Journal of Child Psychology and Psychiatry 61:5 (2020), pp 565–574
doi:10.1111/jcpp.13145
Trajectories of depression symptom change during
and following treatment in adolescents with unipolar
major depression
Sian Emma Davies,1 Sharon A.S. Neufeld,1 Eleonore van Sprang,2 Lizanne Schweren,3
Rogier Keivit,4 Peter Fonagy,5
Bernadka Dubicka,6 Raphael Kelvin,1 Nick Midgley,5
7
Shirley Reynolds, Mary Target,5 Paul Wilkinson,1 Anne Laura van Harmelen,1 and
Ian Michael Goodyer1
1
Department of Psychiatry, University of Cambridge, Cambridge, UK; 2Department of Psychiatry, Amsterdam UMC,
VUmc, Amsterdam, The Netherlands; 3Department of Psychiatry, University Medical Center Groningen, Groningen,
The Netherlands; 4MRC Cognition and Brain Sciences Unit, University of Cambridge, Cambridge, UK; 5Research
Department of Clinical, Educational and Health Psychology, Division of Psychology and Language Sciences,
University College London, London, UK; 6Department of Psychiatry, University of Manchester, Manchester, UK;
7
School of Psychology and Clinical Language Sciences, University of Reading, Reading, UK
Objective: To classify a cohort of depressed adolescents recruited to the UK IMPACT trial, according to trajectories of
symptom change. We examined for predictors and compared the data-driven categories of patients with a priori
operational definitions of treatment response. Method: Secondary data analysis using growth mixture modelling
(GMM). Missing data were imputed. Trajectories of self-reported depressive symptoms were plotted using scores
taken at six nominal time points over 86 weeks from randomisation in all 465 patients. Results: A piecewise GMM
categorised patients into two classes with initially similar and subsequently distinct trajectories. Both groups had a
significant decline in depressive symptoms over the first 18 weeks. Eighty-four per cent (84.1%, n = 391) of patients
were classed as ‘continued-improvers’ with symptoms reducing over the duration of the study. A further class of
15.9% (n = 74) of patients were termed ‘halted-improvers’ with higher baseline depression scores, faster early
recovery but no further improvement after 18 weeks. Presence of baseline comorbidity somewhat increased
membership to the halted-improvers class (OR = 1.40, CI: 1.00–1.96). By end of study, compared with classes, a
clinical remission cut-off score (≤27) and a symptom reduction score (≥50%) indexing treatment response
misclassified 15% and 31% of cases, respectively. Conclusions: A fast reduction in depressive symptoms in the
first few weeks of treatment may not indicate a good prognosis. Halted improvement is only seen after 18 weeks of
treatment. Longitudinal modelling may improve the precision of revealing differential responses to treatment.
Improvement in depressive symptoms may be somewhat better in the year after treatment than previously
considered. Keywords: Depression; therapy; longitudinal studies; outcome.
Background
Adolescence denotes the highest incidence risk rate
period for the emergence of major depression (MD) over
the lifecourse (Avenevoli, Knight, Kessler, & Merikangas, 2008). The effectiveness of current treatment
strategies, both psychological and SSRI medication
alone or in combination (NICE, 2015), have moderate
effect sizes of between 0.3 and 0.6 (March et al., 2004;
Weisz et al., 2017). At least 20% of adolescents with MD
show no treatment response (Goodyer et al., 2008) but
the reasons for this are unclear. Currently, there are
large variations in the definition of response between
trials (Berlim & Turecki, 2007). Such discrepancies
lower comparability between studies and impact the
proportions of patients considered responders or nonresponders, respectively (Uher et al., 2010; Vitiello
et al., 2011). Response definitions are based on
Conflict of interest statement: See Acknowledgements for full
disclosures.
The copyright line for this article was changed on 7 Nov 2019
after original online publication.
percentage symptom reduction (treatment response)
or final scores below an a priori cut-off (clinical
remission). These methods are arbitrary and may lack
clinical meaning (Thibodeau et al., 2015; Uher et al.,
2010). Furthermore, there can be an overlap of patients
meeting criteria for nonremission (e.g. a final Hamilton
Rating Scale for Depression (HRSD) score of ≥7) and a
positive clinical response (e.g. reduction of ≥50% in
HRSD) (Fu et al., 2008).
Empirical modelling techniques, such as growth
mixture modelling (GMM), may address some of the
validity issues with a priori definitions, by categorising patients post hoc (Ram & Grimm, 2009). This
computational technique searches for naturally
occurring heterogeneity to categorise patients into
particular latent classes that follow similar trajectories and make no a priori assumptions on what
constitutes a meaningful response (Bri
ere, Rohde,
Stice, & Morizot, 2016; Thibodeau et al., 2015; Uher
et al., 2010). GMM describes the trajectory of relatively homogeneous behavioural groups and how
they differ from each other in their shape over time
(Gueorguieva, Mallinckrodt, & Krystal, 2011; Ram &
© 2019 The Authors. Journal of Child Psychology and Psychiatry published by John Wiley & Sons Ltd on behalf of Association for Child and
Adolescent Mental Health.
This is an open access article under the terms of the Creative Commons Attribution License, which permits use, distribution and reproduction in any
medium, provided the original work is properly cited.
566
Sian Emma Davies et al.
Grimm, 2009). Homogeneity of groups also aids
investigating predictors of response types that are
likely to have small effect sizes.
Growth mixture modelling analyses of treatment
trials data in depressed adults reveal a variety of
multiple, qualitatively distinct classes, specific predictors and differential therapeutic responses (Cuijpers,
van Lier, van Straten, & Donker, 2005; Gueorguieva
et al., 2011; Stulz, Thase, Klein, Manber, & CritsChristoph, 2010; Thibodeau et al., 2015; Uher et al.,
2010). One recent report from the Treatment of Adolescent Depression Study (TADS) noted that, at 12 weeks,
there were two groups that improved over the trial, and a
further group showing limited change (Scott, Lewis, &
Marti, 2019). A depression prevention study noted two
groups where symptoms gradually reduced over time;
one group showed no change, and the other reported
resurgent symptom count within 6 months of the study
end (Bri
ere et al., 2016). Unfavourable trajectories have
been associated with older age, psychosocial function,
higher anxiety symptom levels and psychotic experiences (Gueorguieva et al., 2011; Jeppesen et al., 2015;
Perlis et al., 2010; Thibodeau et al., 2015).
The duration of follow-up plays a contributory role
in determining response classes. Thibodeau et al.
(2015) found that short-term follow-up mistakenly
classified some responders as nonresponders, whilst
Bri
ere and colleagues (Bri
ere et al., 2016) found a
subgroup of adolescents that showed a significant
decline in symptoms up to 6 months, but relapsed
only after this point. Longer-term follow-ups past the
end of the treatment may improve the precision of
denoting true responders, sustained nonresponders
and relapsing patients.
Objectives
Our primary objective was to reveal trajectories of
depression symptoms from randomisation to the
final assessment, approximately one year following
the end of treatment. The specific aims were to: (a)
define the number and shape of longitudinal classes
of patients revealed from depression symptoms only;
and (b) compare the defined groups with standard a
priori definitions of response/remission.
Our second objective was then to test whether
selected baseline demographic and clinical characteristics would predict class membership.
Methods
Study design
This study was a reanalysis of data from the Improving Mood
with Psychoanalytic and Cognitive Therapies (IMPACT) trial
(Goodyer et al., 2017a; 2017b). As there we no difference in
clinical effects between psychological treatments, these were
collapsed for the present study, to investigate whole population. Self-reported depressive symptoms were measured at six
nominal time points: baseline, 6, 12, 36, 52 and 86 weeks
postrandomisation. The last two time points were post treatment which was completed by 36 weeks in >95% of the cohort.
J Child Psychol Psychiatr 2020; 61(5): 565–74
Participants
Adolescents aged between 11 and 17 years, with major
depression (DSM-IV American Psychiatric Association, 2000),
were enrolled from 15 UK National Health Service (NHS)
clinics, 5 each in North London, North West England and East
Anglia. Patients with a lifetime history of mania were excluded.
Full details on patient inclusion and exclusion criteria can be
found in the study protocol (Goodyer et al., 2011). Four
hundred and sixty five patients who were included in the trial
had data available for the current analysis (Goodyer et al.,
2017a; 2017b).
Variables
Symptom trajectory class membership was defined using the
self-reported Mood and Feelings Questionnaire (MFQ) score
across all time points. This is a 33-item Questionnaire
(Burleson Daviss et al., 2006) of depressive symptomatology
covering the past 2 weeks measured on a 3-point scale (almost
never, sometimes, often/almost always). Higher sum scores
(range of 0–66) indicated more depressive symptoms and were
positively correlated with greater psychosocial impairment
(Goodyer et al., 2017a; 2017b).
Baseline variables investigated for their potential predictive value for trajectory class membership were as follows:
sum scores from self-report measures for anxiety (the
Revised Children’s Manifest Anxiety Scale, RCMAS, Reynolds
& Richmond, 1978), obsessionality (the short Leyton Obsessional Inventory for adolescents (LOI), (Bamber, Tamplin,
Park, Kyte, & Goodyer, 2002) and psychosocial impairment
(the Health of the Nation Outcome Scales for Children and
Adolescents, HoNOSCA, Gowers et al., 1999). Lifetime suicide attempts were defined as binary variables (yes, no) from
data derived from the Columbia Suicide Severity Rating Scale
(Posner et al., 2011). Lifetime nonsuicidal self-injury was
measured using the self-report Risk and Self Harm Inventory
for adolescents (Vrouva, Fonagy, Fearon, & Roussow, 2010).
The Kiddie-Schedule for Affective Disorder and Schizophrenia
(K-SADS, Kaufman et al., 1997) interview assessed psychiatric symptoms and diagnoses. Comorbidity was defined on
an ordinal scale, as the number of concurrent diagnoses.
Psychotic symptoms were counted from an ordinal scale
(absent, present: subthreshold, or present: threshold)
obtained from either of the two K-SADS screening questions
for psychosis.
Statistical methods
Imputation. Multiple imputation was used in order to
maximise sample size and reduce selection bias. Due to a
wealth of auxiliary variables predicting missingness, data were
presumed to be missing at random. Additional nonmissing
variables were also included to improve model prediction. Full
details of imputation rationale and statistical procedure are
given in Appendix S1.
Growth mixture models. There was substantial variation in the timing of each assessment and to model symptom
change accurately, the mean time of each actual assessment
(weeks) was taken as the focal point for GMMs. This corresponded to baseline, 12, 18, 43, 60 and 95 weeks postrandomisation. Variation in time of assessment was further
included as a covariate in all growth models.
We tested GMMs using the Mplus program version 8.0
(Muth
en & Muth
en, 2017). Four growth trends were considered: a linear and quadratic growth trend and two linearpiecewise growth trends. We investigated piecewise growth
trends because we hypothesised that different rates of
© 2019 The Authors. Journal of Child Psychology and Psychiatry published by John Wiley & Sons Ltd on behalf of Association for
Child and Adolescent Mental Health.
doi:10.1111/jcpp.13145
Depression, treatment, trajectories, outcome
improvement might occur during treatment versus follow-up
stages of the trial. Due to the average length of treatment
falling between the means of two assessment time points, we
considered two possible transition points in the models: the
first was placed at the third assessment (18 weeks on average
from baseline) and the second at the fourth assessment
(43 weeks from baseline).
Classes were incrementally added to the single class model
to determine the best fit. All models allowed for within-class
variation, whereby patient’s symptom scores vary around the
mean of the group. However, upon testing, there was no
evidence of significant variation between classes, (See
Appendix S1 and Table 1), meaning that the variation around
each group mean was not significantly different for the
respective classes. Consequently, between-class variation
was held equal for all growth factors. Only solutions that
were replicated with different starting values were accepted.
We considered models with 1–5 trajectory classes, and
retained the most parsimonious model. Full details of statistical rules for determining parsimony are given in
Appendix S1.
Baseline clinical characteristics and predictors of
class membership. Univariate analyses were conducted
(chi-square, or t-tests) to determine whether there were significant differences between classes on baseline demographic and
clinical characteristics. Mann–Whitney–Wilcoxon tests were
performed where data were non-normal. Multinomial logistic
regression was then used to determine which variables
predicted class membership. R-squared statistic indicated
how much variance the regression model explained. Odds
Ratios are reported for all predictors.
Agreement between categorical definitions and
GMM model result. Cohen’s Kappa coefficient (Cohen,
1960) was used to test the agreement between trajectory
classes and the definitions of: (a) treatment response/nonresponse (50% reduction/or not in MFQ score by end of study) or
(b) clinical remission/nonremission (MFQ score at 12 months
below or above 27) (Kent, Vostanis, & Feehan, 1997).
Results
Participants
All 465 participants who entered the trial were
available for the longitudinal analysis. Across all
assessments, 65% or more of the sample (304/465)
had full data on all MFQ items.
567
Outcome data
Fit information for all piecewise class models tested
is provided in Table 1. A two-class, piecewise model
that separately modelled the change in depressive
symptoms, linearly, over the first 18 weeks of treatment (on average; assessments 0–2), and then the
remaining period of the trial (assessments 3–5), was
identified as the optimal model. This is illustrated in
Figure 1. BIC showed a favourable decrease of
approximately 42 with the addition of a second class
from the single class solution. Although the 3-class
solution yielded a slightly lower BIC (DBIC = 4.039),
the decrease in entropy below commonly accepted
thresholds suggests poor classification precision in
the 3-class solution (entropy = 0.844 vs. 0.734). The
Mplus code for the two-class model is provided in
Appendix S2.
The two-class piecewise GMM divided patients into
a comparatively large class of 391 (84.1%; class 1)
and a small class of 74 (15.9%; class 2). The mean
depression scores and the percentage change
between time points are shown in Table 2.
Class 2 on average showed significantly higher
baseline levels of MFQ scores than class 1 (Wald
X2(1) = 25.577, p < .001). Both classes showed a significant decrease in MFQ score over the first 18 weeks
of the trial ( 6.466, p < .001, and 8.794, p < .001,
respectively). There was however, a significantly faster
rate of MFQ reduction of class 2 compared with class 1
over the first 18 weeks (Wald X2(1) = 5.446, p = .0196).
For class 2, there was no further statistical difference in rate of change over the rest of the trial (0.899,
p = .183). Conversely, class 1, on average, continued
to show a significant decline in MFQ score, albeit
slower than in the initial 18 weeks ( 1.639,
p = .014). This difference between the second linear
slopes of the two classes was statistically significant
(Wald X2(1) = 167.075, p < .001). By the end of the
trial, Class 1 showed on average a 60.5% improvement in depressive symptoms, compared with 11.0%
in class 2. We labelled class 1 as ‘continued-improvers’ and class 2 as ‘halted-improvers’.
Table 1 Model fit information for piecewise GMM-CIs
Fit Statistics
GMM-CIs
LL (No. of parameters)
AIC
BIC
Entropy
Group size (%)
C1
C2
C3
C4
C5
1 Class
2 Classes
3 Classes
4 Classes
5 Classes
10,487.996
21,015.992
21,098.832
1
10,454.836
20,957.672
21,057.081
.844
10,440.533
20,937.065
21,053.042
.734
10,431.253
20,926.506
21,059.051
.718
10,423.269
20,918.538
21,067.651
.729
465 (100%)
–
–
–
–
391 (84.1%)
74 (15.9%)
–
–
–
329 (70.7%)
77 (16.6%)
59 (12.7%)
–
–
191 (41.1%)
161 (34.6%)
57 (12.3%)
56 (12.0%)
–
219 (47.1%)
109 (23.5%)
56 (12.0%)
54 (11.6%)
27 (5.8%)
© 2019 The Authors. Journal of Child Psychology and Psychiatry published by John Wiley & Sons Ltd on behalf of Association for
Child and Adolescent Mental Health.
568
Sian Emma Davies et al.
J Child Psychol Psychiatr 2020; 61(5): 565–74
Figure 1 Sample (shown in large grey/black) and estimated (shown in small black) means for the two-class piecewise growth mixture
model. Class 1 reveal a continued-improver class, n = 391 (84.1%) of the population. Class 2 reveal halted-improver class, n = 74 (15.9%)
of the population. Behind plots every individual patient’s trajectory
Table 2 Estimated and observed mean values for MFQ scores and observed mean percentage improvement in MFQ scores are given
for both latent classes
Assessment point in
average weeks from
baseline
0
12
18
43
60
95
Class 1: Continual improvement (n = 391)
Class 2: Halted improvement (n = 74)
MFQ sum scores
MFQ sum scores
Estimated
Weighted
estimates
Observed
44.828
37.073
32.881
25.877
23.039
17.247
44.810
34.649
32.752
23.938
22.291
17.797
44.774
34.623
32.689
25.920
22.224
17.673
% observed
improvement
from baseline
22.672
26.991
42.109
50.364
60.528
Estimated
Weighted
estimates
Observed
51.638
41.090
35.390
39.232
40.789
43.966
51.721
38.191
36.134
38.243
39.034
44.978
52.096
38.420
36.558
38.669
39.835
46.357
% observed
improvement
from baseline
26.252
29.826
25.774
23.535
11.016
Baseline characteristics of each class
Predictors of trajectory class membership
The characteristics of patients following each trajectory class are described in Table 3.
There were more females in the ‘halted’ compared
with the ‘continued’ improvers class (85% vs. 73%).
No other demographic characteristics were significantly different between groups.
Compared with. ‘continued-improvers’, ‘haltedimprovers’ on average showed significantly higher
obsessionality (LOI) and HoNOSCA scores, more
nonsuicidal self-injury (NSSI), and greater
numbers of comorbid diagnoses at baseline.
Interestingly neither baseline levels of suicidal
attempts, psychotic symptoms nor treatment
group discriminated between the classes. There
were no significant differences between the two
classes on any baseline characteristic (Table 3).
Trajectories of classes were similar for HoNOSCA
as for MFQ, as a concurrent validation check
(Appendix S3).
Class 1 acted as the reference class for the logistic
regression (Table 4). When controlling for variables
included in the model, only the number of comorbid
psychiatric diagnoses at baseline significantly predicted
a higher probability of membership to halted-improvers
compared with continued-improvers (Table 4). With
each increasing number of comorbid diagnoses the
odds of a patient belonging to the ‘halted’ compared with
the ‘continued-improvers’ class increased by a factor of
1.4 (X2(4) = 46.03, p < .001). The finding however only
explained 5.4% of the total variance in class membership (Cox and Snell R2 = .054).
Agreement between classes and categorical
definitions of response
The agreement in classification between data-driven
classes and a priori categorical classifications are
shown in Figure 2.
© 2019 The Authors. Journal of Child Psychology and Psychiatry published by John Wiley & Sons Ltd on behalf of Association for
Child and Adolescent Mental Health.
doi:10.1111/jcpp.13145
Depression, treatment, trajectories, outcome
569
Table 3 Characteristics of subjects following the two latent trajectories
Demographics
Female
Age
Region
East Anglia
North London
North West England
Ethnicity (white)
Index of multiple deprivation (IMD)a
Baseline clinical characteristics
RMASb
LOIc
Suicidal thoughts
Suicidal attempts
NSSI
HONOSCA (available for 435 patients)
Comorbidityd
1
2
3
4
Psychotic symptoms
Subthreshold
Threshold
Treatment characteristics
Treatment arm
BPI
CBT
STPP
Baseline SSRI prescription
Class 1: Continuedimprovers (n = 391)
Class 2: Haltedimprovers (n = 74)
Comparison
Mean (n)
SD (%)
Mean (n)
SD (%)
X2/t/W
p
285
15.6
–
161
105
125
325
23.4
72.8%
1.4
–
41.2%
26.9%
32.0%
83.1%
–
63
15.7
–
24
22
28
65
27.7
85.1%
1.3
–
32.4%
29.7%
37.8%
87.8%
–
4.955
0.459
2.035
–
–
–
1.024
13,446
.026
.647
.361
–
–
–
.312
.336
40.7
9.6
345
131
218
18.3
–
121
59
5
0
–
87
32
7.3
5.1
88.2%
33.5%
55.8%
6.0
–
30.9%
15.1%
1.3%
0%
–
23.6%
8.8%
42.3
11.8
69
28
53
19.9
–
26
18
3
1
–
16
10
6.7
5.6
93.2%
37.8%
71.6%
6.3
–
35.1%
24.3%
4.1%
1.4%
–
22.5%
14.1%
1.863
3.124
1.600
0.519
6.443
2.018
10.20
–
–
–
–
2.024
–
–
.065
.002
.206
.471
.011
.046
.006
–
–
–
–
.363
–
–
–
127
127
137
87
–
32.5%
32.5%
35.0%
22.3%
–
28
27
19
10
–
37.8%
36.5%
25.7%
13.5%
2.463
–
–
–
2.877
.292
–
–
–
.090
a
IMD: Median reported for nonparametric tests.
Revised Manifets Anxiety Scale- self-report.
c
Leyton Obsessional Inventory- self-report.
d
Due to insufficient cell size, variable was recorded as 0,1 and 2+ to meet assumptions of chi-square test.
b
Table 4 Baseline predictors of trajectory class membership:
clinical characteristics
Class 1: Halted-improvers
Gender
Age
RCMAS
LOI
HONOSCA
Attempts
NSSI
Psychotic symptoms
Comorbidity
OR
95% CI
0.50
1.02
0.98
1.06
1.03
0.98
1.77
0.97
1.40*
0.23–1.02
0.84–1.26
0.94–1.03
0.99–1.12
0.98–1.08
0.53–1.79
0.95–3.41
0.64–1.45
1.00–1.96*
*<.05.
Percentage symptom reduction
There was moderate agreement between trajectory
class membership and clinical categorical outcome
when ‘treatment response’ is defined by percentage
reduction of 50% by end of study (k = .412,
p < .001). All halted-improvers were also treatment
nonresponders by this definition. However, only 269
of 391 (69%) of continued-improvers were also
treatment responders. The remaining continuedimprovers (122 of 391; 31%) were classified as
treatment nonresponders on the per cent reduction
category.
Clinical cut-off
There was stronger, albeit still moderate, agreement
between trajectory membership and clinical remission, when defined by a cut-off score of 27 on the
MFQ (k = .642, p < .001) at end of study. All haltedimprovers were also clinical nonremitters. However,
only 332 of 391 (85%) of continued-improvers were
classified as in clinical remission. The remaining
continued-improvers (59 of 391; 15%) were classified
as not clinically remitted.
Overall, if either a priori category definitions had
been used in this study a false negative classification
rate of either 15% or 31% would have been
reported.
© 2019 The Authors. Journal of Child Psychology and Psychiatry published by John Wiley & Sons Ltd on behalf of Association for
Child and Adolescent Mental Health.
570
Sian Emma Davies et al.
J Child Psychol Psychiatr 2020; 61(5): 565–74
Figure 2 Sample means for the three categorical approaches. (A) Trajectory classes. (B) Percentage reduction for treatment response. (C)
Clinical cut-off for remission. Behind plots every individual patient’s trajectory
© 2019 The Authors. Journal of Child Psychology and Psychiatry published by John Wiley & Sons Ltd on behalf of Association for
Child and Adolescent Mental Health.
doi:10.1111/jcpp.13145
Discussion
Key results and interpretation
Using a cohort of depressed adolescents recruited
into a clinical trial we computed a depression symptom trajectory that revealed a piecewise function,
with two separate linear trajectories. The best fit
model was for two classes defined as a large (84.1%)
group of continued-improvers and a small (15.9%)
group of halted-improvers. We noted that both
groups improved significantly over the first 18 weeks
of the trial. The halted group showed a cessation in
improvement from thereon, and this may index a
putative relapsing group but this requires replicaton.
Although there are studies that support the existence of more than two classes, these applied constraints on investigated trajectory shape (Stulz et al.,
2010; Thibodeau et al., 2015) and eliminated withinclass variation from their models (Bri
ere et al.,
2016). These methodological choices are often necessary when models struggle to converge (Wickrama
et al., 2016). To assume that no individual variation
exists within classes in depressive patients, we felt
was not representative of real data. We therefore
favoured a GMM analysis at outset, which would
allow for within-class variation. Our results were
stable across different sets of random starting values
without these constraints, offering a much more
representative model of patient experience of depressive symptom change.
A striking finding in the present study was the great
contrast between trajectories of the two groups across
both parts of the model. The faster decline of haltedimprovers may itself be due to the higher depression
ratings at baseline. This does not necessarily index
‘good treatment response’: indeed clinicians may need
to consider that a fast reduction in depressive symptoms may imply subsequent halted improvement.
Further, treatment response cannot be prognostically
assessed adequately before approximately 18 weeks.
The precise psychological treatment implications however cannot be determined as there was no placebo
control group: we do not know whether this period of
rapid improvement is due to the psychological therapy,
a nonspecific response to receiving assessment and
treatment, or regression to the mean. Nevertheless,
these findings indicate that clinical care should be
taken for those adolescents with greater levels of
comorbidity and depression severity at baseline, even
when there is an initial rapid response to treatment.
Previous studies reporting two classes describe
their trajectory groups as either rapid and gradual
responders, or responders and nonresponders
(Gueorguieva et al., 2011; Uher et al., 2010). One
recent study of adolescents reported three trajectory
classes with the two improving classes merging by
end of treatment (Scott et al., 2019). However, those
studies were limited to short trial durations of
12 weeks or less and were not able to assess
Depression, treatment, trajectories, outcome
571
longer-term outcome. The shape of the halted-improvers trajectory in this clinical study resonates with
the symptom resurgence group reported by Briere and
colleagues in their community-sample depression
prevention study (Bri
ere et al., 2016). Their resurgence group showed a similar rapid initial improvement in symptoms, followed by a rapid decline in
symptoms over time. Whilst the slope value for the
second section of our trajectory in the halted-improvers did not reach significance with this cohort,
visual inspection suggests this trend. This trajectory
shape suggests that different underlying therapeutic
mechanisms may be activated in early and subsequent treatment responses. For example, the haltedimprovers may contain patients who are clinically
relapsing as well as not responding. What factors
account for the 18-week assessment as the optimal
break point of responding is not clear. Longer-term
follow-up is essential in future studies to disentangle
group trajectory patterns more accurately and to
reveal the most valid prognostic markers for treatment response both early and later in follow-up.
Mediation analyses will be required to examine factors
and mechanisms involved in early relapse and to
separate these from nonresponse.
Despite baseline univariate differences, only
comorbidities at baseline was retained as a significant independent predictor with a small odds ratio
and low predictive power. We conclude that baseline
demographic and clinical observations are insufficient to fully predict depression symptom change
over time. We also note that self-reported illness
episode duration was not a moderator of treatment
response in the primary results nor here (Goodyer
et al., 2008). Although data were not available in this
study, we conjecture that, consistent with a number
of other randomised controlled trials, a history of
childhood maltreatment could have been a moderator of trajectory group membership, with a positive
history increasing the liability for being a haltedimprover (Goodyer & Wilkinson, 2019). Further we
note that, although patients with a history of mania
were excluded from the study, such symptoms may
have emerged over the treatment phase. Indeed, we
recognise that a next step from this report is to
consider what factors over the course of treatment
and then follow-up phases may influence membership of the continued or halted-improvers groups.
We plan to conduct mediation analyses to examine
the effects of treatment dose, symptom and psychosocial changes and SSRI prescribing over the
treatment phase and their possible impact on membership of continued or halted outcome groups.
Finally, the findings showed only moderate agreement between empirical and a priori definitions of
symptom change indexing clinical response or
remission. This is similar to results of previous
studies (Gueorguieva et al., 2011; Uher et al.,
2010). The findings here have provided a clear cut
© 2019 The Authors. Journal of Child Psychology and Psychiatry published by John Wiley & Sons Ltd on behalf of Association for
Child and Adolescent Mental Health.
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Sian Emma Davies et al.
consistent group of halted-improvers by end of
study, potentially predicted by higher depression
scores and levels of comorbidity at entry combined
with halted improvement by 18 weeks. This suggests
caution with the use of cut-off or percentage change
measures as indices of achieving clinical remission,
as it is possible that a significant percentage of
potentially responding patients might be misclassified too quickly as nonresponders to treatment. Until
more is known about subclasses of depression,
researchers must take care in their choice of outcome measure and in particular to try to minimise a
false negative result.
Limitations
We cannot use these findings to generalise to
population-based trials where recruitment takes
place from schools, community settings or patients
with distinct cultural differences to those in this UK
NHS study. We do not know if the addition of parent
or teacher reporting, as opposed to self-reported
measures would have aided these research findings.
GMM are large-sample statistical techniques, and
whilst a sample of 465 is sufficient, a sample of 600
or more may have seen the emergence of a more
stable 3-class model. The lack of nonsymptom
driven 86-week outcome validators, such as interpersonal function, or data on history of childhood
maltreatment, are other limitations of this paper.
However, investigation of HONOSCA, a measure of
functioning as well as psychiatric symptomatology,
showed similar trends to MFQ (Appendix S3), which
provides preliminary external validity for our findings. In future, GMM trajectories in clinical populations with multiple symptom profile could utilise
nondepressive symptoms in their longitudinal
analyses.
J Child Psychol Psychiatr 2020; 61(5): 565–74
Supporting information
Additional supporting information may be found online
in the Supporting Information section at the end of the
article:
Appendix S1. Statistical methods.
Appendix S2. Mplus code for the two-class, piecewise
model.
Appendix S3. Validation with HONOSCA.
Acknowledgements
The study was funded by the National Institute for Health
Research (NIHR) Health Technology Assessment (HTA)
programme (project number 06/05/01) and Medical
Research Council (SUAG/04/ G101400). E.D. was
funded by a doctoral studentship awarded by the Neuroscience in Psychiatry (NSPN.Org) Consortium itself
funded by a strategic award from the Wellcome Trust
(095844/Z/11/Z) awarded to I.M.G. and P.F.. A.V.H. is in
receipt of a Dorothy Hodgkin Research Fellowship from
the Royal Society. S.E.D. and I.M.G. wrote the initial draft
of the manuscript. All authors contributed to and
approved the final manuscript. The authors would like
to thank the project manager and all research assistants
who collected and curated the impact data over the three
research sites. The views expressed in this publication are
those of the authors and do not necessarily reflect those of
the HTA programme, NIHR UK, National Health Service,
or the Department of Health, UK. I.M.G., P.O.W. and R.K.
received consulting fees from Lundbeck. The remaining
authors have declared that they have no competing or
potential conflicts of interest.
Correspondence
Sian Emma Davies and Ian Michael Goodyer, Department of Psychiatry, School of Clinical Medicine, University of Cambridge, 18b Trumpington Road, Cambridge,
CB2 8AH, UK; Emails: sed48@medschl.cam.ac.uk
(S.E.D.) and ig104@cam.ac.uk (I.M.D.)
Key points
At least 20% of moderate to severely depressed adolescents show no response to current treatments.
Modelling of depressive symptom trajectories of change suggested that 84.1% of depressed adolescents
could be considered as ‘continued responders’, whilst 15.9% could be classed as ‘halted responders’, that is
showing significant decrease in depression severity over the first 18 weeks, which then halted over the
subsequent year.
By the end of study, ‘continued responders’ showed on average a 60.5% improvement in depressive
symptoms, compared with 11.0% amongst the halted-improvers.
A fast reduction in depressive symptoms may not indicate a good prognosis.
Clinical progress continues in the year after treatment and overall may be better than previously considered.
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Accepted for publication: 24 September 2019
First published online: 24 October 2019
© 2019 The Authors. Journal of Child Psychology and Psychiatry published by John Wiley & Sons Ltd on behalf of Association for
Child and Adolescent Mental Health.