we can see the coefficient magnitudes increasing and
the standard errors decreasing as we move from cluster 5 to cluster 7. As these clusters are
distinguished only by the length of the training period prior to securing stable employment,
this again suggests that academically poor students are quick to find a job, presumably in
an unskilled capacity. Similar conclusions can be drawn for clusters 8 and 9, i.e. subjects
who secured employment of some kind after some time in further education rather than
third-level education. Finally, those who achieved 5 or more high GCSE grades are less
likely to experience persistent spells of joblessness (cluster 10).
The optimal G= 11 UUN model contains a uniform noise component. The BIC chooses
such a model over G= 10 models without a noise component and G= 11 models with all
non-noise components. Detecting outliers in this way allows the remaining non-noise clusters
to be modelled more clearly. Figure 8focuses on the noise component, which isolates errant,
directionless subjects who don’t neatly fit into any of the defined clusters and transition
quite frequently between states. This includes transitions in and out of further education,
employment, and training. Most subjects here are early school-leavers. Under the model’s
NGN gating network, the probability of belonging to this noise component is constant
(0.025) and independent of the included GCSE5eq covariate.
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Figure 8: Observations assigned to the noise component of the optimal G= 11 UUN model with the
GCSE5eq covariate in the NGN gating network.
7 Conclusion
The Status Zero Survey followed a sample of Northern Irish youths over a six-year pe-
riod, recording their employment activities at monthly intervals, in order to explore their
unfolding career trajectories and identify those at risk of prolonged unemployment. Here
we present a model-based clustering approach, with the aims of assessing how many typ-
ical trajectories there are, what kinds of typical trajectories there are, and what kinds of
individuals are more likely to experience which kinds of trajectories. Our approach is con-
trasted to heuristic approaches previously employed in analyses of these data. In McVicar
and Anyadike-Danes (2002), Ward’s hierarchical clustering algorithm is applied to an OM
dissimilarity matrix to identify relevant patterns in the data, with subjective costs. Notably,
reference is not made to the associated covariates until the uncovered clustering structure
is examined. In particular, MLR is used to relate the hard assignments of the sequences to
clusters to a set of baseline covariates. It is also notable that the sampling weights are incor-
porated only in the MLR stage and not in the clustering itself. This is arguably a three-step
approach, comprising the computation of pairwise string distances using OM (or some other
distance metric), the hierarchical or partition-based clustering, and the (weighted) MLR.
MEDseq models, conversely, offer a more coherent ‘one-step’ model-based approach. The
sequences are modelled directly using a finite mixture of exponential-distance models, with
the Hamming distance and weighted variants thereof employed as the distance metric. A
range of precision parameter settings have been explored to allow different time points con-
tribute differently to the overall distance. Thus, varying degrees of parsimony are accommo-
dated. Sampling weights are accounted for by weighting each observation’s contribution to
the pseudo likelihood. Dependency on covariates is introduced by relating the cluster mem-
bership probabilities to covariates under the mixture of experts framework. Thus, MEDseq
models treat the weights, the relation of covariates to clusters, and the clustering itself si-
multaneously. Hence, MEDseq provides a coherent framework for estimating the number
of clusters, identifying the relevant features of these patterns, and assessing whether these
patterns are somehow influenced or shaped by the subjects’ background characteristics.
Model selection in the MEDseq setting identifies a reasonable solution for the MVAD
data and shows that clustering the sequences in a holistic manner allows new insights to
be gleaned from these data. In particular, 11 distinct components are found, of which
10 have interpretable typical trajectories and one is an additional noise component which
captures deviant cases. Thus, supported by the use of an information criterion appropriate
for this model-based analysis, a more granular view of the MVAD cohort than the 5groups
uncovered in McVicar and Anyadike-Danes (2002) is provided. Furthermore, allowing for
the other covariates with which the sampling weights used here are defined, GCSE exam
performance at the end of the compulsory education period is found to be the most single
most important predictor of cluster membership.
Opportunities for future research are varied and plentiful. Co-clustering approaches
could be used to simultaneously provide clusters of the observed sequence trajectories and
the time periods (Govaert and Nadif,2013). Such an approach could be especially useful for
the UUN model type identified as optimal for the MVAD data, as it would reduce the number
of within-cluster period-specific precision parameters required. Indeed, parsimony has been
achieved in a similar fashion in the context of finite mixtures with Markov components
(Melnykov,2016a). Additionally, grouping trajectories across time as well would enable
more efficient summaries of the durations of the spells in specific states, which tend to be
long for the MVAD data. In particular, using co-clustering approaches which respect the
ordering of the sequences by restricting the column-wise clusters to form contingent blocks
would be particularly desirable. Indeed, failure to fully account for the temporal ordering of
events, due to the invariance of the Hamming distance to permutations of the time periods,
is a general limitation of our framework which future work will endeavour to address.
It may also be of interest for other applications to extend the MEDseq models to ac-
commodate sequences of different lengths, for which the Hamming distance is not defined.
These different lengths could be attributable to missing data, either by virtue of sequences
not starting on the same date, shorter follow-up time for some subjects, or non-response
for some time points. While the Hamming distance is only defined for equal-length strings,
adapting the MEDseq models to such a setting would be greatly simplified if aligning the
sequences of different lengths is straightforward. Another limitation of MEDseq models is
that time-varying covariates are not accommodated in the gating network. Notably, neither
of these concerns are relevant for the MVAD data.
However, our analysis of the MVAD data is limited by two aspects of the gating network
portion of our framework. The first substantive limitation relates to the WLBS approach
used for quantifying uncertainty in the MLR coefficients. As the sampling weights arise
from stratification, the standard errors obtained in this fashion are approximate. Thus,
examining alternative approaches to produce fully valid variance estimates in the MEDseq
setting in the presence of complex sampling designs is an interesting future research avenue.
The second limitation relates to the stepwise procedure used to identify relevant covari-
ates. As this strategy depends on an information criterion, namely the BIC, whose penalty
term is based on a parameter count, it may be prudent to relax the assumption that gating
covariates must affect all components. As the number of components chosen here (G= 11)
is moderately large, a large number of extra parameters are associated with each extra
covariate (see Appendix A). Thus, only GCSE5eq is identified as optimal, as it is signifi-
cantly associated with many of the typical trajectories. However, we note, for example, that
Catholics are largely underrepresented in cluster 7 and largely overrepresented in cluster 10
(characterised by persistent employment and persistent joblessness, respectively) despite the
omission of the covariate indicating religious affiliation from the optimal model. Incorpo-
rating regularisation penalties into the MLR to shrink certain gating network coefficients to
zero could thus be a fruitful alternative to the present stepwise covariate-selection method.
Another potential extension is to consider MEDseq models with alternative distance
metrics. The distance metric in García-Magariños and Vilar (2015), which accounts for
the temporal correlation in categorical sequences, is of particular interest; so, too, is OM.
In general, heuristic distance-based clustering (including fuzzy methods) can more easily
accommodate more sophisticated distances, while changing the MEDseq distance metric
fundamentally alters the model, which needs the normalising constant and the conditional
maximisation steps for parameter estimation to be tailored to the choice of metric.
MEDseq models, by virtue of being based on the Hamming distance for computational
reasons, implicitly assume substitution-cost matrices with zero along the diagonal and a
single value common to all other entries. The relationship between the exponent of an
EDM based on the Hamming distance and the Hamming distance itself (with a common
cost, typically equal to 1) is apparent from the fact that multiplying the substitution-cost
matrix by any positive scalar, as per normalised variants of the Hamming distance (Elzinga,
2007;Gabadinho et al.,2011), yields the same model, because its value is absorbed into λ.
This is also the case for models employing weighted Hamming distance variants under which
the precision parameters, and hence the otherwise common substitution costs, vary across
clusters and/or time points. However, all model types in the MEDseq family cannot account
for situations in which some states are more different than others — e.g. one where the cost
associated with moving from employment to joblessness is assumed to be greater than the
cost associated with moving from school to training — as they assume that substitution
costs are the same between each pair of states. Such concerns are most pronounced when
there is an explicit ordering to the states, e.g. education levels (Studer and Ritschard,2016).
Basing MEDseq on OM, for instance, would require the subjective specification, or
preferably estimation, of the v(v1)/2off-diagonal entries of symmetric substitution-cost
matrices. Potentially, as per the range of precision settings used for the MVAD application,
the substitution-cost matrices could also be allowed to vary across clusters and/or time
points. However, the normalising constant under an EDM using OM depends both on the
heterogeneous substitution costs and on θand is unavailable in closed form, thereby greatly
complicating model fitting. Indeed, dependence on θrenders even offline pre-computation of
the normalising constant infeasible for even moderately large Tor v. Truncation of the sum
over all sequences or importance sampling techniques could be used to address the intract-
ability. Though not a concern for the MVAD data, as one substitution is equivalent to a
deletion and an insertion for equal-length sequences, considering insertions and deletions
also would present further challenges. In any case, some level of approximation would be
required, while the ECM algorithm for MEDseq models based on simple matching is exact.
As well as removing the normalising constant’s dependence on θ, another positive con-
sequence of the homogeneity of substitution costs with respect to pairs of states under the
Hamming distance is that the ECM algorithm used for parameter estimation scales well
with the sequence length Tand the size of the alphabet v, especially since such normalising
constants need to be computed once, Gtimes, or G1times per iteration, depending on the
precision parameter settings. Though potentially restrictive, having only one parameter as-
sociated with each substitution-cost matrix, regardless of its order v, helps address concerns
about overparameterisation (Studer and Ritschard,2016), especially when the substitution
costs implied by the precision parameter(s) vary across clusters and/or time points.
Furthermore, it is likely that results on the MVAD data would not differ greatly with OM
used in place of the Hamming distance, particularly for models where λvaries across clusters
and/or time points, save for a solution with potentially fewer clusters being found. Indeed,
McVicar and Anyadike-Danes (2002) also consider a setting with common substitution costs
and find that their results do not greatly differ from their solution with state-dependent
costs. This implies that the notion that some states in the MVAD data are closer to each
other than others can be questioned. Ultimately, the UUN model adopted here preserves
the timing of events, by prohibiting time-warping insertion and deletion operations, while
accounting (in a cluster-specific fashion) for the timing, as well as the number, of element-
wise mismatches between sequences, in such a way that all states are assumed to be equally
different. Given the correspondence between Hamming distance weights, precision param-
eters, and implicit substitution costs in MEDseq models, it is notable that these are treated
as parameters rather than inputs, and are thus estimated as part of model fitting rather
than pre-specified along with the matrix of pairwise distances between sequences.
Overall, our analysis of the MVAD data provides a more granular view of the cohort of
Northern Irish youths than previously available, supplemented by interpretable parameter
estimates achieved through a coherent model-based framework. The MEDseq model family
appears promising from the perspective of reconciling the distance-based and model-based
cultures within the SA community. Indeed, the results for the MVAD data are encouraging
in this respect; they seem to suggest that the unconstrained precision parameter settings
adequately address the misalignment issues inherent in the use of the Hamming distance.
It remains to be seen if this holds for more turbulent sequences, e.g. those related to
employment activities tracked over longer periods.
This publication has emanated from research conducted with the financial support of Science
Foundation Ireland under Grant number SFI/12/RC/2289_P2. Additionally, R. Piccarreta
acknowledges the support from MIUR-PRIN 2017 project 20177BRJXS. For the purpose of Open
Access, the authors have applied a CC BY public copyright licence to any Author Accepted
Manuscript version arising from this submission. The authors also thank Matthias Studer and
members of the Sequence Analysis Association for helpful discussions.
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Appendix A The MEDseq Model Family: Parameter Counts
The models in the MEDseq family differ only in their treatment of the precision parameters,
which differentiate the Hamming distance and the weighted variants thereof. The BIC is
used in order to choose between the 8model types, identify the optimal G, and guide the
inclusion of gating covariates. Table A.1 summarises the number of free parameters kin the
BIC penalty term under each MEDseq model type, in order to demonstrate the increasing
level of complexity in moving from the most parsimonious CCN model to the most heavily
parameterised UU model.
The number of parameters contributing to each b
θgestimate notably depends on the
number of states represented across all cases in each time point. Note also that parameters
relating to b
θg,t corresponding to estimated precision parameters are counted, while those
associated with fixed precision parameter values of 0are not counted. Similarly, precision
parameters estimated as 0are counted, but precision parameters fixed at 0associated with
the noise component are not.
The number of gating network parameters is not accounted for in Table A.1. When
covariates are included, there are (r+ 1) ×(G1) or (r+ 1)×(G2) +1 extra parameters
— under the GN and NGN settings, respectively — where r+ 1 is the dimension of the
associated design matrix, including the intercept term. When τis not covariate-dependent,
there are G1extra parameters when τis unconstrained or only 1extra parameter if τis
constrained and the model includes a noise component, in which case τ0is allowed to vary.
Table A.1: Number of estimated parameters under each MEDseq model type. Models with names ending
with the letter N, indicating the presence of a noise component for which the single precision parameter is
fixed to 0, behave like the corresponding model without this component for all other components. Thus, λ
and all subscript variants thereof refer here to the non-noise components only.
Model Precision λg(Clusters) λt(Time Points) Number of Parameters
Central Sequence(s) Precision
CC λg,t =λConstrained Constrained GPT
t=1 (vt1) 1
t=1 (vt1) 1(G > 1)
UC λg,t =λgUnconstrained Constrained GPT
t=1 (vt1) G
t=1 (vt1) G1
CU λg,t =λtConstrained Unconstrained GPT
t=1 (vt1) T
t=1 (vt1) 1(G > 1)T
UU λg,t =λg,t Unconstrained Unconstrained GPT
t=1 (vt1) GT
t=1 (vt1) (G1) T
Appendix B Estimating MEDseq Precision Parameters
For fixed θ, the PMF in (1) belongs to the exponential family with natural parameter λ.
Thus, under any distance metric, the method of moments estimate of λis equal to the MLE.
Hence, with b
θalready estimated as per Section 4.1.2,b
λensures that the expected distance
of observations from b
θis equal to the observed average distance from b
θ, since the solution of
n∂λ =PσST
θ 1
θ)) = PσST
θ =dS,b
Under the Hamming distance, the value of the expectation in (7) holds for any arbitrary
reference sequence in place of b
θ. As the denominator in (7) — corresponding to the normal-
ising constant in (3), under the Hamming distance — is a function of λ, it is crucial that
it exists in closed form in order to estimate the precision parameter. Hence, with known b
the MLE for λfor an unweighted single-component CC model can be obtained as follows:
θnT log(v1) eλ+ 1,
∂ℓ (·)
∂λ =nT (v1)
eλ+ (v1) ndHS,b
λ= log (v1) T
which notably relies on the inverse of the average Hamming distance normalised by the
sequence length T. However, this can yield a negative value for b
λ. Recall that we only
consider λ0. Since all distances are non-negative and typically not identical, ∂ℓ(·)
∂λ is
negative λ > 0in the case where the sufficient statistic dHS,b
θ> v1T(v1), with
limλ→∞ ∂ℓ(·)
∂λ =ndHS,b
θ. Thus,
λ= max 0,log(v1) T
When dHS,b
θ< v1T(v1), such that b
λ > 0, the identity log(c(a/b 1)) = log(c) +
log(ab)log(b)is used for numerical stability, otherwise b
λis set to 0. When sampling
weights are included, following the same steps as above yields the corresponding estimate
λ= max 0,log(v1) + logT n
i=1 widHsi,b
The ECM algorithm is employed when G > 1, in which case the CM-step for b
under a CC MEDseq mixture model with sampling weights is given by
∂λ =T(v1) Pn
i=1 PG
g=1 zi,gwi
eλ+ (v1)
λ(m+1) = max
0,log(v1) + log T n
i=1 PG
g=1 bz(m+1)
i,g widHsi,b
As per (8), this requires the current estimate of each component’s central sequence. When
there are no sampling weights, one need only drop the witerms from (8) and (9) to esti-
mate the precision parameters of unweighted MEDseq models. While b
λcan potentially be
estimated as zero, the inclusion of a noise component in the CCN,UCN,CUN, and UUN
models makes this explicit, by restricting one cluster to have λg ,t = 0 t= 1,...,T.
However, when b
λg,t is estimated as zero rather than fixed to zero, the corresponding
θg,t parameter must be estimated, as it affects the likelihood indirectly through its role in
estimating the precision parameter(s). In particular — taking the UU model as an example
— all state values in the t-th sequence position with non-zero bz(m+1)
i,g are identical to b
when the corresponding denominator in Table B.2 evaluates to zero, such that b
g,t → ∞.
Expressions for the weighted complete data pseudo likelihood functions for all model
types in the MEDseq family are given in Table B.1. All models are written here as though
gating network covariates xiare included. However, the gating networks of models with a
noise component are written in the NGN form employed by the optimal model identified
in Section 5.1 rather than the GN form, i.e. it is assumed that τ0is constant, meaning the
covariates do not affect the probability of belonging to the noise component (see Section 3.4).
Table B.2 outlines the corresponding CM-steps for the precision parameter(s). All deriva-
tions closely follow the same steps as in (9) for the CC model and the normalised sampling
weights are accounted for in all cases. These formulas can be simplified somewhat for un-
weighted models and/or models without gating covariates. Recall that the first letter of
the model name denotes whether the precision parameters are constrained/unconstrained
across clusters, the second denotes the same across time points (i.e. sequence positions),
and model names ending with the letter Ninclude a noise component.
Table B.1: Weighted complete data pseudo likelihood functions for all MEDseq model types, which dif-
fer according to the constraints imposed on the precision parameters across clusters and/or time points.
The expressions for the various weighted Hamming distance metric variants employed, and the associated
normalising constants, are given in full.
Model Weighted Complete Data Pseudo Likelihood
i=1 QG
g=1 τg(xi)exp(λPT
t=1 1(si,t 6=θg,t ))
((v1)eλ+1)Tzi,g wi
i=1 QG
g=1 τg(xi)exp(λgPT
t=1 1(si,t 6=θg,t ))
((v1)eλg+1)Tzi,g wi
i=1 QG
g=1 τg(xi)exp(PT
t=1 λt1(si,t 6=θg,t))
t=1((v1)eλt+1) zi,g wi
i=1 QG
g=1 τg(xi)exp(PT
t=1 λgt1(si,t 6=θg,t))
t=1((v1)eλg,t +1)zi,g wi
i=1 QG1
g=1 τg(xi)exp(λPT
t=1 1(si,t 6=θg,t))
((v1)eλ+1)Tzi,g τ0
i=1 QG1
g=1 τg(xi)exp(λgPT
t=1 1(si,t 6=θg,t ))
((v1)eλg+1)Tzi,g τ0
i=1 QG1
g=1 τg(xi)exp(PT
t=1 λt1(si,t 6=θg,t))
t=1((v1)eλt+1) zi,g τ0
i=1 QG1
g=1 τg(xi)exp(PT
t=1 λg,t1(si,t6=θg,t))
t=1((v1)eλg ,t +1)zi,g τ0