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This article was downloaded by: [Bibliotheek Der Ru]On: 23 February 2010Access details: Access Details: [subscription number 914119905]Publisher RoutledgeInforma Ltd Registered in England and Wales Registered Number: 1072954 Registered office: Mortimer House, 37-41 Mortimer Street, London W1T 3JH, UK
Population StudiesPublication details, including instructions for authors and subscription information:http://www.informaworld.com/smpp/title~content=t713689546
Preferences for the sex-composition of children in Europe: A multilevelexamination of its effect on progression to a third childMelinda Mills a; Katia Begall a
a University of Groningen,
Online publication date: 17 February 2010
To cite this Article Mills, Melinda and Begall, Katia(2010) 'Preferences for the sex-composition of children in Europe: Amultilevel examination of its effect on progression to a third child', Population Studies, 64: 1, 77 — 95To link to this Article: DOI: 10.1080/00324720903497081URL: http://dx.doi.org/10.1080/00324720903497081
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Preferences for the sex-composition of childrenin Europe: A multilevel examination of its effect
on progression to a third child
Melinda Mills and Katia BegallUniversity of Groningen
Comparative research on the preferred sex of children in Western societies has generally focused on women
only and ignored the role of gender equity and the need for children’s economic support in old age. A
multilevel analysis extends existing research by examining, for both men and women and across 24
European countries, the effect of the preferred sex-composition of offspring on whether parents have or
intend to have a third child. Using the European Social Survey (2004/5), a multilevel (random coefficient)
ordered logit regression of that intention (N � 3,323) and a binary logistic multilevel model of the
transition to a third child (N � 6,502) demonstrate the presence of a mixed-sex preference. In countries
with a high risk of poverty in old age, a preference for sons is found, particularly for men. In societies where
there is lower gender equity, both men and women have a significant preference for boys.
Keywords: fertility; children; parents; gender; sex-composition; equality; equity; Europe
[Submitted 29 January 2009; Final version accepted 26 October 2009]
Introduction
There is a long tradition in demography of the study
of parents’ preferences for the sex of their children.
Most studies have focused on Asian countries such
as India and China, where son preference has led to
sex-selective abortions and skewed sex ratios (e.g.,
Park and Cho 1995; Arnold 1997; Chen et al. 2007;
Song and Burgard 2008). Although parents’ prefer-
ences are rarely overtly acknowledged at the in-
dividual level (e.g., Dyson and Moore 1983), they
are a latent manifestation of the level of gender
equity (equity in the opportunities available to men
and women) within a particular society (Pollard and
Morgan 2002). Since unequal status has also been
linked to low fertility (e.g., McDonald 2000, 2006)
and since Bongaarts and Potter (1983) have demon-
strated that the preference for a child of each sex can
operate to increase fertility levels, the impact of sex-
composition preference on the transition to higher-
order births is a particularly relevant topic for
European countries with fertility in the ‘lowest-
low’ category (Kohler et al. 2002).
In recent years, a growing number of studies have
examined this phenomenon within Western societies
(e.g., Yamaguchi and Ferguson 1995; Hank and
Kohler 2000; Brockmann 2001; Pollard and Morgan
2002; Lundberg 2005; Andersson et al. 2006, 2007;
Raley and Bianchi 2006; Kippen et al. 2007). While
concern over a preference for boys has dominated
non-Western research, findings show that this does
not occur in Western countries, with the exception of
a slight boy preference in Finland (Andersson et al.
2006). Rather, there appears to be a preference for a
balanced sex-composition of at least one boy and
one girl, a preference found in Australia (Gray and
Evans 2005; Kippen et al. 2007), the USA (e.g.,
Yamaguchi and Ferguson 1995), and across 17
European countries (Hank and Kohler 2000).
The aim of the study presented here was to extend
previous research by examining the impact of the
existing sex-composition of offspring on the propen-
sity of men and women in European countries to
have a third child, using a cross-national comparative
multilevel design. The central research question was
whether the preference for a mixed-sex family (i.e.,
at least one boy and one girl) drives intended and
actual progression to a third child and how this
varies by key individual and macro-level character-
istics. As well as complementing existing research,
Population Studies, Vol. 64, No. 1, 2010, pp. 77�95
ISSN 0032-4728 print/ISSN 1477-4747 online/10/010077-19 # 2010 Population Investigation Committee
DOI: 10.1080/00324720903497081
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we wished to expand what we know about this topic
in several key ways.
Unlike our study, virtually all previous studies
have, owing to data limitations, focused on women’s
sex-composition preference in relation to fertility
intentions and behaviour, leaving us with a limited
understanding of the effect of men’s preferences (for
an exception see a recent study by Gray et al. 2007).
Although the examination of birth outcomes im-
plicitly includes the behaviour of both male and
female partners, fertility intentions may differ con-
siderably between them, particularly in relation to
sex-composition preference. There may also be
marked differences between men and women in
the way this preference is influenced by education,
the opportunity costs of having another child, and
the extent of gender equity in society.
Previous studies have argued that, in theory, sex-
composition preference is related to country-level
institutional factors such as the extent of gender
equity (Rindfuss et al. 1996; Mason 1997; McDonald
2000, 2006) and the economic need for children in
old age (Brockmann 2001), both of which vary
considerably across nations. Ours was the first study
to operationalize these theoretical constructs and
include them in statistical models.
A third and related advance was this study’s
adoption of a multilevel design. Although cross-
national comparative studies have been conducted
on this topic (e.g., Hank and Kohler 2000; Marleau
and Saucier 2002; Andersson et al. 2006), they used
individual-level regression models rather than multi-
level modelling, whereas only the latter can show
whether there is significant variation between coun-
tries. For example, in Hank and Kohler’s study of
parents’ sex-composition preferences across 17
countries in Europe (2000), individual-level probit
regression models were run separately for each
country, providing us with little statistical corrobora-
tion of the significance of the level of variation
between countries.
Finally, this study complements existing work by
applying multi-sex, multilevel models to examine the
impact of preferences for the sex of offspring on
both fertility intentions and behaviour. Existing
research often examines behavioural outcomes
only in the form of a ‘sex-of-previous-children’ effect
by examining the transition to the second*and most
often*third child. The underlying logic is that
parents who have two children of the same sex are
more likely to progress to a third child than those
with a mixed-sex set of children. Higher rates of
progression to a third child therefore serve as a
proxy for preferences for the sex of children. Fewer
studies have included an analysis of fertility inten-
tions in relation to preferences, and even fewer have
included both intentions and behaviour (for excep-
tions, see Hank and Kohler 2000; Marleau and
Saucier 2002; Pollard and Morgan 2002). Yet we
know that there is often a mismatch between fertility
intentions and behaviour (Quesnel-Vallee and Mor-
gan 2003; Toulemon and Testa 2005).
The next section presents an overview of previous
research and theoretical explanations that have
attempted to explain the phenomenon of parents’
preferences for the sex of children. This is followed
by a description of the European Social Survey
(2004/5), which provided the data we used to study
the phenomenon. The multilevel regression
analyses of third-child intentions and transitions
are then described, followed by the presentation
and discussion of results in relation to our expecta-
tions. We close by reflecting on our findings in light
of existing research and with suggestions for further
research.
Parents’ sex-composition preferences: Previousresearch
Previous research in Western societies has produced
largely complementary evidence of a preference for
mixed-sex offspring. Examining the transition to
second and third children using three waves of the
National Survey for Family Growth (NSFG)
(1973�82) in the USA, Teachman and Schollaert
(1989) demonstrated that having a boy increased the
rate of transition to a second child and that having
two children of the same sex hastened transition to a
third. Pollard and Morgan (2002) found similar
results in the USA for a later period when they
examined the odds of progressing to a third child
using four waves of the CPS (1980�95) and three
waves of the NSFG (1983�95). Women with two
children of the same sex were more likely to
progress to a third child, although the effect wea-
kened over time.
A preference for mixed-sex offspring has also
been found for the transition to a third child in
Australia (Gray and Evans 2005) and predominantly
across Europe (Hank and Kohler 2000). Although
the findings are generally consistent, some studies in
European countries have found deviations from
the tendency to prefer mixed-sex offspring between
countries or over time. Using data from population
registers in Finland, Norway, Sweden, and
Denmark, Andersson et al. (2006) compared the
78 Melinda Mills and Katia Begall
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probabilities of a second and third child by the sex
of previous children. In all four countries, a pre-
ference for a balanced composition at parity two
was found. At parity one in Sweden, Norway, and
Denmark, there was evidence of a girl preference
with parents of a daughter progressing to a second
child more slowly than parents of a son, but in
Finland the opposite was found. Three studies have
revealed a slight girl preference: in the Czech
Republic and Lithuania (Hank and Kohler 2000),
in Denmark, Sweden, and Norway (Andersson et al.
2006), and among first-time pregnant women in
the USA, England, Sweden, Israel, and Canada
(Marleau and Saucier 2002).
There are also numerous studies of parents’
preferences for the sex of children in the biomedical
literature, largely related to prenatal and ultrasound
examinations. The medical literature offers interest-
ing insights because measures are sometimes more
direct and couples are interviewed at the time of the
pregnancy. Shipp et al. (2004), for instance, verified
that the sex of previous children had an influence on
couples’ future childbearing plans. From a study
using a sample of pregnant couples in the USA (N�1,300), the investigators concluded that having only
one boy or one girl was associated with a higher
proportion wishing to know the sex of the foetus,
whereas the response of parents who already had
one child of each sex was ‘sex neutral’.
In another study, Dahl et al. (2006a) found a
preference for a mixed-sex family in the USA and
the UK, while the majority of German respondents
showed no preference for the sex of their children.
In a study of 1,197 men and women in the USA,
Dahl et al. (2006b) reported that only 27 per cent
stated no sex preference for their children. The
remaining parents showed a clear preference for a
mixed-sex set of children, ranging from 50 per cent
who wanted an equal number of boys and girls to
7 per cent desiring more boys than girls, 6 per cent
more girls than boys, 5 per cent only boys, and 4 per
cent only girls. In a study of sex-selection in IVF
implantation, 40.8 per cent of parents stated that
they would want to select the sex of their next child
if there was no added cost (Jain et al. 2005). Of
these, almost 46 per cent had no previous children
and 48 per cent already had children of one sex.
Another key finding of this study was that there was
a significant preference for a child of the opposite
sex for those who already had one or more children
of the same sex. The fundamental questions are how
these preferences emerged and why they exist in
Western societies.
Parents’ sex-composition preferences:Theoretical explanations
The development of preferences for the sex of
children may be a result of both institutional and
individual-level factors. To develop a theoretical
explanation of why a mixed-sex preference has
emerged in Europe, we combine previous explana-
tions from (largely non-Western) literature on pre-
ferences for sons or daughters with more general
institutional and individual-level explanations.
Son preference in Western societies
Most literature on preferences for the sex of
children has focused on son preference in non-
Western societies, but the fact that parents in
Western societies usually want to have ‘at least one
boy’ indicates the existence of a son preference in
these societies too. The tradition of son preference
in East and South Asia, the Middle East, and North
Africa (Arnold 1997; Chen et al. 2007), has been
attributed largely to three factors. First, sons are
more useful economically, which makes them parti-
cularly valuable in agrarian economies (Basu 1989).
A second vital role of sons is to continue the family
line and carry on the family name, a factor that is
arguably also relevant in many Western societies. A
third reason is the fact that men are often the
recipients of inheritance (Arnold et al. 1998).
In contrast to the findings of research in non-
Western countries, there is no contemporary evi-
dence of an aversion to daughters in Western
countries. Where the dowry system exists, daughters
are often viewed as an economic burden. As Leone
et al. (2003) show, in some societies girls have little
value when parents become older. If they become a
member of the husband’s household after marriage,
they are no longer in a position to care for their
parents in their old age and are not expected to do
so. Rather, this is the duty of the son and his wife. In
most Western and European societies, the function
of daughters in elderly care is often very different, a
topic that we will now address.
Daughter preference in Western societies
The findings of Cleland et al. (1983) about the
important role played by daughters in a non-
Western society point to reasons for the emergence
of a preference for at least one daughter in Western
Sex-composition preference and the third child 79
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societies: (i) they help to care for younger siblings;
(ii) they engage in household tasks that reduce the
workload of the parents; (iii) they engage in paid
labour and contribute economically to the house-
hold; and (iv) they provide care for their parents in
old age that is more reliable and of higher quality
than is available from sons.
The emergence of more supportive welfare re-
gimes in many European countries has lessened the
need for children to provide financial assistance to
their parents in old age. The fact that it is no longer
necessary to have at least one son to provide
financial assistance in old age, coupled with the
growing economic independence of women, could
result in the weakening of son preference and even a
growth in a preference for girls. That development
would be consistent with Brockmann’s findings
(2001) of the existence of a son preference in the
generation of women born in Germany before 1910,
its disappearance in the next generation, and the
subsequent emergence of a girl preference after the
Second World War. These changes, she argues, can
be explained by changes in the welfare regime.
While the support of male offspring (and possibly
the care of the daughter-in-law) was still required in
the period before the war, the post-war socialist
regime established a generous state support system
for families and strongly encouraged women to
participate in the labour market. This, according to
Brockmann (2001), led to a shift in the favouring of
daughters over sons because daughters could pro-
vide the same economic support as sons and the
additional benefit of more emotional support in old
age. We know from existing research that women are
the primary care-givers for elderly parents in Wes-
tern societies (Brody 2004). Therefore, a preference
for girls could emerge in countries where there is
more equity in the way men and women are treated,
a more equal division of household labour, higher
levels of paid employment, and a stronger and more
positive role for women in society (see Hammer and
McFerran 1988).
Mixed-sex preferences in Western societies
As pointed out previously, most research in Western
countries has not found a preference for only boys or
only girls, but rather that parents prefer a mixed-sex
family (Teachman and Schollaert 1989; Yamaguchi
and Ferguson 1995; Hank and Kohler 2000; Pollard
and Morgan 2002). But why does this preference for
mixed-sex offspring exist?
Hypotheses
Institutional-level hypotheses
Previous studies (e.g., Hank and Kohler 2000;
Andersson et al. 2007) have demonstrated that there
appear to be specific preferences for the sex of
children within clusters of homogeneous countries,
such as a son preference in societies that have a
limited welfare system and conservative gender
roles. With that in mind, we first developed institu-
tional-level explanations for a mixed-sex preference
that included the level of gender equity and income
support in old age (and the consequent economic
need to have children). We reasoned that, in relation
to Brockmann’s study (2001), it might be that
parents desired both a boy and a girl as a means of
‘uncertainty reduction’ in old age. Further, building
on economic theories of the utility of children (e.g.,
Becker and Barro 1988) and Friedman et al.’s
uncertainty reduction theory (1994), we thought
parents in Western societies might desire a balanced
sex-composition in their children, not only to reduce
economic uncertainty in old age, but also to secure
economic, physical, and emotional care. For most
societies, a plausible institutional explanation for the
desire to have at least one boy is a boy’s greater
economic value and stronger guarantee of financial
support for elderly parents, together with his ability
to carry on the family name and inherit its assets. On
the other hand, girls in most European societies
could offer not only economic support (because they
were doing paid work), but also the additional
advantage of more help with housework and the
care of siblings. These unique contributions of boys
and girls are probably the reasons for the observed
preference for mixed-sex families. The first and
central hypothesis of this study was therefore that
those with two children of the same sex are more
likely to have or to intend to have a third child
than those who already have one boy and one girl.
We describe this as the mixed-sex preference hypoth-
esis (H1).
The reasoning about the need to have children for
support in old age as a way of reducing economic
uncertainty leads us to our second set of hypotheses,
which we term the old-age security hypotheses (H2).
The first of these (H2a) was that individuals in
80 Melinda Mills and Katia Begall
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countries with a relatively high risk of falling into
poverty in old age have a higher economic need to
have boys for economic support in old age, and that
for this reason those with two girls are more likely to
intend to have or to have a third child. The second of
our old-age security hypotheses invoked Maslow’s
classic theory of human motivation and the hier-
archy of needs (1943), which argues that when
individuals’ physiological and safety needs are
fulfilled, they desire the realization of love and
belonging from family in the form of emotional
care, which we assume is more likely to be provided
by daughters. Thus sub-hypothesis H2b states that,
for those with two boys, higher security and a lower
risk of falling into poverty in old age will translate
into a higher likelihood to have or to intend to have
a third child.
Gender equity (equity in society’s allocation of
opportunities to men and women) is another rele-
vant institutional factor shaping fertility behaviour.
Pollard and Morgan (2002) were the first to empha-
size the importance of modernization in shaping sex-
composition preferences, with an explanation that
had its origins in research on gender equity
and fertility (Rindfuss et al. 1996; Mason 1997;
McDonald 2000; Bongaarts 2001). The ‘gender
system’ of a society shapes prevalent gender roles
and dictates which opportunities are equally avail-
able to men and women (e.g., opportunities for
health care, education, the labour market) (Mason
1997; Mills et al. 2008). Bongaarts (2001) and Pollard
and Morgan (2002) maintain that in societies with
relatively high levels of gender equity, parents can
be expected to be ambivalent about the sex of their
children because boys and girls are seen as sub-
stitutes for each other in what they can provide. We
therefore framed a gender equity hypothesis (H3),
according to which, owing to the equal capacities of
boys and girls in nations with high levels of gender
equity, parents with two girls and no boys are
significantly less likely than those in societies with
low gender equity to have or to intend to have a third
child. Conversely, in the low-gender-equity nations
of Europe, parents of two girls are more likely than
those in high-gender-equity nations to have or to
intend to have a third child, that is, to show a son
preference.
Because we adopted a multilevel research design,
we were also able to test a cross-national variation
hypothesis (H4). According to this hypothesis, owing
to institutional differences and variation in cultural
norms across Europe, which have been documented
in many previous studies (e.g., Hank and Kohler
2000; Kohler et al. 2002; Sobotka 2005; Andersson
et al. 2006, 2007), there are significant differences in
both intention to have and actually having a third
child across the different countries of Europe.
Individual-level explanations
As well as institutional factors, the motivation to
have children is also related to individual goals for
them. Since Aries’ seminal book (1962) on the
transformation of the function of children in Wes-
tern societies, others such as Hoffman and Hoffman
(1973) have demonstrated that children provide a
social identity for parents, an expansion of the self,
primary group ties, stimulation, fun, creativity, and a
feeling of power (as a means of social competition
and comparison). Zelizer (1994) refers to the shift in
the value of children in the USA from 1870 to 1930
from the economically ‘useless’ to the emotionally
‘priceless’. In many households in the economically
advantaged Western countries, children are not
required to contribute to household earnings, per-
form an economic role within the household, or
support their parents financially in their old age.
Instead, parents apply what Zelizer describes as
‘sentimental criteria’, and treat a child as a priceless
commodity. Parents see children as a means of
extending their own emotional satisfaction and
self-actualization, which may lead them to prefer
to have a child of their own sex.
The tendency to seek and bond with similar
others*homophily*has been extensively studied
and confirmed in social network research. McPher-
son et al. (2001) demonstrated that the homophily
principle structures all types of network ties (e.g.,
work, friendship, advice, exchange) resulting in
individuals developing highly homogeneous perso-
nal networks. It therefore seemed plausible to apply
the homophily principle in interpreting preferences
for sex of offspring, and to assume that a preference
for similarity would also prompt a preference for
same-sex children.
There is some support for homophily in evidence
that parents have a preference for, and a stronger
affiliation with, a child of the same sex (Kippen et al.
2007). Marleau and Saucier (2002) demonstrated that
women consider girls to be easier to bring up and to be
more rewarding as companions. Morgan et al. (1988)
found that fathers appear to have stronger attach-
ments and obligations to their children and to the
Sex-composition preference and the third child 81
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maintenance of marital harmony when they have
sons, which hints at an intrinsic same-sex preference.
A preference for the same sex has also been found in a
variety of other areas of study, such as same-sex hiring
in the employment sphere (e.g., Gorman 2005). If this
same-sex preference exists across heterosexual part-
ners, the preferences of each partner would effec-
tively cancel each other out, resulting in the mixed-
sex preference that we observe. This reasoning led to
our fifth hypothesis, the homophily hypothesis (H5):
in order to have at least one child of their own sex,
fathers of two girls and mothers of two boys are more
likely to intend to have a third child than are those
who already have a child of their own sex.
Largely owing to the limitations of registration
data and the fact that fertility data are often only
collected from women, little attention has been paid
to differences between men and women in preferred
sex of offspring (for an exception see Kippen et al.
2007). We know that the costs associated with
childbearing and rearing are higher for women
than for men, especially where state support such
as childcare facilities is scarce (Rindfuss et al. 1996).
Women also lose income when they leave the labour
force for such reasons as maternity leave; in contrast
to first and second children, the third child probably
prompts parents to decide if it is economically
feasible for one of them to leave the labour force
temporarily. Owing to the higher costs for women,
we also assumed that they would adjust their sex
preference for children sooner and more definitely
than men, which might diminish observed prefer-
ences. These considerations led us to formulate an
opportunity cost hypothesis (H6): women are less
likely to intend to have a third child irrespective of
the sex of their previous children.
Other individual-level factors that we controlled
for and considered in our model were education
level, age, age at first birth, and children living
outside the household. Studies mentioned previously
led us to assume that those with higher levels of
education are less likely to have or intend to have a
third child owing to such factors as higher commit-
ment to the labour market. On the other hand
Kravdal (1992) found a positive association between
women’s educational level and progression to a third
child, which would suggest that higher education
represents a higher economic status and allows
individuals to have a larger family. Recent evidence
by the same author, however, suggests that the
positive relationship between higher levels of educa-
tion and a greater likelihood of a third child is due to
selection (Kravdal 2001). As a more direct proxy of
economic position, we also included home owner-
ship in initial analyses, but since the coefficients
were not significant and the model was not im-
proved, this variable was excluded from the final
model. We included the number of children living
outside the household to control for the fact that
families might be split up as a consequence of
divorce or separation. In fact they are relatively
few in number. A recently published study by Beets
(2009) found that in the Netherlands not more than
3�4 per cent of parents were living with the offspring
of different parents.
Data and methods
The data used in this study came from the second
wave of the European Social Survey (ESS), a
large-scale quantitative survey administered in 26
European countries in 2004/5. Since Turkey and the
Ukraine were excluded from the current analyses, 24
countries were included in the study; a list of the
countries is shown in the Appendix in Table A3.
Turkey was excluded because results for the country
resemble those for non-Western countries and
reveal large cultural and socio-economic differences.
Ukraine was excluded owing to a lack of data on
macro-level variables required for comparative
analyses.
In each country a representative random prob-
ability sample was drawn with strict quality controls
employed to ensure that all national samples met the
requirements of the study. Each wave of the ESS
consisted of a core questionnaire on attitudes and
values and rotating modules on other phenomena.
The 2004/5 wave contained a module on family,
work, and well-being that included data on family
life and fertility intentions, making it particularly
suitable for this study. The total sample of 49,066
respondents was reduced to those for whom it was
possible to examine fertility intentions at parity two
and the transition to a third child. Descriptive
statistics for all the variables described here are
shown in the Appendix in Table A1 (fertility
intentions) and Table A2 (transition to third child).
Measures
Dependent variables. The two dependent vari-
ables in our analyses were fertility intentions at
parity two and whether the respondent actually had
a third child. The intentions variable was measured
by asking whether the respondent intended to have
82 Melinda Mills and Katia Begall
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another child within the next 3 years, using the
following five-point scale: definitely not, probably
not, do not know, probably yes to definitely yes.
Explanatory variables. The central explanatory
variable*sex of previous two children*was entered
as three dummy variables: two boys, two girls, and,
as the reference category, one boy and one girl.
When models were run using the entire sample,
the sex of the parent was included, using men as the
reference category. Age of respondents at the
time of the interview was included in the fertility-
intentions analysis, using 5-year dummy variables to
account for the non-linear effect. Age at first birth
was included in models that examined the transition
to a third child with centred coding for ease of
interpretation.
The control for the educational attainment of the
respondent was measured in years of full-time
education, which was also centred. In previous
models (not shown here) educational attainment of
the partner was also examined but owing to high
correlations between individuals’ and partners’ edu-
cation and the absence of a difference in the
direction of the findings, partners’ education was
not included in the final models. Educational attain-
ment may be a proxy for a family’s economic
capacity to continue to a third child. Although
income would have been a more direct indicator, it
could not be included owing to the large number of
cases with missing values. Employment status was
also included in initial models (not shown here) but
owing to non-significant results is not shown in the
final models. As a more direct indicator of economic
position, we also controlled for home ownership,
which measured whether the house was owned by
one of the members of the household, but this was
left out of the final analysis because the results were
not significant.
To examine the impact of institutional effects, two
macro-level variables were included to operationa-
lize the theoretical constructs of gender equity and
economic need for children in old age. The values
and categorizations of each are shown in the
Appendix in Table A3. The risk of poverty above
the age of 60 was included as a proxy for whether
children would be required for economic support in
old age. The variable measured the proportion of
people older than 60 who had less than 60 per cent
of the median national income at their disposal
(median equivalent income after social transfers)
(Eurostat 2007; OECD 2007). Unfortunately, not
enough comparable data were available to make it
practical to include a more direct measure such as
the proportion of elderly living in institutions. In
previous models (not shown here), the percentage of
gross domestic product (GDP) spent on pensions
and then, separately, the median income of people
older than 65 years compared with the rest of the
population were included, and produced significant
estimates in the expected directions. The proportion
of GDP spent on pensions was not included because
it did not take into account the large differences
between European countries in both GDP and the
proportion of elderly people. The risk of becoming
poor was therefore the best proxy available that had
comparable data for all 24 countries.
A second macro-level variable was the Gender
Gap Index, developed by the World Economic
Forum (Hausmann et al. 2006). Building upon
and improving previous measures such as the
Gender-Related Development Index (GDI), Gender
Empowerment Measure (GEM), and the OECD
database on Gender, Institutions and Development,
the Gender Gap Index measures the level of
women’s economic participation and opportunity,
educational attainment, health and survival, and
political empowerment.
Both macro-level variables are included as three
dummy variables, with �1 representing one stan-
dard deviation under the mean, the reference
category representing the mean, and �1 indicating
the category of one standard deviation above the
mean. Owing to concerns about correlation between
the macro-variables, we also ran separate models
(not shown here) introducing the macro-level vari-
ables separately. Since the correlation between the
two macro-variables was only �0.1828, this proce-
dure made little difference to the results.
Methods of analysis
Fertility intentions at parity two. The purpose of
the first analysis was to estimate the effect of
preferred sex of children on fertility intentions at
parity two. A sub-sample was used consisting of all
respondents who (i) had two children who were
living in the household, (ii) resided with their
partner at the time of the interview (including both
married and unmarried cohabiting couples), and (iii)
were aged 18�40 (women) and 18�45 (men). This
sample was chosen to ensure that fertility intentions
were as realistic as possible. The age range reflects
the age difference in family formation between men
Sex-composition preference and the third child 83
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and women (men are on average 3 years older than
their partner). As shown in the descriptive statistics
in Table A1, we were left with a sub-sample of 3,459,
which was further reduced to 3,323 when cases with
too many missing values were omitted. Respondents
were divided equally between men and women, with
an average age of 36 years (SD�4.97 years).
The regression analyses were used to estimate the
fertility intentions of the entire sample (N�3,323)
and then separately of men (N�1,679) and women
(N�1,644) in two models. Model 1 included
the individual-level variables and the two macro-
variables of the Gender Gap Index and risk of
poverty in old age. Model 2 included the interaction
effects of the sex of the previous two children by
each of the macro-level variables.
The multilevel ordered logit model used was a
two-level random coefficient model with respon-
dents (i) nested in the country cluster j, which
included a random intercept uj for clusters in the
latent response model. For example, in order to
assess the magnitude of variation among countries,
the combined level 1 (individual) and level 2
(country) model 1 is
log
�Pr(yij � cjxij;zij; uij)
Pr(yij � cjxij;zij; uij)
�
�gc�(g01 sexpref �g01 childhhld�g01 female
�g01 age�g01 education�g01 house�g01 gendergap
�g01 riskpoverty�u(2)0
0 j )
where c�1, 2, 3, 4, 5, which are the five ordered
categories of fertility intentions.
We also tested whether we had violated the strong
assumption of this model, which is the parallel
regression assumption that the relation between
each pair of outcome groups is the same (Long
and Freese 2006). In the event, the assumption had
been (marginally) violated in the full model for the
sex-of-parent variable only, and for this reason we
included sex-specific thresholds in the full model,
which made it possible to relax the parallel regres-
sion assumption for sex of parent (Rabe-Hesketh
et al. 2004; Rabe-Hesketh and Skrondal 2008).
Transition to third child. For the analysis of the
transition to a third child, a different sub-sample of
respondents was used, comprising all respondents
who had two or more children living in the house-
hold and were not older than 45 years. This restric-
tion was necessary because the year of birth and
sex were recorded only for children living in the
household, and respondents older than 45 were
relatively likely to have children who were not living
in the household. We were left with a sample of
6,646, which was reduced to 6,502 when we had
taken missing values into account. The descriptive
statistics for variables used in this analysis are shown
in the Appendix in Table A2. The reason why the
number of respondents is much larger in this sample
is due to the fact that the dependent variable
includes not only those who reached parity two,
but also those with three or more children. The
respondents’ age at first birth was 25.8 years on
average (SD�4.74 years). The average number of
children was 2.4 and respondents completed 12.6
years of education on average (SD�3.50 years). The
modelling procedure was identical to that for the
analysis of fertility intentions, with the exception
that age at first birth was included instead of age. We
did not run separate models for men and women
since there was no theoretical or logical reason to
expect that they would have different birth out-
comes. The analysis employed a binary logistic
multilevel model with 1 indicating that a third child
had been born, and 0 indicating no third child.
Results
The results of the analysis of intention to have a
third child at parity two are shown in Table 1 and
the estimates of the progression to a third child in
Table 2. The results are discussed in relation to the
central hypotheses.
The first central hypothesis proposed that Eur-
opeans prefer a mixed-sex family: that is, that
compared with those who already have a mixed-
sex family, those with children of the same sex are
more likely to have or to intend to have a third child.
The results generally support this hypothesis, parti-
cularly in the case of the transition to a third child:
those with two same-sex children are significantly
more likely to have a third child than those with
mixed-sex families. The interesting differences that
emerge when we compare the estimates by the sex
of the parent are discussed below in relation to the
last two of our hypotheses.
Our old-age security hypothesis also assumed that
institutional factors would play an important role in
shaping parents’ preferences for the sex of children.
We first proposed (hypothesis 2a) that respondents
in countries where there is a higher risk of falling
into poverty in old age eventually have a greater
need for economic support in old age and therefore
have a boy preference, or alternatively that those in
84 Melinda Mills and Katia Begall
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Table 1 Determinants of intention to have a third child at parity two, 24 European countries 2004/5: coefficient estimates for multilevel ordered logit regression
Entire sample Men Women
Model 1 Model 2 Model 1 Model 2 Model 1 Model 2
Variable names and values b (SE) eb b (SE) eb b (SE) eb b (SE) eb b (SE) eb b (SE) eb
Sex of previous children�oneboy, one girl
ref 1.00 ref 1.00 ref 1.00 ref 1.00 ref 1.00 ref 1.00
Sex of previous children�twoboys
0.114(0.088)
1.12 0.288(0.149)
1.33� 0.193(0.123)
1.21 0.576(0.203)
1.78** �0.007(0.126)
0.99 �0.0726(0.224)
0.93
Sex of previous children�twogirls
0.207(0.090)
1.23* �0.085(0.154)
0.92 0.214(0.126)
1.24� 0.086(0.221)
1.09 0.175(0.130)
1.19 �0.247(0.218)
0.78
One or more children livingoutside the household
�0.124(0.190)
0.88 �0.128(0.191)
0.88 �0.306(0.239)
0.74 �0.333(0.241)
0.72 0.167(0.316)
1.18 0.135(0.319)
1.14
Sex�male 0.306(0.075)
1.36*** 0.293(0.075)
1.34***
Age 18�25 ref 1.00 ref 1.00 ref 1.00 ref 1.00 ref 1.00 ref 1.00Age 26�30 �0.452
(0.235)0.64� �0.445
(0.236)0.64� �0.070
(0.486)0.93 0.015
(0.490)1.01 �0.600
(0.273)0.55* �0.580
(0.275)0.56*
Age 31�35 �0.973(0.225)
0.38*** �0.979(0.226)
0.38*** �0.480(0.471)
0.62 �0.387(0.474)
0.68 �1.192(0.262)
0.30*** �1.186(0.263)
0.31***
Age 36�45 �1.941(0.225)
0.14*** �1.950(0.226)
0.14*** �1.448(0.468)
0.23** �1.360(0.471)
0.26** �2.173(0.267)
0.11*** �2.181(0.268)
0.11***
Years of full-time education(centred)
0.083(0.011)
1.09*** 0.082(0.011)
1.09*** 0.050(0.015)
1.05*** 0.050(0.015)
1.05** 0.122(0.016)
1.13*** 0.123(0.016)
1.13***
Gender Gap Index �1 SD 0.342(0.147)
1.41* 0.296(0.183)
1.34 0.334(0.226)
1.40 0.376(0.262)
1.46 0.508(0.252)
1.66* 0.357(0.293)
1.43
Gender Gap Index mean ref 1.00 ref 1.00 ref 1.00 ref 1.00 ref 1.00 ref 1.00Gender Gap Index �1 SD 0.113
(0.149)1.12 0.135
(0.199)1.14 0.219
(0.272)1.24 0.284
(0.305)1.33 �0.088
(0.260)0.92 �0.101
(0.296)0.90
Risk of poverty in old age�1 SD
0.423(0.154)
1.53** 0.195(0.209)
1.22 0.219(0.222)
1.25 0.162(0.254)
1.18 0.497(0.224)
1.64* 0.155(0.258)
1.17
Risk of poverty in old agemean
ref 1.00 ref 1.00 ref 1.00 ref 1.00 ref 1.00 ref 1.00
Risk of poverty in old age�1 SD
�0.526(0.155)
0.59*** �0.435(0.188)
0.65* �0.434(0.199)
0.65* �0.193(0.226)
0.82 �0.192(0.211)
0.83 �0.270(0.240)
0.76
Gender Gap Index �1 SD *two boys
�0.171(0.254)
0.84 �0.337(0.338)
0.71 �0.072(0.393)
0.93
Gender Gap Index �1 SD *two girls
0.356(0.255)
1.43 0.0816(0.343)
1.09 0.653(0.395)
1.92
Gender Gap Index �1 SD *two boys
�0.024(0.290)
0.98 0.0151(0.425)
1.02 0.009(0.403)
1.01
Sex
-com
po
sition
preferen
cea
nd
the
third
child
85
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Table 1 (Continued)
Entire sample Men Women
Model 1 Model 2 Model 1 Model 2 Model 1 Model 2
Variable names and values b (SE) eb b (SE) eb b (SE) eb b (SE) eb b (SE) eb b (SE) eb
Gender Gap Index �1 SD *two girls
�0.131(0.299)
0.88 �0.524(0.445)
0.59 0.099(0.415)
1.10
Risk of poverty in old age �1SD * two boys
0.001(0.249)
1.00 �0.496(0.376)
0.61 0.444(0.343)
1.56
Risk of poverty in old age �1SD * two girls
0.863(0.246)
2.37*** 0.756(0.354)
2.13* 0.978(0.349)
2.66**
Risk of poverty in old age �1SD * two boys
�0.548(0.225)
0.58* �0.991(0.319)
0.37** �0.091(0.323)
0.91
Risk of poverty in old age �1SD * two girls
0.191(0.229)
1.21 �0.037(0.326)
0.96 0.345(0.322)
1.41
Level-2 variance 0.215(0.073)
0.201(0.082)
0.091(0.049)
0.086(0.047)
0.102(0.054)
0.096(0.052)
N 3,337 3,337 1,685 1,685 1.652 1,652Log likelihood �3,449.1 �3,438.5 �1,734.2 �1,734.3 �1,705.9 �1,700.4
Notes: �pB0.10, *pB0.05, **pB0.01, ***pB0.001. Standard errors in parentheses.Source: ESS (2004/5, wave 2, excluding Turkey and Ukraine).
86
Melin
da
Mills
an
dK
atia
Beg
all
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countries where there is a low risk of falling into
poverty in old age have less of a preference for boys.
In general the results demonstrate that when
individuals live in a country where there is a higher
risk of poverty in old age, they are significantly more
likely to have a third child or to intend to have one.
To address our central research question of the
preferred sex of children, we need to turn to the
interaction effects, which are introduced in model 2
in Tables 1 and 2. For ease of interpretation, Figure 1
shows the interaction effects of poverty risk at 60
years of age and above for men and women by sex of
previous children for intention to have more chil-
dren at parity two. This graph shows support for this
hypothesis since we see that men and women who
live in countries where there is a high risk of poverty
in old age are significantly more likely to intend to
have more children when their previous two children
are girls. While hypothesis 2a receives support for
fertility intentions, particularly for men (see Table 1,
model 2), it is not supported by fertility behaviour:
sex of children does not have a significant effect on
the relationship between risk of poverty in old age
and transition to a third child.
Table 2 Determinants of transition to a third child, 24 European countries 2004/5: coefficient estimates for binary logisticmultilevel regression
Model 1 Model 2
Variable name and values b (SE) eb b (SE) eb
Sex of previous children�one boy, one girl ref 1.00 ref 1.00Sex of previous children�two boys 0.269 (0.066) 1.31*** 0.178 (0.118) 1.19Sex of previous children�two girls 0.201 (0.071) 1.22** 0.105 (0.121) 1.11Sex�male 0.221 (0.060) 1.25*** 0.219 (0.060) 1.25***Children living outside the household 0.188 (0.109) 1.21� 0.188 (0.109) 1.21�Age at first birth (centred) �0.093 (0.007) 0.91*** �0.093 (0.007) 0.91***Years of full-time education (centred) �0.026(0.009) 0.97** �0.026 (0.009) 0.97**Gender Gap Index �1 SD 0.075 (0.085) 1.08 0.149 (0.118) 1.16Gender Gap Index mean ref 1.00 ref 1.00Gender Gap Index �1 SD �0.173 (0.097) 0.84� �0.079 (0.136) 0.92Risk of poverty in old age �1 SD 0.157 (0.089) 1.17� 0.234 (0.113) 1.26*Risk of poverty in old age mean ref 1.00 ref 1.00Risk of poverty in old age �1 SD 0.078 (0.073) 1.08 0.013 (0.104) 1.01Gender Gap Index �1 SD * two boys 0.024 (0.187) 1.02Gender Gap Index �1 SD * two girls �0.104 (0.200) 0.90Gender Gap Index �1 SD * two boys �0.026 (0.226) 0.97Gender Gap Index �1 SD * two girls 0.547 (0.229) 1.73*Risk of poverty in old age �1 SD * two boys 0.211 (0.184) 1.24Risk of poverty in old age �1 SD * two girls �0.017 (0.194) 0.98Risk of poverty in old age �1 SD * two boys 0.155 (0.169) 1.17Risk of poverty in old age �1 SD * two girls 0.204 (0.175) 1.23Constant �1.095 (0.063) *** �1.088 (0.075) ***Level-2 variance 0.340 (0.048) 0.303 (0.041)N 6,502 6,502Log likelihood �3,787.4 �3,782.5
Notes: �pB0.10, *pB0.05, **pB0.01, ***pB0.001. Standard errors in parentheses, SD�standard deviation.Source: ESS (2004/5, wave 2, excluding Turkey and Ukraine).
5
–1 SD
Inte
ntio
n (d
efin
ite)
to
have
thi
rdch
ild (
per
cent
)
Poverty risk > 60 years
10
15
20
0
25
30
35
40
45
50
Mean +1 SD
Two boys Two girls Boy and girl
Figure 1 Cross-level interaction effect of poverty risk atage 60 years and over by sex of previous children forintention (definite) to have a third child at parity two, menand women, 24 European countries 2004/5Source: As for Table 1
Sex-composition preference and the third child 87
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There appears to be little support for the related
hypothesis (2b), according to which a lower risk of
falling into poverty in old age (i.e., those in the
category Risk of poverty in old age��1 SD)
creates a stronger desire to have at least one girl
(i.e., a greater likelihood of having a third child or
intending to have one for those who have two boys).
Turning again to the same interaction effects and
Figure 1, we see that in fact those who have two boys
where the risk of poverty in old age is low are less
likely to intend to have a third child than those with
a mixed-sex family. There also appears to be little
support for this hypothesis when we examine the
interaction effects of transition to third child in
Table 2. What is particularly interesting is that men
in nations that have a lower risk of falling into
poverty at older ages appear to stop having more
children once they have at least one boy and are
significantly more likely to do so if they have two
boys; they proceed to the third child only if they
have two girls. Instead of showing a waning of son
preference for countries with a lower poverty risk,
the results suggest that, at least for men, son
preference is related to economic security in old
age. There are of course alternative interpretations,
and we consider them in the discussion.
The third hypothesis was that level of gender
equity manifests itself in sex preferences for children
(H3). If gender equity is high, boys and girls should
be equally valuable, and there should be no clear
preference for the sex of children. Where gender
equity is low, however, we should expect a son
preference (i.e., a greater likelihood of having a third
child or intending to have one for parents who
already had two girls). The results are shown in
Tables 1 and 2, with Figure 2 showing the interaction
effects of gender equity by sex of previous children
for the transition to a third child.
We turn first to the interaction effects of gender
equity by sex of previous children introduced in
model 2 in Table 1, which examines fertility inten-
tions. We see that, as expected, there are no clear
significant effects or sex preferences for children in
high-gender-equity societies. This, however, also
appears to be the case for fertility intentions in low
equity societies. When we turn to the actual transi-
tion to a third child (Table 2 and Figure 2), those in
countries where gender equity is low and whose
previous children were girls, are significantly more
likely than those who have at least one boy to have a
third child. This shows that there is a son preference
in low-gender-equity societies in Europe.
When we look at the main effects of the impact of
gender equity on whether parents have a third child
or intend to have one in Europe, these analyses also
offer clear empirical support for previous theoretical
assumptions about how gender equity works in
relation to fertility (McDonald 2000, 2006). In short
the analyses show that higher gender equity results
in a significantly greater likelihood of actually
having a third child for both sexes and of intending
to have one (significantly for women).
Because we adopted a multilevel research design,
we were able to test a cross-national variation
hypothesis (H4). In view of the institutional differ-
ences and cultural norms across Europe documented
in many previous studies (e.g., Hank and Kohler
2000; Kohler et al. 2002; Andersson et al. 2006,
2007), we expected a significant difference in third-
birth intentions and transitions across Europe. The
level-2 variance estimates for actually having and
intending to have a third child shown at the bottom
of each table demonstrate that there is considerable
variation across Europe, particularly in the actual
transition to a third child. We see that, particularly
for the transition to a third child, this variance is
reduced when we control in model 2 for such
contextual factors as gender equity and risk of
poverty in old age.
Our fifth hypothesis, the homophily hypothesis
(H5), proposed that men and women desire to have
at least one child of their own sex. Turning to the
main effects of sex of previous children in Tables 1
and 2, we see that this hypothesis does not appear to
be supported for women, but that men with two girls
are more likely to intend to have a third child. The
5
–1 SD
Pro
babi
lity
thir
d bi
rth
(per
cen
t)
Gender Gap Index
10
15
20
0
25
30
35
45
50
Mean +1 SD
40
Two boys Boy and girlTwo girls
Figure 2 Cross-level interaction effect of Gender GapIndex by sex of previous children for transition to a thirdchild, men and women, 24 European countries 2004/5Source: As for Table 1
88 Melinda Mills and Katia Begall
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results of the analysis of the actual transition to a
third child, however, appear to suggest more support
for a mixed-sex family than one influenced by a
preference for children of the same sex.
Our final hypothesis, the opportunity cost hypoth-
esis (H6), proposed that women are less likely than
men to intend to have a third child irrespective of the
sex of their previous children. There is general
support for this hypothesis when we examine the first
three columns of Table 1. This model includes the
entire sample and shows that men are indeed sig-
nificantly more likely than women to intend to have a
third child when they already have two children.
When we compare the main effects of the sex of
previous children for men and women in Table 1, we
see that men always have a stronger intention than
women to have a third child if they already have
children of the same sex. Comparing the results of the
full model 2 for both sexes, we see that women appear
to be less likely or even ambivalent about intending to
have a third child when they already have children of
the same sex. Men, conversely, are more likely to
report intentions to have a third child, particularly if
they have two girls.
Discussion
To improve our understanding of parents’ prefer-
ences for the sex of their children in Western
countries, the objective of this study was to examine
whether the preference for a mixed-sex family drives
the progression (indicated by intention or actual
behaviour) to a third child in Europe. We examined
the preferences of men as well as women, and have
shown that there are clear differences between
them. By including two macro-level country vari-
ables, we were able to introduce evidence into what
had been only a theoretical discussion about the
impact of gender equity and the need for children in
old age. By examining both fertility intentions and
behaviour, we provided further evidence of simila-
rities and disparities between them in what they
indicate. Finally, we attempted to develop explana-
tions of why a preference for at least one boy, one
girl, or a mixed-sex family would exist in Western
countries.
The results confirm a clear preference for a
mixed-sex family composition in Europe, or in other
words a preference for at least one boy and one girl.
Cultural and institutional factors also appear to play
an important role in shaping preferences for the sex
of children. Parents in countries where there is a
high risk of poverty in old age and who have only
two girls in the household are significantly more
likely than others to intend to have another child.
We are aware that our interpretation assumes that
the macro-measure of old-age security captures an
individual’s perceived threat of poverty in old age. It
assumes that individuals are aware of poverty among
the elderly within their own country and that it
would remain stable. While conceding that this
indicator may in fact capture a broader range of
economic circumstances and that a more nuanced
indicator would be desirable, we hope it will have
the effect of encouraging further empirical research
into what has hitherto been simply a topic of
theoretical speculation.
Differences between men and women in prefer-
ences for the sex of children have rarely been
studied, and there is a need to develop further
explanations of why these differences exist. On this
issue, we can learn from other studies that have
investigated different outcomes, such as studies of
divorce (Morgan et al. 1988) and from recent
reviews that confirm that there are stronger prefer-
ences for the sex of children among men than among
women (e.g., Lundberg 2005; Raley and Bianchi
2006). We need to bear in mind also that the findings
of studies on this issue may be distorted by
deficiencies in the quality of men’s fertility report-
ing. As Vere (2008) warns, men potentially under-
report their past fertility, especially daughters from
previous marriages. For example, a large study using
the Panel Study of Income Dynamics found that
men were more likely to report the births of sons, yet
using a larger and more representative sample of the
Current Population Survey, Vere could not find
similar evidence. It is therefore important to have
more studies that examine the differences between
mothers and fathers in preference for the sex of
children, using other data sources, in different
countries, and across additional historical periods.
Perhaps one of the most intriguing results of our
study is that we have shown, for the first time, that
there is a strong son preference in low-gender-equity
societies in Europe. Individuals living in countries
with low gender equity are significantly more likely to
have a third child if their previous two children are
girls than if they have at least one boy. This supports
the theoretical arguments of Andersson et al. (2006),
who speculated that gender equity might be a key
factor in understanding differences in sex preferences
between countries. In fact, Andersson et al. (2006)
also found evidence of a girl preference at parity one
in Sweden, Denmark, and Norway, which might be
indicative of how parents’ preferences evolve in
relation to advances in gender equity.
Sex-composition preference and the third child 89
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By running multilevel (random-effects) models,
we were able to test whether there was significant
cross-national variation in preferences for the sex of
children in the 24 European countries. As expected
in the light of previous fertility research, there is
considerable variation across European countries in
whether parents have a third child or intend to have
one. In future research it would be important to
address the issue of how culture and cultural
differences between countries shape parents’ pre-
ferences for the sex of children.
There does not appear to be strong evidence for
the existence of a clear-cut homophily mechanism,
or in other words, a preference by parents for at least
one child of their own sex. The results appear to
suggest more support for a mixed-sex family (i.e., at
least one child of each sex). It should be noted,
however, that men are significantly more likely to
want a third child if they have two daughters than if
they already have a son. The results for women
confirmed our expectation that, possibly owing to
their higher opportunity costs of having a third child,
they are less likely to intend to have a third child,
though many eventually do so. Another explanation,
one proposed by evolutionary biologists, is that men,
particularly virile men, have a greater need to
procreate and reach higher parity. Since the biolo-
gical time span in which men are capable of
becoming a parent is longer than that of women, it
might also be the case that men are more likely than
women to believe that they will be able to have more
children (possibly with a younger partner). The
preferences of women seem to be highly ambivalent.
What issues would it be useful to consider in
future research on this subject? Studies that used
couple data could make an important contribution
to knowledge, since fertility decisions are obviously
joint decisions. Such studies would also help us to
understand why men are significantly more likely
than women to intend to have a third child. It would
also be valuable to examine the intentions and
behaviour of the same individuals over time, using
a longitudinal panel design. Finally, studies are
needed that use a measure of the need to have
children in old age that is more sophisticated than
the measure we used.
Notes
1 Melinda Mills and Katia Begall are at the University
of Groningen, Department of Sociology/ICS, Grote
Rozenstraat 31, 9712 Groningen, the Netherlands.
E-mail: [email protected]
2 The authors are grateful for comments received from
anonymous reviewers, the ISOL discussion group, and
the Netherlands ESS Symposium and Sociology Collo-
quium at the University of Utrecht. We are especially
grateful for the comments of Matthijs Kalmijn, Harry
Ganzeboom, Ritsert Jansen, Jeroen Wessie, Vincent
Buskens, Werner Raub, Beate Volker, Francesco Billari,
Arnstein Aassve, and Frank Furstenberg.
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Appendix
Table A1 Descriptive statistics for variables used in analysis of intention to have a third child
Total Men Women
Variable name Value Frequency Per cent Frequency Per cent Frequency Per cent
Intention to have a third childDefinitely not 2,092 60.48 1,037 59.53 1,055 61.44Probably not 782 22.61 437 25.09 345 20.09Do not know 134 3.87 60 3.44 74 4.31Probably yes 241 6.97 117 6.72 124 7.22Definitely yes 154 4.45 62 3.56 92 5.36Missing 56 1.62 29 1.66 27 1.57
Sex of previous childrenTwo boys 862 24.92 423 24.62 439 25.99Two girls 755 21.83 380 22.12 375 22.20Boy and girl 1,790 51.75 915 53.26 875 51.81Missing 52 1.50 24 1.38 28 1.63
Sex Male 1,742 50.36Female 1,717 49.64
House owned by member of householdYes 2,688 77.71 1,364 78.30 1,324 77.11No 755 21.83 372 21.35 383 22.31Missing 16 0.46 6 0.34 10 0.58
Children living outside the householdNo 3,307 95.61Yes 152 4.39 99 5.68 54 3.08
Risk of poverty in old age�1 SD 678 19.60 316 18.14 362 21.08Mean 1,744 50.42 904 51.89 840 48.92�1 SD 1,037 29.98 522 29.97 515 29.99
Gender Gap Index�1 SD 503 14.54 295 16.93 208 12.11Mean 2,531 73.17 1,256 72.10 1,275 74.26�1 SD 425 12.29 191 10.96 234 13.63
N 3,459 100 1,742 100.00 1,717 100Mean (SD) Missing Mean (SD) Missing Mean (SD) Missing
Years of full-time education 12.8 (3.49) 15 12.77 (3.51) 4 12.82 (3.49) 11Age 36.12 (4.97) 0 37.87 (5.01) 0 34.35 (4.26) 0
Notes: Sub-sample of only those who reached parity two and are aged 40 (women) and 45 (men) years or younger at thetime of the survey. There were 51 missing cases for the intention to have a child, and these were omitted from the analysis.Source: ESS (2004/5, wave 2, excluding Turkey and Ukraine).
Sex-composition preference and the third child 93
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Table A2 Descriptive statistics for variables used in analysis of transition to a third child
Variable name Value Frequency Per cent
Transition to third child 0 4,636 69.761 2,010 30.24
Sex of previous children Two boys 1,752 26.75Two girls 1,476 22.53Boy and girl 3,322 50.72Missing 96
Sex Male 2,616 39.41Female 4,022 60.59Missing 8
Children living outside the household No 6,201 93.3Yes 445 6.7
Risk of poverty in old age �1 SD 1,400 21.07Mean 3,393 51.05�1 SD 1,853 27.88
Gender Gap Index �1 SD 1,044 15.71Mean 4,799 72.21�1 SD 803 12.08
N 6,646 100Mean Std. dev. Missing
Years of full-time education 12.638 3.497 51Age at first birth 25.752 4.735 0
N 6,646 51
Notes: Sub-sample includes only those with parity two and higher.Source: As for Table 1.
94 Melinda Mills and Katia Begall
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Table A3 Description of macro-indicators of Gender Gap Index, and risk of poverty rate at age 60 years and over by country, 24 European countries 2005
Country N transition to thirdbirth analysis
Per cent transition tothird birth analysis
N fertility inten-tion analysis
Per cent fertility in-tention analysis
Gender GapIndex
Gender Gap Indexcategories
Risk of povertyrate�601
Risk of povertyrate�60 categories
Austria 357 5.37 175 5.06 0.6986 1 13 1Belgium 280 4.21 137 3.96 0.7078 1 20 1Switzerland 299 4.50 165 4.77 0.6997 1 18 1Czech Rep. 386 5.81 249 7.20 0.6712 1 4 �1Germany 353 5.31 184 5.32 0.7524 1 14 1Denmark 235 3.54 119 3.44 0.7462 1 14 1Estonia 252 3.79 144 4.16 0.6944 1 20 1Spain 186 2.80 120 3.47 0.7319 1 27 �1Finland 293 4.41 148 4.28 0.7958 �1 16 1France 304 4.57 152 4.39 0.6520 �1 15 1UK 304 4.57 132 3.82 0.7365 1 25 �1Greece 320 4.81 186 5.38 0.6540 �1 26 �1Hungary 220 3.31 120 3.47 0.6698 1 6 �1Ireland 385 5.79 115 3.32 0.7335 1 31 �1Iceland 122 1.84 50 1.45 0.7813 �1 8 �1Italy 179 2.69 87 2.52 0.6456 �1 21 1Luxembourg 271 4.08 139 4.02 0.6671 1 8 �1Netherlands 274 4.12 140 4.05 0.7250 1 6 �1Norway 317 4.77 148 4.28 0.7994 �1 15 1Poland 292 4.39 175 5.06 0.6802 1 8 �1Portugal 205 3.08 125 3.61 0.6922 1 26 �1Sweden 312 4.69 157 4.54 0.8133 �1 9 1Slovenia 212 3.19 128 3.70 0.6745 1 19 1Slovakia 288 4.33 164 4.74 0.6757 1 6 �1Total 6,646 100.00 3,459 100.00
1Data refer to 2005. Cut-off point 60 per cent of median equivalent income after social transfers people 60 years and over.Source: Eurostat and, for Switzerland, OECD.
Sex
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po
sition
preferen
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nd
the
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