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The Intergenerational Transmission of Divorce: A Fifteen-Country Study with the Fertility and Family Survey

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  • University of Leipzig and ETH Zurich

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Studies mainly from the United States provide evidence that children of divorced parents face a higher risk of divorce in their own marriages. We estimate and analyze the effects of divorce transmission using comparative individual data from the United Nations for 13 eastern and western European countries as well as for Canada and the United States. We find substantial and highly statistically significant transmission effects in all samples. This shows that the intergenerational transmission of divorce is a widespread phenomenon observed without a single exception in our data covering a large number of countries with differing historical, institutional, and cultural contexts.
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© Koninklijke Brill NV, Leiden, 2013 DOI: ./-
Comparative Sociology 12 (2013) 1–14 brill.com/coso
COMPARATIVE
SOCIOLOGY
The Intergenerational Transmission of Divorce:
A Fifteen-Country Study with the
Fertility and Family Survey*
Andreas Diekmannand Kurt Schmidheiny
 Swiss Federal Institute of Technology (ETH)
diekmann@soz.gess.ethz.ch
University of Basel
kurt.schmidheiny@unibas.ch
Abstract
Studies mainly from the United States provide evidence that children of divorced
parents face a higher risk of divorce in their own marriages. We estimate and ana-
lyze the efects of divorce transmission using comparative individual data from the
United Nations for 13 eastern and western European countries as well as for Can-
ada and the United States. We nd substantial and highly statistically signicant
transmission efects in all samples. This shows that the intergenerational transmis-
sion of divorce is a widespread phenomenon observed without a single exception
in our data covering a large number of countries with difering historical, institu-
tional, and cultural contexts.
*The authors wish to thank the Advisory Group of the FFS program for its permission,
granted under identication number 81, to use the FFS data on which this study is based. We
are also very much indebted to Elisabeth Coutts and Hartmut Esser for valuable comments.
We started the analysis of the data of the Fertility and Family Survey (FFS) in 2001 when
we received approval by the U.N. Population Activities Unit to use the Fertility and Family
Survey data. Our ndings concerning a) the transmission efects in various countries and b)
a strong and signicant correlation of divorce rates and transmission efects were presented
on conferences (e.g. in Florence 2002) and in a working paper (2004). We are grateful to
Hartmut Esser who objected a causal interpretation of b) arguing that this relation might be
an artifact. Theoretical reasoning and robustness tests by the authors (see below) supported
this view. Dronkers and Härkönen (2008) replicated our ndings on country efects. Also,
they reported the seemingly divorce rate efect on the transmission efect using a multi-
level model. We believe this being an artefact having no causal meaning. For a critique of
Dronken and Härkönen (2008) see our comment Diekmann and Schmidheiny (2008).
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2 A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14
Keywords
divorce, divorce risk, intergenerational transmission, consequences of divorce,
child well-being
1 Introduction
Studies from the United States provide evidence that children of divorced
parents face a higher risk of divorce in their own marriages, although there
is controversy about the reasons for this efect (Amato 1996; Bumpass and
Sweet 1972; Glenn and Kramer 1987; Greenberg and Nay 1982; Keith and
Finlay 1988; McLanahan and Bumpass 1988; Mott and Moore 1979; Mueller
and Pope 1977; Pope and Mueller 1976; Teachman 1982; Wolnger 1999).
The efect has also been found in the Netherlands (Traag, Dronkers and
Vallet 2000), West Gemany (Diekmann and Engelhardt 1999), Great Britain
(Kiernan and Cherlin 1999) and France (Traag, Dronkers and Vallet 2000),
although no efect was found in East Germany (Diefenbach 1997). The
intergenerational transmission of divorce has contributed to the upward
trend in divorce rates, and a better understanding of its efect is crucial
not only to the analysis and prediction of divorce trends but also to insight
into the development of children after their parents’ divorce. Up to now it
has not been established whether the transmission efect and its proposed
explanations are generalizable to other Western countries and beyond. In
this study, we estimate and analyze the efects of divorce transmission for
13 eastern and western European countries as well as for Canada and the
United States. Our study is the rst to conduct a systematic investigation
covering a large number of countries with difering historical, institutional,
and cultural contexts, thus allowing us to answer the question of whether
divorce transmission is a stable and robust phenomenon observable across
diferent cultures. Also, the analysis of patterns in divorce-related charac-
teristics among children from divorced families sheds light on the explana-
tion of the transmission efect.
It is by no means natural to assume that transmission efects are a univer-
sal phenomenon. One can hypothesize that these efects depend on many
factors, some of them related to institutional setting (Amato and Keith
1991a,b). Custody regulations, nancial support by the non-resident parent,
and other aspects of divorce laws may have an impact on how well children
are able to cope with their parents’ divorce. Of course, these regulations
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A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14 3
vary over time and across countries. The nations in our sample also difer
in terms of cultural patterns, religious beliefs and the degree to which a
divorce is considered a “normal” event in the life course.
In the current study, we therefore examine the divorce transmission
efect in countries that vary in terms of the above factors. We use the ret-
rospective data on family histories from the Fertility and Family Survey
to estimate transmission efects for fteen countries, including various
Western and Eastern European countries, Canada, and the United States.
We also control for additional independent variables that may well
afect the survey’s respondent’s risk of divorce. The set of control vari-
ables includes well-known divorce risk factors such as age at the start of
a union that led to marriage, birth of a child, the wife’s educational level
premarital cohabitation and membership in certain marriage cohorts
(e.g. White 1990). Previous studies have shown that divorce risks decrease
with the age at which the union that leads to marriage is commenced,
while childless couples and spouses who lived together before marriage
exhibit higher divorce risks than couples who have children and who did
not share a household before marriage. This set of control variables may
also mediate the transmission efect. Empirical studies show that if par-
ents divorce, their ofspring complete less education, marry earlier, have
a greater tendency to cohabit before marriage and may invest less in a
partnership than children from non-divorced families (e.g. Keith and Fin-
lay 1988, Amato 1996, Diekmann and Engelhardt 1999). A further strength
of the present study is that our data allow us to investigate whether the
means of possible mediating variables difer systematically across coun-
tries. Finally, we have also accounted for marriage cohort membership
because divorce risk in general has increased in Western countries in
recent decades.
Our estimation employs the techniques known as survival analysis in
statistics, duration modelling in econometrics, or event-history analysis
in sociology. First, we investigate the presence or absence of transmission
efects in the countries included in our study. Second, we examine whether
the ndings are explainable by the other divorce-related covariates men-
tioned above. Finally, we explore whether there are systematic diferences
in the means of divorce risk factors for respondents with divorced and non-
divorced parents, diferences which could explain at least part of the social
inheritance of divorce.
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4 A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14
2 Data
The study is based on data from the Fertility and Family Survey. The FFS
comprises surveys in 21 countries, but the necessary information on either
the duration of the respondent’s marriage or whether the respondent’s par-
ents divorced is lacking for ve of those countries. Thus, with West and East
Germany analyzed separately, our estimates are based on 16 data sets col-
lected in the early 1990s in 13 European countries, Canada, and the United
States. We conne our analysis to female respondents who were married
or have previously married. With these restrictions, net sample sizes vary
from 1,279 (Czech Republic) to 6,844 (U.S.).
Table A.1 in the Appendix displays the variables used and their means.
Tables A.2 and A.3 show the means of the variables separately for respon-
dents whose parents did not divorce and those whose parents divorced.
The variable of main interest is duration of rst marriage in months. We
consider a marriage terminated when it ends in divorce or permanent
separation. For these purposes, we consider the termination date to be
the date of dissolution of a common household.
The main explanatory variable is the parents’ relationship during the
respondent’s childhood. The dummy variable parents’ divorce is set to 1 if
the respondent’s (natural or adoptive) parents divorced or separated after
her birth. In addition to the divorce/separation of the respondent’s parents,
the analysis includes the family structure of the home of origin, specically
information on whether the respondent grew up with both parents, one
parent or without either parent.
Further independent variables include the respondent’s marriage
cohort, age at start of the union that led to her rst marriage, birth of a
child, her educational level and her cohabitation history. We use ve-year
marriage cohorts from 1970 to 1990. The age at start of union is the age of the
Samples were drawn from the population within certain age limits. The Belgian sample
covers only Flanders and the region of Brussels. For more information on the FFS and its use
in comparative research, see Festy and Prioux (2002).
The FFS Standard Recode File does not distinguish between legal divorce and separation.
This denition seems reasonable as the time between the end of co-residence and the
date of the legal divorce varies substantially across the diferent jurisdictions. Furthermore,
the date of legal divorce is not reported in most FFS data sets. See Festy and Prioux (2002,
p. 32) for a discussion of the comparability of FFS partnership data across countries.
The information on composition of the home of origin varies across countries.
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A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14 5
respondent at the time she began living with her rst marriage partner. The
birth of the respondent’s rst child is included as a time-dependent covari-
ate. The respondent’s educational level is the one attained by the date of the
interview, and is measured in accordance with the international standard
classication of education (ISCED). This scale covers seven educational
levels from pre-primary (0) to the second stage of tertiary (6). We distin-
guished among three levels: ‘lower’ (values 0, 1 or 2 of the ISCED classica-
tion), ‘medium’ (ISCED 3 or 4), and ‘higher’ (ISCED 5 or 6). Cohabitation
denotes whether the respondent had already shared a household with
her rst spouse before they married (see Appendix A1 for the means of
the covariates).
3 Methods
We use a proportional hazard rate model to estimate the efects of a respon-
dent’s parents’ divorce and sociodemographic covariates on the respon-
dent’s own divorce risk.
It is well known that divorce risk increases during the initial years of
marriage and decreases thereafter (see Figure 1). Because of this non-mono-
tonic duration dependence, we model the hazard rate of divorce risk as
r(t) = atet / λ (1)
with marriage duration t, parameter λ measuring time in months elapsed
until maximal risk and a = exp(β0 + x1β1 + ... + xkβk + ... xKβk); x1 is a dichot-
omous variable indicating whether parents remained married (x1=0) or
were divorced (x1=1), x2,..., xm are further covariates, and β1, β2,..., βm
are empirically estimated parameters. βk · 100% is approximately and
[exp(k β)1] · 100% is exactly equal to the percentage change in the respon-
dent’s divorce risk r(t) when the covariate xk increases by one unit. We use
Unfortunately, educational attainment at marriage is either not reported or very poorly
reported for most countries. See Festy and Prioux (2002, p. 32) for a discussion of the limited
comparability of education variables across countries in the FFS.
The other levels of the ISCED scale are: (1) primary education or rst stage of basic educa-
tion, (2) lower secondary or second stage of basic education, (3) (upper) secondary educa-
tion, (4) post-secondary non-tertiary education, and (5) rst stage of tertiary education. See
UNESCO (1997) for more details.
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6 A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14
maximum likelihood to estimate the β parameters of covariate efects and
the parameter λ. Apart from the birth of the rst child, all independent
variables are treated as time constant. We estimate the parameters of the
time-dependent covariate in the likelihood function using the method of
episode splitting (Blossfeld and Rohwer 1995).
The complete length of the episode can be observed only in marriages
that ended in divorce before the interview took place. Marriages still in
efect at the time of the interview or those ended by the death of a spouse
are treated as censored data. Both the complete episodes and the censored
ones were used to estimate the β and λ-parameters. In the presence of
censored data, the maximum likelihood method provides consistent and
asymptotically normally distributed estimates of the parameters. Partial
likelihood estimation (Cox 1972) of β1, β2,..., βm which does not require
specication of the baseline hazard rate function r(t) leads to practically
identical estimates as the ones reported.
Episode splitting is a method for decomposing an episode like marriage duration into
subintervals. Covariates remain constant within subintervals, and the likelihood function
can therefore be rewritten as a product of the subinterval-specic likelihoods. For technical
details see, for example, Blossfeld and Rohwer (1995).
Figure 1
The sickle model of the divorce risk function. Hazard rate curves for difer-
ent values of the parameters λ and a.
0.004
0.003
0.002
Divorce Risk
0.001
0 60 120 180 240
Time in Months
 = 180, a = 5.10-5
 = 120, a = 8.10-5
 = 120, a = 5.10-5
 = 120, a = 2.10-5
 = 60, a = 5.10-5
300 360 420
0
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A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14 7
4 Results
We estimate three diferent models. The estimated transmission efects
from those three models are summarized in Table 1 and the respective per-
centage efects visualized in Figure 2. Model 1 controls only for marriage
cohorts, Model 2 controls for these cohorts and home of origin, and Model 3
includes the additional covariates of date of the rst child’s birth, cohabita-
tion, age at start of union, and educational level (see Figure 2 and Table 1, for
more detailed information see Tables A.4 to A.6 in the Appendix).
800%
700%
600%
500%
Transmission efect, [exp()‒1]
Austria
Belgium
Canada
Czech
Estonia
E-Germany
W-Germany
Hungary
Italy
Latvia
Lithuania
Slovenia
Spain
Sweden
Switzerland
USA
400%
300%
200%
100%
Model 1: controlled for cohorts
Model 2: additionally controlled for parental home
Model 3: additionally controlled for education, age at start of union, cohabitation, children
0%
Figure 2
The intergenerational transmission efect of divorce. Summary of estimates
of β1 from country specic maximum likelihood estimations of hazard rate
models with diferent sets of control variables. Plotted is the percent efect
exp(β1)–1. Reading the transmission efect (e.g., Austria): the percent efect
of 101% means that children whose parents divorced have a 101% higher
risk of divorce than children whose parents did not divorce. 95% con-
dence interval based on (non-linear) transformation of condence bounds
of the estimated β1. Full estimation results are provided in Tables A.4 to A.6
in the Appendix.
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8 A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14
Austria Belgium Canada Czech Estonia E-Germ W-Germ Hungary Italy Latvia Lithu. Slovenia Spain Sweden Switz. USA
Intergenerational transmission effect in model 1: controlled for cohorts
Parents divorced 0.70
***
1.04
***
0.71
***
0.55
***
0.47
***
0.57
***
0.86
***
0.38
**
1.15
***
0.42
***
0.58
***
0.84
***
0.80
*
0.72
***
0.73
***
0.45
***
(0.11) (0.17) (0.09) (0.15) (0.12) (0.13) (0.17) (0.12) (0.29) (0.09) (0.13) (0.24) (0.31) (0.14) (0.12) (0.05)
Intergenerational transmission effect model 2: additionally controlled for parental home
Parents divorced 0.62
***
0.83
***
0.71
***
0.40
*
0.41
**
0.44
**
0.89
***
0.36
**
1.55
***
0.40
***
0.56
***
0.91
***
0.82
*
0.48
*
0.70
***
0.43
***
(0.12) (0.21) (0.09) (0.18) (0.13) (0.15) (0.19) (0.14) (0.32) (0.11) (0.16) (0.27) (0.33) (0.20) (0.13) (0.05)
Intergenerational transmission effect model 3: additionally controlled for education, age at start union, cohabitation, children
Parents divorced 0.55
***
0.72
***
0.62
***
0.28 0.32
*
0.38
**
0.83
***
0.30
*
1.42
***
0.33
**
0.50
**
0.84
**
0.59 0.42
*
0.66
***
0.31
***
(0.12) (0.21) (0.1) (0.18) (0.13) (0.15) (0.19) (0.14) (0.32) (0.11) (0.16) (0.27) (0.33) (0.2) (0.13) (0.05)
All marriages in sample
Number 3217 2373 2429 1268 1237 1733 1289 2756 3163 2041 2145 1958 2674 1839 3059 6518
Of which divorced 18% 11% 29% 21% 27% 20% 18% 17% 5% 29% 15% 8% 5% 19% 15% 35%
Marriages with non-divorced parents
Number 2868 2192 2043 1058 955 1436 1141 2317 3070 1579 1779 1822 2564 1628 2705 4915
In % of all marriages 89% 92% 84% 83% 77% 83% 89% 84% 97% 77% 83% 93% 96% 89% 88% 75%
Of which divorced 17% 10% 27% 19% 25% 18% 16% 16% 5% 28% 14% 7% 5% 18% 15% 33%
Marriages with divorced parents
Number 349 181 386 210 282 297 148 439 93 462 366 136 110 211 354 1603
In % of all marriages 11% 8% 16% 17% 23% 17% 11% 16% 3% 23% 17% 7% 4% 11% 12% 25%
Of which divorced 28% 22% 41% 30% 34% 28% 31% 21% 14% 35% 21% 15% 10% 29% 23% 42%
Notes: Reported is the parameter β
1
of the maximum likelihood-estimations of the hazard rate r(t) = a t exp(-t/λ) where a = exp(β
0
+ x
1
β
1
+ x
2
β
2
+ … + x
m
β
m
) and x
1
indicates whether
the parents have been divorced. Standard errors in parentheses. Parameters with (***,**,*) are significantly different from 0 at p < .001 resp. p < .01, p < .05. The complete maximum
likelihood-estimations of the sickle models for each country are reported in Tables A.4 to A.6 in the Appendix.
Table 1
Survival analysis of divorce risk, summary of estimations by countries
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A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14 9
Our rst nding is the universal existence of the transmission efect. Model
1 in Figure 2 shows the percentage transmission efect, i.e. exp(β1)-1, for chil-
dren from divorced families versus respondents from non-divorced families
for all samples. The efects range from 50% for Hungary and Latvia to 220%
for Italy with an average efect of 103%. This means that children whose
parents divorced face – on average – a 103% higher risk of getting divorced
than children whose parents did not divorce. The efects for all 16 data sets
are substantially larger than 0 and highly signicant (Spain at p < 0.05, Slo-
venia at p < 0.01; all other countries at p < 0.001). In contrast to a previous
analysis (Diefenbach 1997), we also nd a highly signicant transmission
efect for East Germany. Our analysis clearly shows that the intergenera-
tional transmission of divorce is a widespread phenomenon observed with-
out a single exception in formerly communist Eastern Europe, Southern
(Catholic) Europe, Western Europe and North America.
Children of divorced parents usually grow up with one parent only. In
the next step we therefore disentangle the efect of the parent’s divorce
from that of the parental home. Model 2 controls for whether the person
grew up with one parent only. The two efects can be separately identied
as some children lose a parent for reasons other than divorce. Our results
show that the transmission efect is only moderately changed in all coun-
tries but Italy, where it is substantially increased. The transmission efect
remains highly signicant in all countries.
In model 3 we include educational level, age at start of union, cohabita-
tion and the birth of children as additional control variables. Transmission
efects are again only moderately reduced compared to the unconditional
efects in model 1, except in Italy (where few of the respondents’ parents
were divorced), and remain signicant at p < 0.05 in all countries but Spain
(p < 0.1) and the Czech Republic. This reduction is a result of controlling for
cohabitation and age at start of the union, since in all countries the chil-
dren of divorce cohabit more and begin unions earlier.
The pattern of union formation and cohabitation sheds light on the
explanation of the transmission efect. In all countries but one, the esti-
mated coecient indicates a higher risk of divorce for couples that lived
together before marriage (see Table A.6). In addition, the well-known nega-
tive efect on divorce of age at union formation shows up in all countries. At
the same time, the children of divorce cohabit more frequently than respon-
dents from non-divorced families and enter into unions at younger age in
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10 A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14
all countries (compare Tables A.2 and A.3). Although cohabitation does
likely not exert a positive causal efect on divorce, the covariate is corre-
lated with the divorce risk. Previous studies have shown that spouses living
together before marriage form a selective group that is more divorce-prone
than those entering into traditional marriages (Brüderl and Kalter 2001).
For older marriage cohorts in particular, cohabitation signals that spouses
are less committed to marriage than spouses who did not live together
before marriage. The observed patterns t well with the “low-commitment
hypothesis” (Amato and DeBoer 2001, Wolnger 2005). Accordingly, one
causal pathway is that children of divorce develop more sceptical attitudes
toward a long-lasting partnership, therefore choose less binding commit-
ments when starting a union, and are ultimately more likely to divorce.
The systematic efect of parental divorce on the pattern of educational
attainment is also noteworthy. In all countries except one, the children
of divorced parents participate in advanced education (ISCED 5 or 6) to
a lesser extent than respondents with non-divorced parents. Yet, as far as
possible efects of this disparity on marriage stability are concerned, edu-
cation is not consistently related to the respondent’s divorce risk and is
therefore not a mediator variable in explaining part of the transmission
efect in almost all the countries studied. Women’s education has mul-
tiple consequences on such divorce-related risk factors as labour-force
participation, personal income, household income and cultural prefer-
ences. Some of these factors increase divorce risks while others may have
a positive impact on marital stability. Thus, it is no surprise that we do
not nd a robust efect for a woman’s education on marital stability for all
the national data sets. Previous studies have also not reported consistent
ndings on the efect of a woman’s level of education on her divorce risk
(Dourleijn and Lieroer 2002).
5 Explaining Cross-Country Variation of Transmission Efects
Our estimations show substantial variation in the magnitude of the divorce
transmission efect ranging from 0.38 in Hungary to 1.34 in Italy. Can these
diferences across counties be explained by historical, institutional and cul-
tural diferences?
We did not nd any systematic pattern that related the magnitude of the
transmission efect to, for example, catholic religion, former communist
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A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14 11
countries, etc. However, there is one striking relationship (see Figure 3):
the transmission efect is highly negatively correlated with the divorce
rates of the parent population (slope = 3.46, t-value = 4.38, R2 = 0.58). This
result is especially important as it seems to conrm the hypothesis that the
detrimental efects of divorce on children are stronger in societies where
divorce is rare and thus more likely stigmatised.
While highly appealing, we believe that this nding is an artefact for two
reasons. First, the negative correlation between transmission efect and
parental divorce rates found across countries is not found within countries
over time: while divorce rates increased in all countries, the transmission
efect decreased in only half of the countries and increased in the others.
Our scepticism is in line with Li and Wu (2004) who show that Wolnger’s
(1999) nding of a decrease in the U.S. divorce transmission efect was
awed. Second, the negative correlation between transmission efect and
Detailed estimation results are available from the authors on request.
Figure 3
Intergenerational divorce transmission and divorce rates by country.
Slo
Ita
Spa
0.1.2
Divorce Rate
.3
Transmission Efect, beta
0 0.5 1 1.5
Bel
WGer
Swi
Aus
Swe
Can
Lit
Cze
Lat
Est USA
HunEGer
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12 A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14
parental divorce rates found for the relative increase in divorce risk is not
found for the absolute increase in divorce risk: while the relative efects dif-
fer by a factor of almost 3 across countries, the absolute efects are of simi-
lar magnitude. Relative increases are the result of dividing absolute efects
by the baseline divorce risk. The signicant negative relationship between
the relative transmission efect and parental divorce rates can simply be
explained by the self-evident positive relationship of the left-hand side
numerator (the divorce rate of the observed generation) with the explana-
tory variable (the divorce rate of the parents’ generation).
6 Conclusions
This study investigates the intergenerational transmission of divorce in
13 European countries, the United States and Canada. We analyze the
cross-national data from female respondents in the Fertility and Family
Survey, applying techniques of event history analysis. We nd substantial
and highly statistically signicant transmission efects in all samples. This
shows that the intergenerational transmission of divorce is a widespread
phenomenon observed without a single exception in our data covering a
large number of countries with difering historical, institutional, and cul-
tural contexts.
Our study also demonstrates the presence of systematic patterns in the
consequences of divorce on children’s marital behaviour. Women whose
parents had divorced were, in all countries, also more likely to cohabit with
the men they eventually married than women who grew up with both of
their parents. This nding is in keeping with previous studies, which have
also noted greater rates of cohabitation among respondents with divorced
parents (e.g. Amato 1996, Diekmann and Engelhardt 1999, Kiernan and
Cherlin 1999). Our study adds to this nding by showing that the increased
tendency to cohabit is a rather universal phenomenon, one observed – at
least for female children – in all the countries we studied. This result ts
with Amato and DeBoer’s (2001) suggestion that the children of divorce
have less favourable attitudes toward marriage and therefore choose less
binding commitments when starting a union.
Detailed estimation results are available from the authors on request.
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A. Diekmann, K. Schmidheiny / Comparative Sociology 12 (2013) 1–14 13
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... V Evropě se mezigeneračním přenosem rozvodu zabývali například ve Francii [Traag, Dronkers, Vallet 2000], Švédsku [Gähler, Härkönen 2014], Německu [Engelhardt, Trappe, Dronkers 2002] či Velké Británii [Kiernan, Cherlin 1999] nebo Norsku [Lyngstad, Engelhardt 2009]. Výzkum byl v posledních letech rozšířen i o mezinárodní srovnání [Dronkers, Härkönen 2008;Diekmann, Schmidheiny 2013]. ...
... Hlavním předmětem výzkumů ovšem není samotná existence mezigeneračního přenosu rozvodu. Například Amato označuje rozvod rodičů za "dobře zdokumentovaný rizikový faktor pro ukončení manželství" [Amato 1996: 628], který můžeme pozorovat napříč různými kulturami [Diekmann, Schmidheiny 2013;Dronkers, Härkönen 2008]. Výzkumníci se proto snaží především přijít na vysvětlení vztahu mezi rozvodem rodičů a rozvodem dětí, jež zatím zůstává nejasné [Amato 1996]. ...
... Nízký věk při první svatbě je přitom jedním z dobře zdokumentovaných rizikových faktorů rozvodu [Booth, Edwards 1985;Härkönen, Dronkers 2006]. Zároveň jsou to právě děti rozvedených rodičů, kteří před vstupem do manželství častěji žijí v kohabitaci [Thornton 1991;Diekmann, Schmidheiny 2013;Härkönen, Brons, Dronkers 2020], což je další faktor, který zvyšuje riziko rozvodu [Dush, Cohan, Amato 2003]. Kromě toho dosahují tyto děti nižšího vzdělání [Kreidl, Štípková, Hubatková 2017], přičemž nízký socioekonomický status je opět spojen s vyšším rizikem rozvodu [McLeod 1991;Gähler, Palmtag 2015]. ...
... Divorce is a commonly studied example of relationship instability, and the research on this topic has found that individuals who experience a parental divorce are more likely to divorce in adulthood themselves (e.g., Amato, 1996;Bumpass et al., 1991;Pope & Mueller, 1976). This intergenerational transmission has been consistent in studies across the United States, a variety of European countries, and Canada (e.g., Diekmann, & Schmidheiny, 2013;Dronkers, & Härkönen, 2008). Additionally, the number of parental relationships an offspring witnessed during childhood is positively associated with their own number of partners in early adulthood, demonstrating an intergenerational transmission of union transitions (Amato & Patterson, 2017;Kamp Dush et al., 2018). ...
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