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Australian data and psychometric properties
of the Strengths and Difficulties
Questionnaire
David J. Hawes, Mark R. Dadds
Objective:
We examine the Australian psychometric properties of the Strengths and
Difficulties Questionnaire (SQD), a brief screening measure of behavioural and emotional
problems in children and adolescents.
Method:
Using a large community sample (n = 1359) of young Australian children
(4–9 years), we assessed the internal consistency, stability, and external validity of the
parent-report SDQ. Normative data and cut-offs were also produced.
Results:
Moderate to strong internal reliability was exhibited across all SDQ subscales, and
support was found for the original five-factor structure of the measure. Adequate validity was
evidenced in the relationship of these scales to one another, while correlations between the
SDQ subscales, teacher ratings, and diagnostic interviews demonstrated sound external
validity. SDQ total difficulties scores were associated with concurrent treatment status and
scores over a 12-month period were stable.
Conclusions:
The current study of the SDQ with Australian children presents evidence of
sound psychometric properties. Being the first study to empirically support the use of the SDQ
in Australia, it is recommended that the youth and teacher-report forms of the measure
receive similar attention in the future.
Key words:
attention deficit disorder with hyperactivity, child behaviour disorders, child
Australian and New Zealand Journal of Psychiatry 2004; 38:644–651
behaviour, clinical psychology, psychometrics.
The Strengths and Difficulties Questionnaire (SDQ)
[1], is a brief screening measure that is being employed
increasingly for the purpose of identifying behavioural
and emotional problems in children and adolescents.
Developed in the UK, the instrument produces scores for
each of five subscales: conduct problems; hyperactivity;
emotional symptoms; peer problems; and prosocial
behaviour. Each of these consists of five items. A ‘total
difficulties’ score is calculated by totalling the four
deficit focused subscales (i.e. all except for prosocial
behaviour), and an impact score is produced based on
five items, such as: ‘Do the difficulties upset or distress
your child?’ Parent and teacher-forms of the SDQ are
available for 3–16 year-olds, while a youth report form
is available for 11–16 years olds. At less than one-
quarter the length of the Child Behaviour Checklist
(CBCL) [2], and with evidence of comparable reliability/
validity [3], expedience and parsimony appear to under-
lie the SDQ’s increasing popularity. Within Australia,
interest in the SDQ has escalated following recent inclu-
sion of the measure in the Mental Health Outcomes and
Assessment Training (MH-OAT) protocol, a state-wide
data collection protocol being used with patients in New
South Wales mental health services. Data from the SDQ
and other key measures are collected on admission,
review, and discharge, and routinely reported to the
NSW Department for Health.
David J. Hawes, Research Associate; Mark R Dadds, Professor (Cor-
respondence)
School of Psychology, University of New South Wales, Sydney, New
South Wales, Australia 2052. Email: m.dadds@unsw.edu.au
Received 27 November 2003; revised 20 April 2004; accepted 21 April
2004.
Copyright
D.J. HAWES, M.R. DADDS 645
While current interest in the SDQ attests to the demand
for such brief measures, it is only with continued evalu-
ation that the properties of this relatively new instrument
may be understood. Furthermore, the appropriateness of
this measure for use in different cultures is an issue of
current relevance, with the SDQ now available world-
wide in over 40 languages. The SDQ has been evaluated
within the UK [6] as well as with samples of Dutch
(n = 562) [5], Arabic (n = 322) [6], Swedish (n = 900)
[7], Bangladeshi (n = 261) [8], German (n = 273) [9] and
Finnish (n = 703) [10] children. These studies vary con-
siderably with regard to the properties of the instrument
evaluated.
Studies of the SDQ conducted with samples in the
UK [4], Holland [5] and Sweden [7] have offered
largely consistent support for the original five-factor
structure of the SDQ in parent, teacher and youth-
report formats. Unexpected findings consistent across
at least two of these include numerous secondary load-
ings on the prosocial factor [4,7] and a limited associa-
tion between the item ‘Generally obedient . . .’ and the
‘conduct problems’ factor with which it is theoretically
associated [4,5,7]. Only one available study, conducted
with Arabic children [6], has failed to replicate the
predicted factor structure. The authors concluded that
while the original factors were somewhat evident in the
sample, the individual subscales appeared to be more
heterogenous or multifactorial than observed in other
populations.
UK data on the original five-factor structure of the
SDQ has generally suggested sound internal reliability.
Goodman [4] reported mean Cronbach’s
α
coefficients
of 0.73 across the respective scales. In this sample, the
lowest coefficient was reported for youth self-report of
peer problems (0.41), while parent and teacher-reports
across scales ranged from 0.57 (parent-report peer prob-
lems) to 0.88 (Teacher-report hyperactivity).
The internal consistency of the SDQ has also been
assessed in the Dutch [5], Swedish [7], Arab [6] and
Finnish [10] studies. Findings across these populations
have generally supported the internal reliability of the
instrument (Finnish: α= 0.63–0.86; Swedish: α= 0.51–
0.75; Dutch: α= 0.45–0.80), with the only questionable
support coming from the Arabic sample [6], in which the
authors attributed low to moderate coefficients (α = 0.18–
0.65) to the unexpected factor structure observed.
Research into the test–retest reliability of the SDQ is
limited and available findings appear mixed. Only one
UK study appears to have reported such evidence, with
a community sample of 34 parents completing the
measure 3–4 weeks following initial administration [11].
Intra-class correlations ranged from 0.44 (the ‘burden’
item from the impact scale) to 0.85 (total difficulties).
However, coefficients for five SDQ subscales were not
reported.
The study reported by Muris, Meesters, and van den
Berg [5] appears to include the only formal assessment
of test–retest reliability to come from a sample outside
the UK. Parent and youth reports on the SDQ were
obtained for 91 children from the study’s original
sample (n = 562), 2 months after initial administration.
With the exception of the procosial subscale (ICC = 0.59)
of the youth-report form, the intra-class correlations of
all subscales were higher than 0.70.
Support for the concurrent and discriminant validity
of the SDQ is available from a range of UK studies
[1,3,4,11]. Evidence of good concurrent validity against
established self-report measures is also available from
the Dutch [5], German [9] and Finnish [10] studies. In
populations outside the UK, evidence of the discriminant
validity of the SDQ subscales to date has been available
from Bangladeshi [8] and German [9] samples only. In
the latter, SDQ parent and teacher-reports on the sub-
scales of the SDQ and CBCL performed comparably in
distinguishing between community and clinic samples,
while the total difficulties scale on the SDQ discrimi-
nated more accurately than that of the CBCL. The SDQ
also demonstrated some advantage over the CBCL in
discriminating between diagnoses within the clinical
sample, with the SDQ significantly better at predicting
hyperactivity [9]. Less impressive however, were the
findings from the Bangadeshi sample, in which the total
difficulties scale and the peers problems and prosocial
subscales failed to distinguish between clinic and com-
munity groups [8]. As the SDQ was the only measure
employed in this study, the extent to which these find-
ings reflect the specific properties of the SDQ, or the
actual sample, is unclear.
In summary, while evidence to date has been largely
supportive of the reliability and validity of the SDQ in
various populations [5,7–10], mixed findings (e.g. [6,8]).
highlight the need for continued evaluation. With a
notable lack of evidence pertaining to samples of
younger children, further attention to such samples
appears particularly warranted. The aim of this study
was to assess the basic psychometric properties of the
parent-report form of the SDQ with a large community
sample of young Australian children (4–9 years). As
Goodman [4] noted, studies evaluating the psychometric
properties of the SDQ have been characterized by poten-
tially unrepresentative samples, small sample sizes, or
the absence of independent psychiatric diagnoses as
validating criteria. The current study attempts to
addresses each of these criticisms, while assessing the
internal consistency, stability, external validity, and
normative data and cut-offs for the measure.
Copyright
646 AUSTRALIAN PSYCHOMETRIC PROPERTIES OF THE SQD
Method
Sample
A sample of 1359 4–9-year-old children was recruited through
11 primary schools in the city of Brisbane, Australia’s third largest
city. The schools were chosen to represent a range of inner-city and
suburban locations of differing socioeconomic status. Family income
ranged from less than $20 000 (4%), $20–30 000 (8%), $30–70 000
(50%) to over $70 000 (38%). Education in parents ranged from
elementary school (1%) through a mode of ‘finished high school’
(25%) to university educated (20%). The majority of families were two
caregivers; 13% were sole parents. Children with complete data sets
were available as follows: 4–6-year-old boys (n = 404); 7–9-year-old
boys (n = 302); 4–6-year-old girls (n = 398); and 7–9-year-old girls
(n = 255).
Permission to conduct research was obtained from the Griffith
University Human Research Ethics Committee, Education Queens-
land, and Catholic Education Queensland. The test battery (with
attached information and consent forms) was dispersed through the
schools to all children within the relevant age ranges and sent home to
parents. Mean return rates ranged from 32.5% to 74.8% across schools.
Of the original 1359 children, 900 were randomly selected to com-
plete follow-up assessments at 12 months; 780 (86.7%) completed and
returned the measure and 450 of these were randomly selected for
telephone diagnostic interviews. Of these, 327 (72.6%) were success-
fully contacted and interviewed.
Measures
In addition to the parent-report SDQ, standard demographic details
were collected. and school teachers completed a rating form for each
participating child. This measure consists of 5-point Likert scales
assessing the child on dimensions of anxiety (shy, nervous, afraid,
inhibited), aggressive, impulsive-hyperactive, language, reading and
writing problems. Previous research has supported the ability of
teachers to accurately report on these dimensions (Strauss, Frame and
Forehand, 1987) and our previous research has supported the validity
of this specific system in the context of large school-based studies [12].
Diagnostic telephone interviews were conducted using the Diag-
nostic Interview Schedule for Children, Adolescents, and Parents
(DISCAP) [13], a semistructured interview based on DSM-IV criteria
with good reliability [14]. The DISCAP is used to assign DSM-IV
diagnoses and identify subclinical features of DSM-IV disorders, both
of which are assigned severity ratings using a six-point scale
(1 = minimal impairment in functioning/symptoms rarely problematic,
to 6 = very severe impairment in functioning/symptoms always prob-
lematic). As such, the diagnostic data collected during follow-up
assessments were both categorical (presence or absence of respective
DSM-IV diagnosis) and continuous (severity of diagnosis or sub-
clinical diagnostic features). In the current study, a severity rating scale
cut-off of 4 was used as the criteria for clinically significant problems
warranting diagnosis.
Interviews were conducted by clinical psychologists using the
Diagnostic Interview Schedule for Children, Adolescents, and
Parents (DISCAP) [13], and completed for 327 (72.9%) of this
sample. Twenty-five percent of interviews were conducted by two
interviewers, positioned on separate telephone lines and kept blind to
each other’s written notes and diagnoses, in order to check interrater
reliability of diagnoses. Interrater reliability for DISCAP interviews
was high with 100% agreement on externalizing disorders and only
one disagreement of internalizing disorders (
κ
= 0.87). The correlation
between raters’ severity ratings for primary diagnosis was r = 0.96.
Results and discussion
Statistical analyses revealed no differences in the demographic and
adjustment profiles of participants according to participation rates
achieved in each school. Analyses were also conducted to check that
there were no demographic or adjustment differences between partici-
pants selected for follow-up phases, participants who completed these
measures, and the larger pool. A series of
ANOVA
s using sample as the
independent variable and demographic measures and SDQ scores
at time 1 as dependent variables confirmed the equivalence of the
samples.
Table 1 shows coefficient alphas for each of the five SDQ subscales,
and the total difficulties and impact scales. These range from 0.59
(peer problems) to 0.80 (hyperactivity), indicating a moderate to strong
internal reliability across the subscales. These alphas are very similar
to those reported by Goodman [4], with both studies showing that for
parent-report, internal reliability was strongest for the hyperactivity
subscale, and weakest for the peer problems subscale.
Factor structure
The factor structure of the SDQ was examined using SPSS principal
components analysis with oblimin rotation, performed for girls and
boys separately. Table 2 shows the pattern matrix after a five-factor
solution was forced. The five factors produced are consistent with the
original subscales of the SDQ. For boys, the hyperactivity factor
accounted for most of the total variance (22.45%), while conduct
problems explained the least (5.09%). A different pattern was observed
in girls, with the prosocial factor accounting for the most variance
(19.71%), and peer problems the least (5.11%).
For both genders, most items loaded moderately to strongly onto
their predicted factors, with the factor loadings for boys found to be
generally stronger that those for girls. Common to both boys and girls
were cross-loadings for the item ‘Generally obedient . . .’ For boys, the
item loaded most strongly (albeit negatively) onto the prosocial factor
(– 0.38), and almost as strongly onto the hyperactivity factor (0.34).
For girls it loaded most strongly onto the prosocial factor (0.46). In
girls, cross-loadings were also observed for the peer problems item
‘Solitary, plays alone . . .’, which loaded weakly onto this predicted
factor (0.39) and almost as strongly onto the emotional symptoms
factor (0.30). The hyperactivity item ‘Thinks things out before acting’
loaded moderately onto the predicted factor (0.48) and to a lesser
extent onto the prosocial factor (0.39).
Table 3 presents correlations between the five subscales of the SDQ.
Each scale correlated significantly (p < 0.01) with every other, with
these correlations ranging from –0.14 (prosocial and emotional
symptoms) to 0.52 (conduct problems and hyperactivity). While the
observed pattern of correlations suggested mutual associations across
the five SDQ subscales, the strengths and directions of these corre-
lations are conceptually meaningful and consistent with current
Copyright
D.J. HAWES, M.R. DADDS 647
knowledge of comorbidity. For example, both the conduct problems
and hyperactivity subscales correlated most strongly with each other
(0.52), while prosocial correlated most strongly with conduct problems
(– 0.46), with the expected negative relationship.
Table 3 also presents the correlations between teacher ratings of
child behaviour and parent report on the SDQ. These reveal consistent
cross-informant agreement between subscales of common symptom
areas. Teacher-rated aggression, for example, correlated positively
with SDQ conduct problems (0.35, p < 0.01), and negatively with
prosocial (– 0.24, p < 0.01). As would be expected, the SDQ peer
problems subscales correlated positively with teacher ratings of
aggression (.15, p < 0.05), and hyperactivity (0.19, p < 0.05). With
regard to internalizing symptoms, teacher ratings of anxiety correlated
positively with the SDQ emotional symptoms (.21, p < 0.01).
Stability
SDQ scores at time 1 and time 2 were used to examine test–retest
reliability. Clearly, 12 months is too long to conduct a traditional esti-
mate of measurement stability. Over such a period, correlations will
reflect both measurement instability as well as real changes in the
child’s behaviour due to maturation, environmental changes and
the like. Thus, correlations between time 1 and time 2 scores will be at
the lower end for stability estimates. However, we deemed these worth
reporting due to the high values obtained for the current sample: hyper-
activity, r = 0.77; conduct problems, r = 0.65; emotional symptoms,
r = 0.71; peer problems, r = 0.61; prosocial, r = 0.64; total difficulties;
r = 0.77; impact scores, r = 0.63. These show that parents’ ratings on
the SDQ are remarkably stable over a 12-month period and are only
marginally lower than those reported for test–retest intervals of
1–2 months [5,11].
Validation against clinical diagnoses
DSM-IV diagnoses assigned to the interviewed sample were separ-
ated into four diagnostic groupings corresponding to the main symptom
subscales of the SDQ. These were:
conduct disorders
(conduct disorder
and/or oppositional defiant disorder);
hyperactivity
(attention deficit/
hyperactivity disorders);
internalizing disorders
(separation anxiety,
specific phobia, overanxious disorder, generalized anxiety disorder,
panic disorder, social phobia). Finally, diagnoses of disorders not
subsumed by these categories (elimination disorders, adjustment dis-
order, school refusal, nightmare disorder) formed the
other disorders
category.
Table 4 presents the prevalence of diagnoses when cases were
grouped according to high or low risk based on SDQ subscale and
total scores. Cases scoring within the most extreme 10% of each
subscale were regarded as ‘high risk’, while those scoring below the
90th percentile were regarded as low risk. The exception to this
was the prosocial subscale, on which the lowest scoring 10%
were thought most likely to exhibit psychopathology. The diagnoses
examined for each SDQ score were those most closely associated
Table 1. Mean scores and banding for SDQ subscales and total difficulties and impact scales
Subscale Coefficient
Alpha n = 1359
Banding of
Scores
4-6 years mean (SD) 7-9 years mean (SD)
Boys (n = 404) Girls (n = 398) Boys (n = 302) Girls (n = 255)
Hyperactivity 0.8 3.84 (2.65) 2.91 (2.15) 3.93 (2.72) 3.10(2.37)
Alpha 0.82 0.76 0.82 0.76
Top 5%
≥
9
≥
7
≥
9
≥
8
Top 10%
≥
8
≥
6
≥
8
≥
6
Emotional Symptoms 0.66 1.85 (1.87) 1.90 (1.75) 2.22 (2.19) 2.31 (2.19)
Alpha 0.64 0.56 0.73 0.7
Top 5%
≥
6
≥
5
≥
7
≥
7
Top 10%
≥
5
≥
4
≥
5
≥
6
Conduct problems 0.66 1.83 (1.74) 1.45 (1.46) 1.95 (1.97) 1.45 (1.62)
Alpha 0.66 0.58 0.72 0.65
Top 5%
≥
5
≥
4
≥
6
≥
4
Top 10%
≥
4
≥
3
≥
4
≥
4
Peer problems 0.59 1.52 (1.66) 1.27 (1.45) 1.45 (1.67) 1.35 (1.57)
Alpha 0.62 0.54 0.63 0.6
Top 5%
≥
5
≥
4
≥
5
≥
5
Top 10%
≥
4
≥
3
≥
4
≥
4
Prosocial 0.7 7.53 (1.88) 8.22 (1.60) 7.76 (1.85) 8.37 (1.62)
Alpha 0.7 0.71 0.65 0.67
Top 5%
≤
5
≤
5
≤
5
≤
6
Top 10%
≤
5
≤
5
≤
6
≤
6
Total difficulties 0.82 9.04 (5.52) 7.53 (4.73) 9.55 (6.27) 8.22 (5.57)
Alpha 0.81 0.85 0.78 0.82
Top 5%
≥
20
≥
16
≥
22
≥
18
Top 10%
≥
16
≥
14
≥
19
≥
16
Impact 0.79 2.09 (2.86) 1.31 (2.24) 2.77 (3.52) 2.08 (3.01)
Alpha 0.79 0.81 0.72 0.76
Top 5%
≥
8
≥
7
≥
10
≥
8
Top 10% >6 >5 >8 >7
Copyright
648 AUSTRALIAN PSYCHOMETRIC PROPERTIES OF THE SQD
with the respective subscale score. For scores on the emotional
symptoms, conduct problems, and hyperactivity subscales of the SDQ,
these diagnoses were those grouped in the three respective diagnostic
categories described earlier (conduct disorders, hyperactivity, internal-
izing disorders). As total difficulties, peer problems, prosocial, and
impact scores were thought relevant to a range of diagnoses, the
prevalence of any diagnosis was examined for each. This method
is based on Goodman’s [4] analysis of concordance between the
SDQ and clinical diagnoses, and as such was thought optimal for
comparing data from the current study with previous SDQ
research. Following Goodman [4], the discrete scores of the SDQ were
divided as closely as possibly into groups of 10% and 90% (e.g. at
Table 2. SDQ factor analysis
Total variance
explained
Hyper Emotion Prosocial Peer Conduct
Boys
19.71%
Girls
8.02%
Boys
8.15%
Girls
9.14%
Boys
7.13%
Girls
22.45%
Boys
5.28%
Girls
5.09%
Boys
5.11%
Girls
5.54%
Factor loadings
Distractible
0.84 0.78
Persistent
0.79 0.66
Restless
0.68 0.71
Fidgety
0.74 0.72
Reflective
0.61 0.49 0.4
Fears
0.73 0.76
Worries
0.77 0.72
Clingy
0.65 0.65
Unhappy
0.57 0.49
Somatic
0.49
0.39
Helps out
0.71
–0.66
Caring
0.76
–0.7
Considerate
0.58
–0.56
Kind to kids
0.59
–0.45
Shares
0.53
–0.61
Good friend
0.63 0.72
Popular
0.65 0.48
Best with adults
0.69 0.55
Solitary 0.3
0.54 0.39
Bullied
0.41 0.49
Lies
0.62 0.66
Fights
0.71 0.54
Tempers
0.54 0.53
Steals
0.63 0.7
Obedient 0.34 –0.38
0.46
Rotated (oblimin) five-factor solution; loadings between +0.3 omitted, loadings greater than +0.4 in boldface type.
Table 3. Correlations between SDQ scales and teacher ratings
SDQ Scale Total
Diff.
Emotional Conduct Hyper Peer Prosocial Impact Teacher
anxiety
Teacher
aggression
Teacher
hyperactive
Total difficulties 1
Emotional symptoms 0.67** 1
Conduct problems 0.72** 0.27** 1
Hyperactivity-inattention 0.79** 0.28** 0.52** 1
Peer problems 0.62** 0.36** 0.29** 0.27** 1
Prosocial behaviour –0.42** –0.14** – 0.46** –0.33** –0.28** 1
Impact 0.71** 0.48** 0.57** 0.53** 0.44** –0.36** 1
Teacher anxiety 0.12** 0.21** 0.09** 0.15** 1
Teacher aggression 0.32** 0.06* 0.35** 0.34** 0.15** –0.24** 0.32** – 0.06* 1
Teacher hyperactivity
–inattention 0.36** 0.32** 0.42** 0.20** – 0.24** 0.34** 0.06* 0.59** 1
*p < 0.05; **p < 0.01.
Copyright
D.J. HAWES, M.R. DADDS 649
times these were 12% and 88%, etc. due to the discrete nature of the
scores).
For each SDQ scale/subscale, there were significant differences in
prevalence between the high and low-risk groups (p < 0.05 for each),
indicating that higher scores were associated with a greater probability
of being assigned a DSM-IV diagnosis. The highest odds ratio was
observed for the conduct problems subscale (and ODD, CD diagnoses)
(30.5), while the lowest was seen for prosocial (and frequencies of any
diagnoses) (2.3).
The concurrent validity of the SDQ was also evaluated against
diagnostic interviews by correlating SDQ scores with the severity of
primary Axis I diagnostic features. Ratings of symptom severity were
collapsed into the same diagnostic variables described earlier (conduct
disorders, hyperactivity, internalizing disorders), in addition to ratings
of symptom severity for any disorder (i.e. those within these categories
as well as those in the ‘other disorders’ category). Results are shown in
Table 5.
The severity of the sample’s overall primary features as rated by
clinicians, correlated strongly with SDQ scores for impact (0.57,
p < 0.01) and total difficulties (0.47, p < 0.01). The conduct problems,
hyperactivity, and emotional symptoms scales of the SDQ correlated
strongly with clinical assessments of related diagnostic features, with
coefficients ranging from 0.33 (p < 0.01) for emotional symptoms
with internalizing disorders, to 0.51 (p < 0.01) for hyperactivity scores
with hyperactivity diagnoses. As would be expected, SDQ ratings
of peer problems correlated positively with both clinical diagnoses of
conduct disorders (0.12, p < 0.05) and internalizing disorders features
(0.14, p < 0.05) as well as the severity of any diagnostic features (0.28,
p < 0.01).
Participant treatment status was examined to further evaluate the
discriminant and predictive validity of the SDQ.
ANOVA
s revealed
that children reported by parents to be currently receiving treatment
for emotional/behavioural problems scored significantly higher on
SDQ total difficulties (mean = 15.0, SD = 6.5), than children
not receiving treatment (mean = 8.0, SD = 5.1), F
1,1398
= 171.21,
p < 0.001).
Shown in Table 1 are the means and cut-offs for the 90th and 95th
percentiles for each of the five SDQ subscales, and the total difficulties
and impact scales. From the total sample (n = 1359), these cut-offs
were produced separately for each gender, and broken into two age
groups (4–6 years and 7–9 years). Due to the positive nature of the
prosocial subscale, the cut-offs presented for this subscale represent
scores at the 10th and 5th percentiles.
A number of similarities are evident between the current findings
and those reported in Goodman’s [4] nationwide UK epidemiological
sample. For example, using the same criteria as Goodman [4], preva-
lence of diagnoses in the high-risk groups of each sample were
comparable for emotional symptoms (UK: 20.5%, Australia: 17.5%)
and ADHD (UK 17.5%; Australia: 20.8%). Also consistent with
Goodman [4], odds ratios in the current sample identified the prosocial
subscale as the SDQ subscale exhibiting the weakest association with
DSM-IV diagnoses (UK: 3.4; Australia: 2.3).
One notable differences between the current study and that reported
by Goodman [4] was the discrepancy between prevalence rates of
ODD and CD in the conduct problems high risk groups within each
sample (UK: 25.7%; Australia: 9.8%). This difference however, is
better explained by differences in the overall prevalence rates of CD
and ODD diagnoses in the respective samples (UK: 4.7%; Australia:
1.5%) than differences in the distribution of diagnoses between risk
groups.
Conclusion
The aim of this study was to assess the basic psycho-
metric properties of the parent-report SDQ with a large
community sample of young Australian children (aged
4–9 years). Moderate to strong internal reliability and
stability was exhibited across all SDQ subscales. Ade-
quate validity was evidenced in the relationship of
these scales to one another, while the pattern of correla-
tions between the SDQ subscales, teacher-ratings and
Table 4. Prevalence of DSM-IV diagnoses within high (extreme 10% of sample) and low risk (90% of sample)
groups based on SDQ scores
SDQ scale
(Cutoff)
Prevalence of Diagnosis
DSM-IV low-risk high-risk odds ratio
diagnosis group group (95% CI)
Total Difficulties Any 4.6% 36.1% 11.7
(15/16) (13/283) (13/36) (4.9–28.2)
Emotional Internalizing 1.75% 17.5% 11.9
(4/5) Disorder (5/285) (7/40) (3.6–39.6)
Conduct Conduct Disorder 0.4% 9.8% 30.5
(3/4) (1/283) (4/41) (3.3–280.1)
Hyperactivity Any ADHD 1.4% 20.8% 17.9
(6/7) (4/276) (10/48) (5.3–59.9)
Peer Problems Any 5.9% 26.5% 5.7
(3/4) (17/287) (9/34) (2.3–14.1)
Prosocial Any 7.5% 15.6% 2.3
(5/6) (22/293) (5/32) (0.8–6.5)
Total Impact Any 4.5% 41.2% 14.9
(6/7) (13/291) (14/34) (6.2–36.1)
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650 AUSTRALIAN PSYCHOMETRIC PROPERTIES OF THE SQD
diagnostic interviews demonstrated sound external
validity. This was demonstrated further by the relation-
ship of SDQ total difficulties scores to concurrent treat-
ment status.
While the original five-factor structure of the SDQ
was generally confirmed, one conduct problems item
(‘Generally obedient . . .’) loaded more strongly onto the
prosocial subscale for both boys and girls. While this
finding may seem conceptually confusing, it is consist-
ent with previous research. Goodman [4], for example,
reported factor loadings on the prosocial subscale for
this item (among others), prompting the suggestion that
the prosocial subscale might be described as a ‘positives’
factor. Studies outside the UK have also noted unex-
pected factor loadings for this item [5,7]. The findings
from the current study can therefore be seen as adding to
existing evidence, questioning the utility of this item as
an indicator of conduct problems in young children.
While not yet evaluated, it is possible that this item
would demonstrate a more unique relationship with
conduct problems if it were negatively worded (e.g.
‘Generally disobedient, usually refuses/ignores what
adults request’).
The following limitations should be noted for the
current study. While teacher ratings of child behaviour
and difficulties were obtained, the teacher version of the
SDQ was not used for this purpose. Inclusion of this
would have allowed for a comparison of cross-informant
reports on the instrument. The present study nonetheless
supports the utility of the parent report SDQ as a
measure of psychopathology in young Australian chil-
dren. As the properties of the teacher and youth-report
SDQs remain unknown in Australian samples, further
research into these forms of the instrument would repre-
sent an important contribution to the growing knowledge
of the measure.
Acknowledgement
This research was supported by the National Medical
Health and Research Council of Australia and Catholic
Education of Queensland. Thanks to all the families who
participated.
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Table 5. Correlations between SDQ scales and severity of diagnostic features
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