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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: m.c.mills@rug.nl

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.

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