/ , Child Psychoi. Psychiat. Vol. 38. No, 7. pp. 867-872, 1997Cambridge University Press

© 1997 Association for Child Psychology and PsychiatryPrinted in Great Britain, All rights reserved

0021-9630/97 $15,00-1-0,00

Parental Divorce and Adult Psychological Distress: Evidence from aNational Birth Cohort: A Research Note

Bryan RodgersThe Australian National University, Canberra, Australia

Chris Power and Steven HopeInstitute of Child Health, London, U.K.

An association was found between childhood parental divorce and adult psychologicaldistress in a British national birth cohort at ages 23 and 33. No moderating effects were foundfor gender, age at separation, or remarriage of the custodial parent. Participants who wereyoung adults when their parents divorced also showed increased levels of symptomatology,whereas those who experienced parental death in childhood showed no increased risk. Aninteraction between parental divorce and own divorce in women, giving particularly highsymptom levels, arose from a selection process in those from divorced families of origin only,with high 23-year scores predicting subsequent divorce. Own divorce was associated with anincrease in distress between age 23 and 33, but this was irrespective of family of origin.

Keywords: Depression, adulthood, divorce, family structure, psychological distress, longi-tudinal study.

Introduction

An association between childhood parental divorce(where the term "divorce" is used to denote permanentseparation as well as divorce) and adult psychologicaldistress has been demonstrated repeatedly. This has beenreported in many countries, from case-control and cohortstudies, and effect sizes are larger than for most otherdomains of adult well-being (Amato & Keith, 1991;Parker, 1992; Tennant, 1988). For such a long-estab-lished association, remarkably little is known aboutcontributory mechanisms. Few studies have investigatedfactors that mediate or moderate the risk arising fromparental divorce (Aseltine, 1996; McLeod, 1991). Onestudy of a British cohort born in 1946, the NationalSurvey of Health and Development (NSHD), reportedon the pathways between parental divorce and adultdepression (Rodgers, 1994). Findings from a follow-up atage 43 included :

(1) the risk of adult depression associated with par-ental divorce was greater in women than men;

(2) there was no effect of remarriage of custodialparents on their offspring's mental health;

(3) parental death in childhood was not associatedwith later depression;

(4) women who were young adults when their parentsseparated also showed increased risk of depression;

Requests for reprints to: Dr Bryan Rodgers, NHMRC Psy-chiatric Epidemiology Research Centre, The AustralianNational University, Canberra, ACT 0200, Australia.

(5) the increased risk of depression in women whoexperienced childhood parental divorce was re-stricted to those who were never married, divorced,or remarried at age 43.

Replication of these results, especially with a largersample, is important for the first four findings. The fifthfinding was ambiguous, being consistent either with avulnerability model whereby parental divorce sensitisedindividuals to a more detrimental impact of breakdownin intimate relationships in adult life, or with selectionmodels, whereby those with depressive symptoms, ortraits predisposing to depression, were at greater risk ofdivorce themselves. Longitudinal analyses, with adultsymptoms and marital status assessed on more than oneoccasion, would have clearer implications.

The British National Child Development Study(NCDS) met the criteria of having a larger sample andappropriate longitudinal data. Analyses using these datawere conducted with the following aims:

(1) to quantify the risk of adult psychological distressassociated with childhood parental divorce;

(2) to identify moderating effects of gender, age atseparation, and remarriage of custodial parent onthis risk;

(3) to evaluate the importance of loss and lack ofparent figures during childhood by separatelycontrasting (i) those who experienced death of aparent up to age 16 and (ii) those who were youngadults when their parents divorced, withparticipants from intact families and with thosewho experienced parental divorce in their child-hood years;

867

868 B. RODGERS et al.

(4) to investigate the interaction of parental divorceand own divorce in predicting psychological dis-tress.

MethodSample, Data Collection, and Derived Variables

The NCDS is a prospective study of individuals born in theweek 3-9 March 1958 (Butler & Bonham, 1963). All those bornin Wales, Scotland, and England, plus immigrant childrenresiding in Britain by age 16, were included in the cohort. Datawere gathered at birth and at ages 7, 11, 16, 23, and 33. At age23, 12537 participants were interviewed, representing 76% ofthe target sample and 11,407 (69 %) were interviewed at age 33.The adult samples showed little bias compared to the originalcohort, but disadvantaged groups were slightly under-represented (Ferri, 1993; Power, Manor, & Fox, 1991).Although there were no direct questions on divorce in thechildhood interviews, items covered relationships of parent-figures to the participant and reasons for absence of naturalparents from the household (Chase-Lansdale, Cherlin, &Kiernan, 1995; Cherlin et al., 1991; Elliott & Richards, 1991).At age 33, participants were asked whether their parents hadever permanently separated or divorced, their age when theirparents last lived together, and whether they were living withtheir parents immediately before the separation. Discrepanciesbetween childhood and adult sources were examined. Partici-pants who had never lived with both natural parents wereexcluded, although it was noted that a proportion did report atage 33 that their parents had divorced. Of the further 1046participants who reported retrospectively that their parentsseparated (by age 16), 435 were not identified from childhoodinformation. Most of these (304) had some missing data. The33-year reports were used to determine age when parentsstopped living together in all cases. However, 97 participantswith a record of parental divorce during childhood did notreport this at age 33. For 52, the discrepancy concerned age atseparation rather than its occurrence. Information for theremaining 45 participants suggested that many had not livedwith both natural parents until age 16, and the prospective datawere considered more reliable. By grouping age when parentsstopped living together into categories corresponding to thetiming of childhood interviews (i.e. 0-7, 8-11, 12-16 years), itwas possible to use earlier information to correct later reports.The 1143 participants who were 16 or younger when theirparents divorced represented 10.1 % of those followed up. Afurther 541 respondents (4.8%) had parents who divorced atlater ages and 505 (4.5%) had a parent die before the 16-yearinterview.

The self-completed Malaise Inventory (Rutter, Tizard, &Whitmore, 1970) provided a measure of psychological distressat ages 23 and 33 years. It was scored by summing endorseditems (range 0-24). Scores were imputed for individuals withone or two missing items, estimated from the relevant percentilepoint on the population distribution for completed items.Participants with three or more missing items (0.4%) wereallocated missing values. A single dimension was found to be anappropriate representation of item scores and yielded accept-able internal consistency (Cronbach's a = .77 and .80 at age23 and 33 respectively). The scale was highly discriminatingbetween those who had or had not sought professional help formental health problems, as indicated by receiver operatingcharacteristic analysis (area under the curve = 0.73 for cons-ultations between ages 23 and 33), and showed relationshipswith demographic and socioeconomic variables similar to thoseexpected for depressive disorders of clinical severity.

Marital status at ages 23 and 33 was classified into categoriesof single (i.e. never married), married, divorced, and remarried.

The married and remarried included those who reported theywere living as married. The widowed were excluded.

Statistical AnalysisThe GLM procedure of SAS (SAS Institute, 1990) and the

MANOVA procedure of SPSS (SPSS, 1990) were used foranalyses, including repeated measures analysis of variance,where Malaise scores at ages 23 and 33 were the continuousdependent variables. When more than one pairwise comparisonwas conducted, a levels were adjusted accordingly (i.e. a = .025for two comparisons). Tests for main effects and interactions,including those involving age at Malaise completion, wereconducted. The LOGISTIC procedure of SAS was used foranalysis of dichotomised Malaise scores to determine oddsratios.

Results

Response Bias

Response bias in relation to family of origin could notbe determined precisely, because parental divorce hadbeen identified in part by retrospective reports. However,estimates indicated that children from divorced familieswere less likely than those from intact famihes to beinterviewed in adulthood. Possible implications for dif-ferences in adult psychological distress were investigatedby comparing childhood behaviour ratings from theRutter Teacher scale (Rutter, 1967), obtained at age 16,for those included and omitted at age 33. This scale is animportant predictor of Malaise score at 23 years (Poweret al., 1991). Behaviour scores were significantly higherfor those missing at follow-up compared to those con-tacted, in offspring from both divorced and intactfamilies. The size of these differences and the extent ofdrop-out indicated that adult data would tend to under-estimate differences in Malaise scores between those fromdivorced and intact families. A similar tendency operatedwhen analyses were restricted to individuals with Malaisescores from both adult contacts. Other analyses indicatedno significant differences in Malaise scores between thoseinstances of parental divorce detected from the childhoodinterviews and those not. It seemed reasonable, therefore,to include all participants in subsequent analyses, ir-respective of the source of information on parentaldivorce.

Risk of Adult Psychological Distress Associatedwith Parental Divorce

Adult psychological distress was greater in those whoexperienced parental divorce by age 16 compared to thosefrom intact families (Table 1). There was no significantinteraction with age at Malaise completion, and theinteraction of parental divorce with gender was alsononsignificant. The strength of association was assessedfor 23- and 33-year scores in the form of continuous anddichotomised measures, using a cut-point between 5 and6 in the latter case. The effect size indicated by continuousmeasures was 0.30 at age 23 and 0.33 at age 33, adjustedfor gender, and odds ratios for the dichotomised scoreswere 1.70 and 1.85 respectively.

The possibility that effects associated with parentaldivorce could reflect lower childhood socioeconomic

PARENTAL DIVORCE AND ADULT PSYCHOLOGICAL DISTRESS 869

Table 1Mean 23- and 33-year Malaise Scores by Type of Separation

No separation**

Divorce (0-16 yrs)" '

Death (0-16 yrs)"

Divorce (17-33 yrs)**

Age 23

1.88(3818)

2.67(390)2.22(205)2.21(202)

Men

Age 33

1.85(3818)

2.89(390)2.17(205)2.42(202)

Age 23

3.19(3947)

4.11(480)3.29(221)3.59(243)

Women

Age 33

2.61(3947)

3.45(480)2.87(221)3.42(243)

Tests of family of origin effects, adjusted for gender (MANOVA):a vs. b: F(2,8631) = 50.9, / ;<.001;a vs. c: F(2,8187) = 2.2, p = .100*;b vs. c: F(2,1292) = 7.2,;; < .001*;avs. d:F(2,8206)= 13.2,/?< .001*;b vs. d: F(2,1311) = 3.7, p = .026*.

A shown in brackets; root MSE = 2.79 at age 23 and 2.87 at age 33; * adjusted a = .025.

status (SFS) was addressed using housing tenure as aproxy measure. This avoided the problem of missing datafor occupational status when there was no father figure inthe household. Housing tenure in childhood is predictiveof adult Malaise scores in this sample (Power & Manor,1992; Power et al., 1991). Parental divorce remainedhighly significant in the model after adjustment fortenure. The effect sizes for 23-year and 33-year Malaisewere reduced to 0.24 and 0.31 respectively.

Age at Separation and Remarriage of CustodialParents

MANOVA for the groups who were 0-7, 8-11, and12-16 years old when their parents stopped livingtogether showed no significant effect of age at separationand no interactions with either age at Malaise completionor gender. Parental remarriage was not ascertainedretrospectively, so this factor could only be consideredfor those identified from contemporary data on familystructure. Remarriage of the custodial parent was foundto be of no significance for adult Malaise scores and wasnot considered further.

Table 2Mean 33-year Malaise Score by Current Marital Statusand Family of Origin

MenNo separation(N)Divorced(N)

WomenNo separation{N)Divorced(TV)

Single

2.37(653)3.96(54)

2.79(429)3.67(48)

Married

1.72(3033)

2.55(322)

2.45(3074)

2.99(354)

Divorced

2.65(191)3.74(51)

3.74(317)5.56(62)

Remarried

2.04(398)3.08(77)

3.08(541)3.98(113)

ANOVASMen:

Marital status, F(3,4771) = 13.4,/? < .001.Family of origin, i^(l,4771) = 35.9, p< .001.Interaction term, F(3,411\) =].4,p = .232.

Women:Marital status, F(3,4930) = 26.0, /? < .001.Family of origin, F(l ,4930) = 31.9,/? < .001.Interaction term, F(3,4930) = 2.1, p = .047.

Parental Death, Parental Divorce in EarlyAdulthood, and Adult Distress

Differences in mean Malaise scores between those whoexperienced parental death by age 16 and those fromintact families were nonsignificant (Table 1), and nointeraction terms involving parental death weresignificant. The parental death group also hadsignificantly lower scores than those whose parentsdivorced during their childhood years.

Those who were aged 17 or older when their parentsseparated had significantly higher scores than those fromintact families (Table 1). There was also a separation byage at completion effect in repeated measures analysis[F(2,8207) = 6.8, p = .00% reflecting the greater dif-ference on 33-year score by comparison with 23-yearscore, although separate ANOVAS for 23-year and 33-year scores both showed significant effects. Comparison

between those for whom parental divorce occurred up toage 16 and those where divorce was later showed adifference of borderline significance, with the latter grouphaving lower mean Malaise scores. Additional analysesexamined whether psychological distress in those aged17-23 years at separation was more pronounced for thegroup still living with their parents immediately beforethe separation, but this was not confirmed.

Parental Divorce., Own Divorce, and Adult Distress

As significant interaction terms were found for genderand own marital status in the prediction of Malaisescores, separate analyses were conducted for men andwomen. For men, the interaction between parentaldivorce and own marital status was nonsignificant in theprediction of the 33-year Malaise score (Table 2). For

870 B. RODGERS et al.

Table 3Mean 23-year Malaise Score by Subsequent Change in Marital Status and Family ofOrigin: Women

Family of origin Married-married Married-divorced Married-remarried

No separation

Divorced(TV)

3.21(1805)

3.64(219)

3.47(180)5.64(36)

3.22(269)4.39(57)

ANOVA: Change in marital status, F(2,2560) = 1.5,p= .001.Family of origin, F(l, 2560) = 25.4, p < .001.Interaction term, F(2,2560) = 4.7, p = .009.

women, this interaction just reached signiflcance and thepattern of mean scores was similar to that predicted.Parallel analysis using a logarithmic transformation ofMalaise score gave a similar pattern of results, but yieldeda nonsignificant value for the interaction term. Thisraised the possibility that the interaction could resultfrom problems of scaling with skewed dependent variables(Rodgers, 1991). Subsequent analyses of Malaise scoresin relation to changes in marital status were conductedusing repeated measures ANOVA, where time effects arebased on difference scores. These difference scores are notprecisely normally distributed, being leptokurtic, buthave a more symmetrical distribution without theskewness shown by raw Malaise scores.

Parental Divorce, Change in Own Marital Status,and Adult Distress

Cross-classiflcation of marital status at ages 23 and 33showed that numbers were sufiEicient to make comparisons(1) between those remaining single and those movingfrom single to first marriages, and also (2) between thoseremaining in first marriages and those who becamedivorced or remarried.

For men, those who remained single had significantlyhigher Malaise scores (at both ages) than those whobecame married, but this was so irrespective of parentaldivorce. There was no evidence of a differential impact ofchange in status on change in Malaise score whencomparing men from divorced and intact famihes oforigin. Comparisons between those who remained mar-ried over the 10-year interval with those who becamedivorced or remarried showed that men who moved intothe divorced category had rather higher Malaise scores(at both ages) than the other groups, but this fell justshort of statistical signiflcance. There were no significanteffects for change in Malaise score associated with owndivorce or parental divorce.

For women, comparisons between those who remainedsingle and those who became married showed the samepattern to that in men. Those remaining single had higherMalaise scores at both ages, irrespective of family oforigin. Changes in Malaise scores were unrelated tochange in status, to parental divorce, or to the interactionof these factors. For women who had been married at age23, Malaise scores were signiflcantly higher (at both ages)in those who subsequently divorced. This was attributableto the women from divorced families of origin, and therewas a signiflcant interaction between change in status andparental divorce (Table 3). Change scores also showed a

marked increase for those who became divorced by age 33[F(2,2560) = 10.2, p < .001], but this occurred irrespec-tive of family of origin. There were no signiflcant changesover time associated with parental divorce.

Discussion

This study conflrmed an association between childhoodparental divorce and adult symptomatology. The strengthof association was diminished somewhat af ter adjustmentfor childhood SES, more so for the 23-year outcome, butremained highly signiflcant. The effect size was consistentwith previous studies (Amato & Keith, 1991), indicatedby odds ratios of 1.70 and 1.85 for high Malaise scores atage 23 and 33 respectively. These are probably under-estimates, given the evidence of response bias in the adultfollow-ups. Other findings showed some similarities andsome differences by comparison with the earlier investiga-tion using the NSHD (Rodgers, 1994). First, a mod-erating effect of gender, speciflcally that parental divorceis a greater risk factor in women, was not conflrmed in thepresent study. There are other studies where genderdifferences have not been found, but no systematicdifference between these studies and those reporting amoderating effect has been suggested. Second, the lack ofmoderating effects of age at parental separation andremarriage of the custodial parent agreed with the NSHDflndings. Other studies of outcomes in childhood, ado-lescence, and adulthood have provided no clear-cutevidence for the importance of these factors (Amato,1993; Amato & Booth, 1991; Aseltine, 1996). Third, theweak association of parental death in childhood withlater symptoms was consistent with reports from theearlier study (Rodgers, 1990, 1994). It is now generallyaccepted that parental death is not an important riskfactor for adult affective disorders (Parker, 1992;Tennant, 1988), but perhaps assumes grater importancein conjunction with other f"actors (Amato, 1991; Harris,Brown, & Bifulco, 1986). Fourth, the increased risk ofdistress in those who were young adults when theirparents divorced replicates previous results.

Collectively, the flndings for age at separation, re-marriage of custodial parent, and loss of parent by deathprovide further evidence that loss and lack of a parentflgure are not sufificient to account for the elevated risk ofadult psychological distress. This concurs with studies ofchildhood sequelae of divorce and of adults who haveexperienced other forms of parental separation, sug-gesting that quality of upbringing (including familyconflict, hostility, and neglect) is more important (Amato,

PARENTAL DIVORCE AND ADULT PSYCHOLOGICAL DISTRESS 171

1993; Emery, 1982; Harris etal., 1986 ;Pryor& Seymour,1996; Rutter, 1971). One study of young adults fromdivorced families provides some direct support for thishypothesis (Black & Pedro-Carroll, 1993).

The present study also identifled a stronger associationof parental divorce in early adulthood with symptoma-tology at age 33 compared to age 23, suggesting that theremay be an additional response to parental divorce (or itsconsequences) as a contemporary stressor.

Analyses of Malaise scores at ages 23 and 33 in relationto the timing of transitions in own marital careers wereinconsistent with the hypothesis that parental divorce (orevents surrounding it) establishes a vulnerable set that isinvoked in the event of termination of intimate relation-ships in later life (O'Neil, Lancee, & Freeman, 1987;Rodgers, 1994). The interaction between parental divorceand own divorce in women arose through a selectionprocess seen only in those from divorced famihes oforigin, whereby psychological distress was predictive ofsubsequent breakdown in their own marriage. Thegeneral issue of a hnk between adult marital breakdownand psychopathology has attracted considerable researcheffort, but the question of which comes flrst has not beenclearly resolved (Bachrach, 1975; Bloom, Asher, &White, 1978). Indications that psychological distressprecedes divorce may be a consequence of increasingstress during the later phases of marital breakdown(Menaghan, 1985; Merikangas, 1984). There is lessequivocal evidence that distress increases followingdivorce (Aseltine & Kessler, 1993; Horwitz, White, &Howell-White, 1996; Menaghan & Lieberman, 1986),and this was indeed found for women in the presentstudy. However, the increase occurred irrespective of pastparental divorce.

The selection factor for women from divorced familiesof origin, where high Malaise score predicted owndivorce, cannot be readily explained by the stress ofimpending marital breakdown, as no comparable effectwas observed in women from intact famihes of origin andthe length of the follow-up period (10 years) wouldminimise the impact of short-term changes. Severalprocesses might have contributed to this pattern of resultsand three general models, not necessarily mutuallyexclusive, are offered.

(1) Psychological distress may have a stronger causalimpact on marital breakdown in women fromdivorced families of origin, compared with thosefrom intact families. This could arise becauserelationships of the former have inherent weak-nesses, making them less able to cope with suchstrain, or because these women are more acceptingof divorce as a way out of an unhappy marriage(Amato, 1996).

(2) Parental divorce may sensitise women to the impactof problems in intimate relationships, rather thantheir dissolution per se. This is a modiflcation ofthe original vulnerability hypothesis. Greaterfamiliarity with the likelihood and consequencesof divorce could be factors involved in promotingsuch sensitivity.

(3) The nature of breakdown in adult intimaterelationships may be different in women from

divorced families of origin, with the factors in-volved being associated with chronic distress.Support for this comes from several studies. Aninvestigation of young adults in Finland reportedincreased levels of conflict in intimate relationshipsfor women, but not men, from divorced families oforigin (Aro & Palosaari, 1991). In one U.S. study,parental divorce was found to increase "the Uke-lihood that offspring exhibit behaviours that inter-fere with the maintenance of mutually rewardingintimate relations" (Amato, 1996). Further,McLeod (1991) reported, from a sample of marriedcouples in Detroit, that measures of marital qualityaccounted for the high rates of depression inwomen whose parents had divorced.

Whatever the mechanisms involved, the continuationof the selection process observed between ages 23 and 33would lead to a stronger pattern of interaction betweenparental divorce and own divorce at later ages. This wasindeed the case in the NSHD sample, where no as-sociation was found between family of origin anddepressive symptoms for women who were still in theirflrst marriage by age 43 (Rodgers, 1994).

Conclusions

This study provides evidence that the risk of adultpsychological distress associated with parental divorcearises from factors additional to loss or lack of a parentflgure during childhood. An interaction between parentaldivorce and own marital status, found for women, cannotbe explained by a simple model of vulnerability to theconsequences of their own divorce, but arises from aselection process, seen only in those from divorcedfamilies of origin, whereby psychological distress ispredictive of marital breakdown. It is possible thatpsychological distress has different consequences formarital breakdown in women from divorced and intactfamilies of origin. However, these flndings are alsoconsistent with mechanisms where interpersonal behav-iour in intimate relationships is implicated in the linkbetween parental divorce and adult psychological dis-tress.

Acknowledgements—The authors received financial supportfrom the National Health and Medical Research Council ofAustralia (Bryan Rodgers), the Department of Health forEngland and Wales (Chris Power and Steven Hope), and theCanadian Institute of Advanced Research (Chris Power). Weare grateful for helpful comments on a draft of this paper fromJan Pryor of the University of Auckland.

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Accepted manuscript received 6 January 1997