Working Paper Series 

Congressional Budget Office 

Washington, DC 

Social Effects, Household Time Allocation, and the 

Decline in Union Formation 

Almudena Sevilla-Sanz 

Congressional Budget Office 

Washington, DC 

E-mail: Almudena.Sevilla-Sanz@cbo.gov 

May 2005 

2005-7 

Working papers in this series are preliminary and are circulated to stimulate 

discussion and critical comment. These papers are not subject to CBO’s formal 

review and editing processes. The analysis and conclusions expressed in them are 

those of the authors and should not be interpreted as those of the Congressional 

Budget Office. References in publications should be cleared with the authors. 

Papers in this series can be obtained at www.cbo.gov/tech.cfm. 
Abstract 

Economic theories of the household and the marriage market provide potential 

explanations for differences in household formation rates over time based in part on the 

evolution of female wages. However, cross-country differences in female market human 

capital are unlikely to account for the current differences in union formation rates across 

developed countries. 

I develop a partial equilibrium model that formally incorporates social effects on a 

woman's decision to enter a household. Social effects are modeled as gender roles that 

constrain potential partners away from the efficient allocation of household labor and 

diminish the gains to forming a union. Thus, women living in less-egalitarian countries 

have, ceteris paribus, a lower union formation probability. 

I test the model using a cross-country data set at the individual level (1994 and 2002 

ISSP), which contains information on attitudes toward gender roles and the allocation of 

time to household production. The time and cross-country variation in the data, as well as 

individual reported attitudes toward gender roles, allow for the identification of the effect 

of gender roles. The empirical findings are consistent with the proposed model of social 

constraints upon the allocation of household time. 

Journal of Economic Literature Classification Numbers: D13, J0, J1, J2, Z13 

Key Words: Marriage, Union Formation, Household Labor, Social Norms 
1 Introduction1

During the past decade, low fertility in most developed countries has drawn the attention of

researchers in a variety of social science disciplines. Special focus has been given to the so

called lowest-low fertility countries, i.e., those countries, with fertility persisting well below

replacement levels (Kohler, 2002). Among these countries, Spain, Italy and Japan are the

leading examples with average total fertility rates of 1.2. Below replacement fertility presents

new economic challenges for a society, as it changes the age structure of the population and

may require structural adaptations with important implications for welfare (Weil, 1999).

These factors have become of special concern for lowest-low fertility countries, in which the

sharp decline in fertility together with a slow increase in female labor force participation

have raised questions about the viability of pay-as-you-go pension systems (Morgan, 2003).

Figure 1: Cross-Country Comparison of Crude Marriage Rates per 1,000

NoNo

No

No

No

NoNoNo

No

S

S

S S

S S S

S

S

Fin

Fin

Fin

Fin

Fin

Fin

Fin

Fin

Fin

D

D

D

D

D

D

D

D

D

4 6 8 10 12

Crude Marriage Rate

1960 1980 2000

Year

North

A

A

A

A A

A A

A

A

B B

B

B

B

B

B

B

B

FrFr

Fr

Fr

Fr

FrFr

Fr

Fr

Ge

Ge

Ge

Ge

GeGeGe

GeGe

Ir

Ir

Ir

Ir

Ir

Ir Ir

Ir

Ir

L

L

L

L

L

L

L

L L

NLNLNL

NLNL

NL

NL

NL

NL

4 6 8 10 12

1960 1980 2000

Year

West

SpSp

Sp

Sp

Sp

Sp

SpSp

I I

I

I

I

I

I

I

I

Gr

Gr

Gr

Gr

GrGr

Gr

Gr

Gr

P

P

P

P

P

P

P

P

P

4 6 8 10 12

1960 1980 2000

Year

South

J

J

J

J

US

US

US

US

US

US

US

4 6 8 10 12

1960 1980 2000

Year

Japan

Source: Council of Europe 2001, US Vital Statistics and Japan Vital Statistics

Crude Marriage Rates per 1,000

1 I would like to thank the participants at the European Econometric Society Meeting in Madrid 2004

and at the Annual Meeting of the Population Association of America in Philadelphia 2005.

1
Most of the research on lowest-low fertility has focused on the number of children in married

households.2 Figure 1 shows that in the past decades marriage rates have substantially

declined in all developed countries. While the decline in marriage has been followed by

an increase in cohabitation and out-of-wedlock fertility in most developed countries, this

situation has not occurred for lowest-low fertility countries.3 Table 1 shows that union

formation rates (either marriage or cohabitation) for women between 30 and 34 years old in

Canada, Sweden and Norway (around 90 percent) are higher than in Italy or Spain (86 and

78 percent respectively). These cross-country differences widen for younger cohorts (those

aged 25-29), who are entering a union at much smaller rates in Italy and Spain (40.7 and 53.3

percent respectively) versus 75 percent in Canada and Sweden. Moreover, the declines in

marriage rates have been compensated by increases in cohabitation in Canada, Sweden and

Norway, where cohabitation rates account for 40 percent of all unions. This circumstance

has not happened in Spain and Italy, where only 9 percent of all unions are in the form of

cohabitation. Finally, out-of-wedlock fertility is only about 5 percent for Italy and Spain,

whereas it amounts to 9.7 and 13.2 for Sweden and Norway. In sum, substantial declines

in marriage in lowest-low fertility countries, together with the fact that these countries have

not experienced the increase in out-of-wedlock fertility and cohabitation characteristic of

other developed countries, call into question previous fertility studies based solely on marital

fertility and draw new attention to household formation decisions as a contributor to fertility

phenomena.

2This approach has been justified by the lack of out-of-wedlock fertility, the high percentage of marriages

and the empirical observation that the drop in fertility was mainly due to declines in third and higher order

births within marriage (Bettio and Villa, 1998).

3See Retherford (2001) for a detailed description of the Japanese case and Frabioni and Rosina (2004)

for the Italian case.

2
Table 1: Cross-Country Comparisons of Union Formation and Out-of_wedlock Fertility

Country Cohort

Out-of-Wedlock Percentage Percent of First

Fertility**** in a Partnership Marriages***

Canada 25-29 -- 74.2 31.9

30-34 1.5 90 53.5

40-44 2.9 91.1 65

Sweden 25-29 -- 75.9 7.4

30-34 9.7 90 7.1

40-44 13.7 90.4 9.4

Norway** 25-29 -- 63.4 23.1

30-34 13.2 88.4 39.5

40-44 9.8 88.3 60.7

Spain 25-29 -- 53.3 43.3

30-34 5 86.2 75.2

40-44 2.5 88.1 87.4

Italy 25-29 -- 40.7 35.5

30-34 5.1 78.3 70.1

40-44 4.8 89.6 82.2

Source: UN Family and Fertility Surveys (1994-95)

(*) Union Formation Indicators are at age 30 for the rest (except 25 for cohort 25-29)

(**) For Norway, reference ages are 23 and 28

(***) Percentage of first unions that are marriages with no cohabitation

(****) Out-of-wedlock fertility: Percent not living in a partnership at first live birth

Union Formation Indicators*

Economic models of the household and the marriage market offer potential explanations for

the differences in household formation rates over time based on the evolution of relative

female market human capital (Becker, 1973) and (Becker, 1974). However, as shown in

section 5, differences in female education are unlikely to account for the observed differences

in union formation rates across countries. Moreover, comparative advantage or bargaining

theories of the household have also failed to explain the current patterns of household time

allocation observed in the data. In particular, the increase in female human capital that led

to an increase in female labor force participation has not, contrary to the predictions of both

theories, led to a more egalitarian division of housework within the household.

Although women continue to provide most of household production, cross-country differences

3
in the division of household work are significant. For example, based on EUROSTAT’s

reports on HETUS (the Harmonized European Time Use Surveys project), the average daily

time spent in domestic work by Swedish women is about 3 hours 42 minutes. Swedish

men spend an average of 2 hours 29 minutes. However, in the case of Spain these figures are

respectively 4 hours 45 minutes and 2 hours 8 minutes. While the average man’s contribution

to household production only differs 21 minutes between the two countries, only 70 percent

of Spanish men versus 92 percent of Swedish men engage in household activities. This

translates into Spanish women doing 1 more hour of domestic work than Swedish per day.

This paper theoretically develops a stylized partial equilibrium model of union formation in

which social effects influence individual decisions to enter a union. Social effects are modeled

as gender roles that constrain how potential partners divide the household surplus generated

through household production. The inability of potential partners to credibly commit to

make transfers of time, rather than private consumption, before the union is formed is

at the root of the argument.4 In this setting, both the non-observability by third parties

of spouse’s time devoted to household production and the absence of credible threats for

certain household production activities (especially those related to caring activities (Folbre

and Bittman, 2004)) constrain potential partners to rely on gender roles when making a

decision on how to divide the household surplus. The model predicts that gender roles result

in inefficient household time allocations, which diminish a woman’s gains to entering a union.

Accordingly, women in less egalitarian countries have, everything else being equal, a lower

probability of entering a union.5

4Time-use studies show that a substantial amount of non-market work is devoted to household production

(Apps, 2003) and (EUROSTAT, 2004). Similarly, Hersch and Stratton (Hersch and Stratton, 2002) show

that the allocation of time to household production can account for a substantial part of the wage gender

gap.

5De Laat and Sevilla-Sanz (De Laat and Sevilla-Sanz, 2004) also show that there exists a negative crosscountry relationship between fertility and non-egalitarian attitudes toward gender roles.

4
Social interactions or social norms are to a large extent enforced through nonmarket interactions and usually difficult to isolate empirically. I provide evidence for the existence of

social constraints upon the allocation of household time by showing that, after controlling

for employment status and other relevant variables, household time allocation patterns have

not substantially changed over time and closely resemble established gender roles. In the

empirical analysis, I first provide an identification strategy of social effects that relies on the

time and cross-country variation in the data, which allows for the isolation of gender roles

from other social interaction effects. A second identification strategy comes from using individual reported attitudes toward gender roles, which allows for the identification of gender

roles net of individual preferences. Empirical results using the 1994 and 2002 International

Social Survey Program (ISSP), a cross-country data set with individual level information

on demographic and economic variables, as well as attitudes toward gender roles and the

allocation of time to household production activities, support the predictions of a household

formation model with social constraints upon the allocation of household time: Whereas,

everything else being equal, more egalitarian women have lower probability of entering a

union, women living in more egalitarian countries have, everything else equal, a higher probability of entering a union.

The paper is organized as follows. Section 2 presents a brief literature review of the different

areas in economics to which the present paper contributes. Section 3 presents a stylized model

that illustrates the role of social effects on individual union formation decisions. Section 4

describes the data used in the empirical analysis and provides evidence of rigidities associated

to the allocation of time to household production. Section 5 offers identification strategies

of the effect of gender roles on individual union formation probabilities. Section 6 concludes.

5
2 Literature Review

Economic models of the household and the marriage market can provide an explanation for

the differences in household formation rates over time based on the evolution of relative

female market human capital. Unitary models of the household (Becker, 1973) predict a

decline in marriage rates after an increase in women’s labor supply (due to, for example,

an increase in female education), which reduces the gains from specialization and exchange

from a union (Bryant, 1995). Bargaining models of the household also predict declines in

marriage due to an increase in female education, which raises a woman’s reservation utility

and declines net gains to marriage (McElroy and Horney, 1981). Each theory perceives the

household in an entirely different way.6 Nevertheless, they share a focus on female market

opportunity cost that does not seem to offer a good explanation to the observed cross-country

differences in household formation rates.

Moreover, contrary to the predictions of comparative advantage or bargaining theories of the

household, a higher female human capital has not led to a more egalitarian allocation of time

within the household.7 The determinants of how spouses allocate time to household production are based on the predictions of either the unitary household production models (Becker,

1991) or bargaining models (McElroy and Horney, 1981). In the unitary framework, family

members cooperate to produce utility for all, either through the purchase of market goods

and services with earnings from market work or through household production. Specializa6On the one hand, unitary models rely on altruism in the family and do not consider conflicts of interest

between partners, which leads to the maximization of a single family utility function. On the other hand,

bargaining models focus on self-interested actors that are not entirely altruistic and, when conflict arises,

resources affect whose interests prevail.

7Time spent producing goods and services within the household has been recognized as important since

Becker (Becker, 1965). However, the time devoted to household production continues to be overlooked in

most household economic models (Juster and Stafford, 1991).

6
tion is thus efficient and the spouse with the lowest opportunity cost (i.e., the lowest human

capital or the highest home productivity) contributes the most to household production and

the least to market work.

Bargaining theories reach the same conclusion, that relative wages will affect relative contribution to housework, but the logic behind it is based on the concept of bargaining power

and threat points.8 Empirical findings contradict the prediction of both theories. For example, when a wife works more hours outside the home, she still undertakes a larger share

of housework (Ackerlof and Kranton, 2000). Similarly, men’s unpaid work increases with

his wife’s wage but only up to the point where the wife contributes as much as the husband to the household income (Bittman et al., 2001). Furthermore, this unequal division

of housework persists after observable characteristics are taken into account (Alvarez and

Miles, 2003). The literature has taken this residual as evidence for the presence of a gender

effect underlying household decisions on the allocation of time.

This paper, this gender effect is traced to an imperfect commitment process within the household, long recognized in the household economics literature (Becker, 1991) and (Lundberg

and Pollak, 1993). Recent discussions in the literature point at the inefficient allocation of

household resources that may emerge upon a couple’s inability to reach binding, legallyenforceable agreements about future behavior (Basu, 2001), (Rasul, 2002). Imperfect commitment is usually characterized as the inability of one spouse to make transfers of private

consumption to compensate the other partner for utility losses such as in Lundberg (Lundberg and Pollak, 2001) and Mazzocco (Mazzocco, 2003), and inefficiencies arise in a dynamic

8Chiappori (Chiappori, 1992) and (Browning and Chiappori, 1997) unified both set of theories into a

"collective" approach to the household, where efficiency in the household maximization problem is secured

due to spousal transfers of private consumption. However, incorporating household production into the

collective framework requires further restrictions on preferences and technology in order to identify the

sharing rule and raises questions on the dichotomization of time into leisure and market work in household

economic models (Apps and Rees, 1996) and (Apps and Rees, 1997).

7
setting once the union has taken place. The present paper differs from previous research in

that it focuses on the inability of potential spouses to credibly commit to make transfers of

time rather than money and on how the inefficiencies that may arise after the union has

been formed affect the decision to enter a union in the first place.

Finally, incorporating gender roles into an economic model of household formation contributes to the growing literature of social interactions in economics interested in how attitudes, social norms or culture influence individual economic decisions. In the last decade,

economists have investigated such interpersonal effects in a multitude of applications, including crime (Sah, 1991), unemployment (Stutzer and Lalive, 2004), investment decisions

(Duflo and Saez, 2000), welfare participation (Bertrand et al., 2000) and numerous decisions

made by teenagers on matters such as education, childbearing, gang membership and drug

use (Becker and Murphy, 2000). More recently, in the case of union formations, Loughram

(Loughram, 2002) analyzes the effect of male wage inequality on female’s marriage probabilities and Drewianka (Drewianka, 2003) exploits variations in a two-sided mate matching

market to identify the externalities associated with spousal search. This paper presents a

potential channel of how social effects (gender roles) might affect individual union formation

decisions by imposing a social constraint on the allocation of household time. Furthermore,

the use of the ISSP data provides a unique opportunity for a direct measure of these social

effects, allowing for the identification of gender roles net of individual attitudes and other

social interaction effects.

3 A Model of Union Formation and Gender Roles

This section presents a partial equilibrium model of union formation that focuses on the social

effects associated to the allocation of household time. Although omitting the male side of

the market might seem unsatisfactory from a theoretical perspective, this partial equilibrium

analysis does not invalidate the empirical results, which can be understood as the general

8
equilibrium outcome of changes in social norms and union formation probabilities.

To begin a baseline model of spousal search is presented. A union is understood as a partnership for the purpose of joint production and joint consumption. I focus on two specific

aspects of the gains to entering a union: efficiency gains from specialization in household

production and the consumption of market public goods.9 Once the surplus from the union

is defined, prospective mates need to form some notion as to whether families realize the

potential gains and how those gains are divided. Whereas this negotiation is done efficiently

under the baseline model, inefficient allocations arise in the presence of gender roles that

diminish a woman’s gains to entering a union. Thus, the model predicts that women living

in countries with less egalitarian gender roles have, with all things equal, a lower probability

of entering a union.

3.1 Baseline Model: Efficient Allocation of Household Formation

Gains

Utility after the union is formed A household is denoted by the superscript U (union)

and is assumed to be formed by two individuals a man m and a woman f. The joint household

utility depends on the consumption of two public goods: household maintenance (which is

privately produced by the household members) and a composite consumption good (which

is purchased in the market). This distinction, which bares the substitutability assumption

between time spent at household production and money, is important on theoretical and

empirical grounds. Empirically, time-use survey data shows low levels of household services

outsourcing, suggesting low substitution between time spent in household production and

9Other dimensions to marriage, such as risk pooling or consumption smoothing, are left out of the analysis

for exposition purpuses.

9
money spent in market goods.10 On the theoretical front, Apps and Rees (Apps and Rees,

1997) discuss the theoretical limitations of the perfect substitutability assumption necessary

to identify the sharing rule in Chiappori’s collective approach (Browning and Chiappori,

1997).

The composite consumption good includes market consumption goods that are jointly consumed by the household CU (such as groceries, housing, child care, etc.) and can be acquired

in the market at a normalized price p = 1. For the remainder of the paper, I will refer to CU

as the market public good. The household joint utility also depends on the production (and

consumption) of a particular public good, household chores or household public good, ZU

(these are the "commodities" in Becker’s language (Becker, 1965) such as a cleaned house

or home-made meals). It differs from CU in that it cannot be purchased in the market. ZU

is produced using both partners’ time in household production Hi for i = m, f such that:11

ZU = HU

f + HUm

with man’s and woman’s time being perfect substitutes in the production of household

services.12 Whereas the output ZU is consumed jointly by both partners, each partner

10Using Australian time-use data, Bittman shows that during the period 1984-1994 real expenditure in

outsourcing for cleaning, for instance, did not increase (Folbre and Bittman, 2004) (p.229-230). Moreover,

only 4 percent of the households bought any cleaning services during the two-week period of the survey.

Similar evidence exists for the case of Britain and the United States, where despite increases in income

inequality the demand for paid domestic services has not increased. Such low levels of outsourcing are also

found in the 1994 and 2002 ISSP in this paper.

11Consider ZU as a lower bound for the amount of household production that needs to be done in the

household, which is equal for all households. Although this assumption might seem to strong, the longitudinal

dimension of the data allows to account for permanent differences across countries with respect to the level

of household production.

12Perfect substitutability assumption is made for exposition purposes. The results are robust to general

10
privately contributes to its production. I assume that each partner derives disutility f(Hi)

from the time devoted to household production Hi for i = m, w, where f(.) is an increasing

and convex cost function. I normalize 0 ≤ Hi ≤ 1.

The household’s utility is defined as the sum of individual utilities such that:

V U = 2U(CU )+2U(ZU ) − f(HU

m) − f(HUf )

and the household’s maximization problem is:

max

Ci,Hi

2U(CU )+2U(ZU ) − f(HU

m) − f(HUf )

st.

ZU ≥ HU

m + HUf

CU = X(1 − HU

i )wi

at the optimum C∗U = wf (1 − H∗

f ) + wm(1 − H∗

m) and Z∗U = H∗

f + H∗

m, where wi i = m, w

are the wages. That is, the household consumes all the joint disposable income and produces

the needed amount of household production.

The amount of time that each partner devotes to household production Hi is given by the

first order conditions:

Hi : −2U0

(CU )wi + 2U0(ZU ) − f0(HU

i )=0

and for the case where wm > wf it is the case that HU

f > HU

m.

13

specifications of the production function.

13Under the assumption of interior solution, the second order conditions are satisfied such that:

Hi : 2U”(CU )w2

i + 2U”(ZU ) − f ”(HUi ) ≤ 0

for i = m, f.

11
Given the household’s utility and the contracted provision of household labor, each partner’s

utility in a union is given by:

V U

m = U(CU ) + U(ZU ) − f(HUm)

V U

f = U(CU ) + U(ZU ) − f(HUf )

and V U = V U

f + V U

m .

It is important to note that in this unitary household, there is a unique distribution of individual utilities. This derives from the assumption that the only private goods are essentially

the disutility of time devoted to household production. Unlike collective (Browning and

Chiappori, 1997) or non-cooperative models of the household (Lundberg and Pollak, 1993),

transfers of private consumption between partners cannot compensate for time devoted to

household production.14

Utility if single The utility for a given individual i if single is given by the following

maximization problem:

max

Ci,Hi

U(CS

i ) + U(ZS) − f(HSi )

st.

ZS ≥ HS

i

CS

i = (1 − HSi )wi

where agent’s i utility depends on market consumption goods CS

i , which can be acquired

14Apps and Rees (Apps and Rees, 1997) point out that most of the goods consumed within the household

are public goods (except for leisure) and show that the range of goods where consumption differs significantly

across individuals in the household is relatively small. Similarly, Fella et al. (Fella et al., 2004) use the nondivisibility of public consumption goods within marriage to explain the relevance of divorce laws on the

probability of divorce. In this setting, the allocation of private allocation of time, rather than private

consumption, becomes the variable of interest.

12
in the market at a normalized price p = 1. It also depends on the amount of household

production, ZS. Without loss of generality, let’s assume that ZS = αZU , for 0 <α< 1.

This assumption implies that the amount of household work that needs to be done in the

single household is less than that in the married household.

The solution to this problem is straight forward and given by:

HS

i = ZS

CS

i = (1 − ZS)wi

and the utility in the single state is denoted by:

V S

i = U(CSi ) + U(ZS) − f(ZS)

for i = m, f.

Decision to form a household In this partial equilibrium setting, a woman enters a

union if V U is greater or equal than the individual utility in the single state such that:

V U

f ≥ V Sf

This condition states that the decision to enter a union will follow if the utility each individual

gets within a joint household V U

i is enough to compensate for the loss of utility if single V Si .

3.2 Social Effects, the Allocation of Household Time and the Probability of Union Formation

This section builds on the baseline model of section 3.1 by explicitly analyzing the role of

social effects characterized as the prevailing gender roles in a given country. Gender roles act

as a social constraint that effectively prevents potential partners to perfectly contract upon

13
the efficient division of household labor before the union is formed.15 Thus, the amount of

time each partner devotes to household production Hm and Hf is the value dictated by the

existing gender roles.

The household maximization problem becomes:

max

Ci,Hi

2U(CU )+2U(ZU ) − f(HU

m) − f(HUf )

st.

ZU ≥ HU

m + HUf

CU = X(1 − HU

i )wi

HU

m ≤ HU

m

where the constraint is characterized by the fact that the male partner cannot increase his

time to household production above what is imperative by gender roles HU

m (or symmetrically,

that a woman cannot decrease the amount of time devoted to household production below

what is prescribed by the existing social norms HU

f ).16 Under this scenario the solution

is straightforward. Given that ZU is fixed and ZU = HU

m + HU

f , HUf = HU

f and CU = X(1 − HU

i )wi.

Gender roles affect the probability of entering a union in two ways: First, they diminish the

household utility and second, it alters the distribution of household surplus. On the one hand

household utility is lower under gender roles because partners are constrained from reaching

15The model specification, where the only private goods are the disutility from time devoted to household

production, implies that it is not possible to compensate a woman for having a "socially constrained" partner.

16 In a context of high relative female wages as the one analyzed here, it is reasonable to assume that

the social norm is always binding. For the remeinder of the paper I consider HU

m ≥ H, i.e. the efficient

amount of a man’s time to household production is above the value that is prescribed by the social norm.

Intertemporal cross-country differences in union formation probabilities can also be explained in light of this

model, assuming that the constraint becomes effective for higher female wages.

14
optimal time allocations. This situation is true despite increases in the consumption of the

market public good.

17

CU

Gender Roles = (1 − HUm)wm + (1 − HUf )wf > CU

Baseline = (1 − HUm)wm + (1 − HUf )wf

On the other hand, social constraints also alter the distribution of the surplus. In particular,

under strong gender norms in the way characterized above, a man has no incentive to deviate

from what is prescribed by gender roles once the union is formed. He not only enjoys a higher

level of market consumption good but also a lower disutility from household labor.

V U

m = U(CU

) + U(ZU ) − f(HU

m) < V U

m = U(CU ) + U(ZU ) − f(HUm)

Proposition 1 Under gender roles that constrain the household allocation of time, a man’s

utility from entering a union rises.

A woman is, however, worse off as a higher consumption of the market public good is not

enough to compensate her for a higher amount of time devoted to household production.18

On the one hand, she gets higher consumption, but on the other hand she has to allocate

more time to household production, which leads to lower utility within marriage.

V U

f = U(CU

) + U(ZU ) − f(HU

f ) < V U

f = U(CU ) + U(ZU ) − f(HUf )

Proposition 2 Under gender roles that constrain the household allocation of time, a woman’s

utility from entering a union falls.

17Under social constraints, the man devotes less time to household production and more to market labor.

Given that the man’s wage is higher, the consumption of market public good increases.

18Given that the household utility decreases in the presence of gender roles, and that man’s utility increases,

woman’s utility must necessarily decrease.

15
Finally, the probability of a woman entering a union under gender roles goes down:

p(V U

f ≥ V Sf ) > p(V U

f ≥ V S

f ) ⇐⇒ p(V Uf ) > p(V U

f )

Also, gender roles that prescribe lower (higher) levels of HU

m (HUf ) lower (increase) a woman’s

probability of entering a union:

p(V U

1,f ≥ V S

f ) > p(V U

2,f ≥ V S

f ) ⇐⇒ p(V U

1,f ) > p(V U2,f )

for V U

i,f = U(CUi ) + U(ZUi ) − f(HUi,f ) for i = 1, 2 and HU1,f > HU2,f .

Proposition 3 Stronger gender roles, which constrain household time allocation to lower

(higher) levels of HU

m (HUf ), lower a woman’s probability of entering a union

From Proposition 3, some women (those with a lower taste for household production or with

lower taste toward forming a union) might be better off not entering a union altogether.

Everything else equal, the probability of entering a union diminishes with the probability of

facing a stronger constraint. That is, women living in less egalitarian countries where gender

norms are well entrenched would have, with all things equal, a lower probability of entering

a union.

4 The Data: International Social Survey Program: Family and Changing Gender Roles (ISSP 1994 & 2002)

The data used for the empirical analysis is drawn from the International Social Survey

Program (ISSP), which is an annual program of cross-national collaboration on surveys

between several social science institutes dating to 1983.19 These data sets offer a unique

19The ISSP is a cross-country effort in which each member state individually collects data and (1) jointly

develops topical modules dealing with important areas of social science, (2) carries a module of a 15-minute

16
opportunity for cross-country analyses in topics such as social inequality, social networks ,

and the role of government, as they coordinate national social science surveys to produce a

common set of questions asked in identical form in the participating nations.20

This analysis is based on a pooled cross-section of the 1994 and 2002 surveys analyzing

"Family and Changing Gender Roles.” In each of the participating countries, a male or

female adult older than 16 or 18 years (depending on the country) from the selected household

was administered (almost) the same questionnaire across all countries. Although the same

individuals are not surveyed in both years, the data allows the researcher to take into account

country and survey-year fixed effects, which are useful for the identification of social effects.

This survey is particularly useful for the purpose of this analysis because it collected general

demographic information, employment and wages, the actual division of labor within the

household and several attitudinal variables toward gender roles.

I restrict the sample to those countries who undertook the survey in both years.21 Table

2 reports the list of countries used in this analysis for both, men and women between the

ages of 18 and 70. It includes households from Australia, Austria, Germany (West), Great

Britain and Northern Ireland, Ireland, Japan, Netherlands, New Zealand, Norway, Spain,

Sweden, and the United States.

A main contribution of this survey is that it asks questions about gender roles. I construct

self-completion supplement to their regular national surveys, (3) includes a common core of background

variables, and (4) makes the data available to the social science community as soon as possible. The number

of member states is currently 39 (although not all members have participated since 1983).

20Examples of the use of the ISSP are Blau and Kahn (Blau and Kahn, 1992) and Albrech et al. (Albrecht

et al., 2000).

21Some Eastern European and developing countries were also part of the ISSP survey. However, due to the

differences in economic systems and demographic processes, this analysis concentrates on western developed

countries.

17
an egalitarian index from a series of eight responses that report the level of agreement with

statements designed to capture attitudes towards the gender division of household labor.22

The responses are coded on a 1 to 5 scale, from ”strongly agree” to ”strongly disagree.” Table

2 shows the average value of the principal component index in each country by survey year.

Countries are ordered from higher to lower values of this index according to 1994 values,

with a higher index meaning more egalitarian. Respondents from Austria, Italy, Japan and

Spain seem to hold attitudes toward the gender division of household labor that are less

egalitarian than those of respondents in countries such as Norway, Sweden, and Canada.

There is not a substantial change with respect to the ranking of countries according to this

index in the two survey periods. There is, however, a clear trend toward more egalitarian

attitudes across countries, as the value of the egalitarian index is higher in 2002 than in

1994. It is also noticeable that among the countries with less egalitarian gender norms are

those with lowest-low fertility and union formation rates, such as Japan, Italy and Spain.

Given that the object of interest is household formation and not marriage per se, I define an

ever-married woman as a woman who is either married or has ever been married (i.e., divorced

or widowed). I also include all women who are currently living with a partner in a long lasting

relationship. I construct this variable using the answers to the questions on marital status

22The egalitarian index consists of a standarized principal component index that is calculated using a

principal component factor analysis. This analysis finds a number of common factors that linearly predicts

the covariance matrix of the responses to the following statements:

"A working mother can establish just as warm and secure a relationship with her children as a mother

who does not work."

"A pre-school child is likely to suffer if his or her mother works."

"All in all, family life suffers when the woman has a full-time job."

"A job is alright, but what most women really want is a home and children."

"Being a housewife is just as fulfilling as working for pay."

"Having a job is the best way for a woman to be an independent person."

"Both the man and woman should contribute to the household income."

"A man’s job is to earn money; a woman’s job is to look after the home and family."

18
and steady life partnership (i.e., whether living as married with a partner). Table 2 reports

summary statistics for these two variables. The first thing to observe is that the percentage

of women ever to cohabit is much higher for those countries with more egalitarian attitudes.

This difference is also more pronounced in 2002 than in 1994. Secondly, partnership rates

seem to increase as we move toward countries with more egalitarian attitudes.

Table 2: Summary Statistics by Country and Survey Year

Gender Roles

Country Principal Yrs. Of Yrs. Of 

(From more % Ever % Ever in Component Schooling Schooling

to less egalitarian) Year Married Partnership Index (males) (female) Obs.

Sweden 1994 0.76 0.76 0.33 11.51 11.50 1177

2002 0.54 0.80 0.54 11.93 12.50 974

Norway 1994 0.66 0.79 0.12 13.22 12.60 1,910

2002 0.70 0.85 0.41 13.50 13.22 1,368

Great Britain 1994 0.81 0.81 0.08 11.46 11.38 867

2002 0.76 0.76 0.08 12.28 12.21 1,672

US 1994 0.77 0.77 0.03 13.46 13.38 1,270

2002 0.74 0.78 0.13 13.59 13.51 1,063

Northern Ireland 1994 0.75 0.75 0.02 11.50 11.33 545

2002 0.72 0.72 0.02 --- --- 845

Netherlands 1994 0.72 0.72 -0.03 12.95 12.10 1,743

2002 0.67 0.78 0.06 14.30 13.19 1,110

New Zealand 1994 0.84 0.84 -0.11 12.03 11.90 932

2002 0.80 0.87 -0.10 13.31 13.19 888

Ireland 1994 0.71 0.71 -0.16 11.09 11.58 866

2002 0.68 0.71 0.08 12.34 12.58 1094

Spain 1994 0.68 0.68 -0.18 --- --- 2,259

2002 0.66 0.71 0.02 11.90 11.50 2,170

Japan 1994 0.79 0.79 -0.18 12.21 11.67 1,143

2002 0.81 0.81 -0.01 13.05 12.34 948

Australia 1994 0.86 0.86 -0.22 11.52 11.30 1,560

2002 0.81 0.86 -0.08 12.22 12.17 1,239

Germany 1994 0.75 0.85 -0.23 --- --- 2,129

2002 0.71 0.86 0.14 11.38 11.00 835

Austria 1994 0.82 0.82 -0.37 12.45 11.18 852

2002 0.71 0.79 -0.05 --- --- 1770.00

Overall 1994 0.75 0.77 0.09 12.18 11.80 19,549

2002 0.72 0.79 0.11 12.57 12.42 21,948

Female education is used as a measure of market female human capital and potential outside

opportunities. Education is given in years and levels, but data on levels applies different

19
criteria among different countries and it is hard to interpret. Instead, I use completed years

of schooling. I lose information on Austria, Germany, Northern Ireland, and Spain, for one

survey year because these countries have no information on this variable.23 As mentioned in

the introduction, differences in education across countries are not significant. Especially the

woman-to-man education ratio is small for all countries in the two survey years.

4.1 The Allocation of Time to Household Production

The data also contains information on the share of household production that each partner

undertakes. Questions on household production include who does the laundry, who shops

for groceries, who prepares meals, and who cares for the sick. Answers to the question "who

does what" are tabulated in 6 different categories: The first two is always and usually the

woman, the third is both and the fourth and fifth are usually and always the man. A sixth

category is a third person. I drop these observations, which amount to less than 2% of the

sample.24 Table 3 shows the share of housework performed by each partner in each country

during the two years the survey was implemented.25 Countries are ordered from more to

less egalitarian according to their 1994 egalitarian index. Table 3 shows that, first, more

egalitarian countries tend to have a higher participation of men in household activities and

second, that the spousal distribution of housework share has not changed much over the two

survey years.

23A dummy is included in all regressions in order to control for missing education in these countries, which

allows to use the remainder relevant information for these countries.

24Low levels of housework outsourcing are consistent with more detailed time-use surveys.

25Section 5 provides further evidence that the division of housework has changed little both, across countries and over time.

20
Table 3: The Division of Household Production

Country Usually/ Both Usually/ Out- Obs. Usually/ Both Usually/ Out- Obs.

Survey Always Always sourcing Always Always sourcing

Year woman man woman man

SW

1994 0.80 0.17 0.03 0.00 831 0.41 0.57 0.02 0.01 773

2002 0.73 0.21 0.07 0.00 731 0.40 0.56 0.04 0.00 649

NO

1994 0.82 0.16 0.02 0.00 1,365 0.44 0.55 0.01 0.01 1,232

2002 0.80 0.16 0.04 0.00 1,033 0.43 0.55 0.02 0.02 918

GB

1994 0.81 0.18 0.01 0.01 547 0.50 0.49 0.01 0.01 520

2002 0.79 0.16 0.05 0.01 998 0.55 0.42 0.03 0.01 885

US

1994 0.68 0.27 0.05 0.02 718 0.48 0.50 0.02 0.01 708

2002 0.62 0.28 0.10 0.02 629 0.49 0.45 0.06 0.01 608

NI

1994 0.90 0.09 0.01 0.02 310 0.53 0.46 0.01 0.01 299

2002 0.87 0.10 0.03 0.02 437 0.60 0.36 0.04 0.01 421

NL

1994 0.88 0.10 0.01 0.00 1,162 0.49 0.50 0.02 0.00 1,132

2002 0.85 0.11 0.04 0.00 748 0.50 0.47 0.02 0.00 721

NZ

1994 0.75 0.22 0.04 0.00 705 0.59 0.40 0.01 0.00 682

2002 0.77 0.19 0.04 0.00 647 0.59 0.38 0.03 0.00 602

IR

1994 0.88 0.12 0.00 0.00 555 0.52 0.47 0.01 0.00 550

2002 0.87 0.10 0.03 0.01 679 0.56 0.40 0.03 0.01 646

SP

1994 0.91 0.08 0.01 0.02 1,416 0.61 0.38 0.01 0.00 1,402

2002 0.83 0.14 0.03 0.01 1,332 0.53 0.43 0.04 0.01 1,271

JA

1994 0.96 0.04 0.01 0.02 827 0.79 0.20 0.01 0.02 790

2002 0.91 0.07 0.02 0.03 689 0.73 0.23 0.04 0.02 630

AUS

1994 0.79 0.16 0.05 0.01 1,415 0.62 0.35 0.03 0.01 1,336

2002 0.75 0.14 0.11 0.03 1,128 0.63 0.31 0.06 0.04 1,065

GE

1994 0.91 0.08 0.01 0.02 1,472 0.57 0.42 0.01 0.01 1,341

2002 0.89 0.09 0.03 0.02 568 0.57 0.40 0.03 0.01 499

OS

1994 0.94 0.04 0.02 0.01 590 0.62 0.37 0.02 0.01 556

2002 0.90 0.07 0.03 0.02 1,082 0.58 0.39 0.03 0.02 937

Laundry Sick Caring

21
Table 3 (cont): The Division of Household Production

Grocery Shopping Meal Preparation

Country Usually/ Both Usually/ Out- Obs. Usually/ Both Usually/ Out- Obs.

Survey Always Always sourcing Always Always sourcing

Year woman man woman man

SW

1994 0.42 0.49 0.09 0.00 830 0.56 0.42 0.03 0.00 829

2002 0.38 0.49 0.13 0.00 732 0.60 0.29 0.11 0.00 732

NO

1994 0.40 0.50 0.10 0.00 1,366 0.53 0.42 0.05 0.00 1,366

2002 0.43 0.48 0.10 0.00 1,035 0.59 0.31 0.09 0.00 1,034

GB

1994 0.43 0.53 0.05 0.00 554 0.59 0.36 0.05 0.00 553

2002 0.46 0.46 0.08 0.00 997 0.59 0.31 0.11 0.01 1,000

US

1994 0.49 0.45 0.06 0.00 729 0.54 0.40 0.06 0.00 726

2002 0.48 0.42 0.10 0.00 642 0.55 0.31 0.13 0.00 639

NI

1994 0.59 0.38 0.03 0.00 315 0.65 0.33 0.02 0.01 314

2002 0.62 0.34 0.04 0.00 442 0.64 0.31 0.05 0.01 442

NL

1994 0.58 0.34 0.08 0.00 1,163 0.66 0.31 0.03 0.00 1,163

2002 0.58 0.30 0.10 0.00 748 0.66 0.24 0.09 0.00 747

NZ

1994 0.60 0.36 0.04 0.00 707 0.70 0.26 0.04 0.00 706

2002 0.58 0.35 0.07 0.00 647 0.65 0.27 0.08 0.00 648

IR

1994 0.67 0.29 0.04 0.00 554 0.75 0.23 0.02 0.00 555

2002 0.63 0.29 0.08 0.01 681 0.71 0.23 0.06 0.01 682

SP

1994 0.68 0.29 0.03 0.01 1,421 0.81 0.18 0.01 0.01 1,431

2002 0.51 0.42 0.07 0.00 1,346 0.75 0.19 0.06 0.01 1,337

JA

1994 0.80 0.20 0.01 0.03 824 0.94 0.05 0.01 0.03 821

2002 0.76 0.22 0.02 0.01 700 0.91 0.08 0.02 0.03 691

AUS

1994 0.60 0.31 0.09 0.00 1,391 0.71 0.23 0.06 0.00 1,412

2002 0.61 0.25 0.14 0.02 1,113 0.68 0.19 0.13 0.03 1,112

GE

1994 0.49 0.44 0.06 0.01 1,489 0.55 0.43 0.02 0.01 1,485

2002 0.51 0.41 0.08 0.01 573 0.76 0.18 0.06 0.00 573

OS

1994 0.52 0.39 0.09 0.01 592 0.58 0.40 0.03 0.00 589

2002 0.46 0.46 0.09 0.00 1,101 0.77 0.17 0.06 0.01 1,096

22
4.2 Household Time Allocation and the Persistence of Gender

Roles

As mentioned in the introduction, despite the increase in relative female human capital (and

labor force participation), men’s time to household production remains low and has not

substantially increased over time. This empirical regularity has been argued to be inconsistent with household models based on either specialization or bargaining theories and has

suggested the existence of a gender effect associated to household production. Based on the

1994 and 2002 ISSP data sets, Table 4 provides further evidence for the existence of rigidities

associated to the allocation of time to household production and thus, for the hypothesized

gender effect. The data does not contain information on the time that each partner devotes

to household production, but it does contain information on each partner’s share of total

time. Table 4 shows the predicted probabilities for the allocation of time to several household

production activities for a representative cohort of women between 30 and 40 years old.26

26See Appendix A for estimates from the ordered logit regression. The predicted probabilities presented

here are derived from the regressions in the second column of Table 6.

23
Table 4: Estimated Ordered Logit Probabilities of the Division of Household Labor

not working working not working working not working working not working working

1994 Always woman 0.718 0.512 0.4 0.256 0.399 0.272 0.491 0.282

Usually woman 0.207 0.316 0.313 0.308 0.293 0.288 0.319 0.35

Both 0.06 0.134 0.273 0.411 0.267 0.369 0.161 0.3

Usually man 0.008 0.021 0.007 0.014 0.028 0.049 0.019 0.045

Always man 0.005 0.014 0.005 0.01 0.011 0.02 0.007 0.02

2002 Always woman 0.651 0.435 0.399 0.261 0.359 0.257 0.498 0.296

Usually woman 0.251 0.354 0.318 0.315 0.302 0.292 0.322 0.364

Both 0.078 0.164 0.27 0.4 0.292 0.379 0.153 0.281

Usually man 0.011 0.026 0.007 0.013 0.032 0.05 0.017 0.04

Always man 0.007 0.018 0.005 0.009 0.012 0.02 0.007 0.017

Sample: Women between 30 and 40 years old 

In 2002: 2,940. In 1994: 2,297

Estimated probabilities from ordered logit regressions

Woman's work status Woman's work status Woman's work status Woman's work status

Laundry Cares for the sick Grocery Shopping Meal preparation

Column 1 analyzes the case where a woman is not working (and only the male partner

is), and column 2 analyzes the case where both partners work. Several interesting results

are worth mentioning. First, working women are less likely than non-working women to

do most of the household production and are more likely to share this activities with her

partner. Similarly, men are more likely to contribute when the woman is working. Second,

differences between genders are significantly apart with women contributing most of the

time to household production. Third, although the probability that a man contributes to

household production activities has increased from 1994 to 2002, it remains significantly small

at around 4 percent if the woman works, and 2 percent if she does not. In conclusion, women

have gone from always doing household activities to usually. For example, for the case of

doing laundry the probability that both partners contribute has increased only 3 percent

if the woman works and close to nothing if she does not (although the probability that a

woman contributes always to this activity has gone from .71 to .65 in the case that she works

and from .51 to .43 if she does not). Finally, the probability that a woman always or usually

does household production activities remains very high at about 40 percent (compared to 4

percent and 2 percent for men). In conclusion, all these findings suggest a social constraint

24
upon the allocation of family time characterized in this paper as the imperative gender roles.

5 Empirical Analysis

The empirical analysis in this section tests the implications for individual behavior that follow

from the partial equilibrium model and evaluates the effect of gender roles on a woman’s

probability of entering a union. Gender roles are captured by the egalitarian index reported

in Table 3 by the principal component index.27 Higher (more egalitarian) values of this index

represent a lower constraint upon the division of household production.

5.1 Empirical Specification of Social Constraints and Household

Formation Decisions

The theoretical model predicts that the mechanism through which gender roles affect a

woman’s probability of entering a union is by constraining the allocation of household time,

which reduces the gains from forming a union. In particular, proposition 3 showed that a

woman who lives in a country with more egalitarian gender roles has, everything else equal, a

higher probability of entering a union than a woman in a less egalitarian country.28 Formally,

the model states that the decision rule to form a union for woman i in country k is given by:

form a union if V U

i,k ≥ V Si,k + εi,k for i = m, f

27The estimates are robust to different specifications of gender roles. A sensitivity analysis is presented in

Table 7 of Appendix B.

28Throughout the analysis, it is assumed that the country’s attitudes toward gender roles are the implicit

reference group that affects the ability of potential partners to efficiently divide the marital surplus. This

assumption avoids self-selection problems given that mobility across countries does not take place.

25
where εi,k is an individual taste parameter toward remaining single. This normally distributed error is independently distributed across countries but correlated within countries for

i = m, f (male or female) and k = 1...13.

29

In effect, the probability of a union may be written as a function of observable individual

and "marriage market" characteristics, which include gender roles:

pi,k = p(V U

i,k − V Si,k ≥ εi,k) =

= p[f(EDUi,k, AGEi,k, GRk) ≥ εi]

the reduced form econometric specification is:

pi,t,k = Xi,t,kβ1 + Et,kβ2 + θt + ωk + +εi,k

where pi,t,k is the probability of entering a union for woman i in survey year t, and country

k. Xi,t,k is a set of individual observable characteristics (education, age and labor force

participation) and θt and ωk are the year and country fixed effects respectively. The error

term captures the taste for marriage that is assume to follow a normal distribution with

variance σk the same for all women in country k.

5.2 Identification of Gender Roles (I)

The identification of social effects on individual union formation probabilities comes from the

differential changes over the survey period and across countries of the variable Et,k, which is

the principal component index that captures the level of gender roles (or social constraint) at

survey year t in country k. Higher values of this variable are associated with more egalitarian

gender roles, and therefore with a lower constraint upon the division of household production

between partners.

29 I choose a relatively large cell size, the woman’s country in a particular survey year, to minimize measurement error in my estimates of marriage-market specific social constraints.

26
Notice that ωk, the country fixed effect, accounts for permanent differences across countries

over the survey period that may systematically affect the probability of women entering a

union in some countries. For example, permanent differences in preferences toward forming a

union, so that everything else equal, women living in some countries are more prone to enter

a union than in others, would be captured by the country fixed effect. Similarly, systematic

differences in preferences toward household production across countries, which would affect

the probability of entering a union, are also captured by the country fixed effect.30 By the

same underlying principle, the survey year fixed effect θt accounts for any common change

to all countries over the survey years. For example, a change in attitudes toward forming

a union that would have similarly occurred in all countries during the survey period, would

be captured by the survey year fixed effect.31 As a result, so long as the country fixed

effect does not vary over the survey period and the survey year fixed effect does not vary

across-countries, this approach yields an unbiased estimate of β2.

32

Column 1 in Table 5 shows the results for this regression, for a sample of women between 20

30For example, if Spaniards preferred (on average) a cleaner household (and therefore had a higher preference for more household production), they would also have a lower probability of entering a union (as they

would require, ceteris paribus, a higher union formation gain than say, Swedes). This effect would however

be captured by the country fixed effect.

31The survey also contains information on attitudes toward marriage and cohabitation. Appendices C

and D present the answers to these questions. In general, more egalitarian countries are more receptive to

cohabitation whereas less egalitarian countries seem to place more value on marriage. This is captured in the

regression by ωk. Moreover, these attitudes change consistenly in all countries over the two survey periods

(in all countries cohabitation has become more acceptable and marriage has become less). This is captured

in the regression by θt. Thus, β2 is not likely to capture a differential change in attitudes toward union

formations.

32Notice that this aproach is similar to a difference in difference approach, where the treatment is a

continuous rather than a discrete variable (mainly, the degree of gender roles in a given country).

27
and 40 years old that are not currently studying.33 The reported coefficients from the probit

models presented here are the change in the probability for an infinitesimal change in each

independent, continuous variable and, by default, the discrete change in the probability for

dummy variables. The coefficients of the woman’s individual variables have the usual sign.

The higher the woman’s education, the lower the probability of entering a union, which is

consistent with the specialization theory of the household proposed in Section 3. A one-year

increase in schooling decreases a woman’s probability of entering a union by around 1.5

percentage points. The small magnitude of this coefficient, together with the fact that the

cross-country educational differences are not significant, further supports the argument that

relative female education is not the appropriate variable to explain cross-country differences

in union formation. The coefficient on age is positive and significant as expected. There

is also a significant change on the probability of entering a union in the survey-years. The

coefficient of survey year is negative and significant, which shows a decreasing trend in the

probability of entering the union over this period.

33Regressions using a sample of women ages 20-65 produced qualitatively similar results (see B), although

the magnitude of the correlation between gender roles and union formation probabilities is lower. This is

not surprising because the vast majority of women aged 45-65 in my sample have been in a union for many

years and so are no longer active participants in the marriage market.

28
Table 5: Gender Roles and a Woman's Union Formation Probability

Woman's Probabiltiy of Entering a Union No individual With Individual 

(Maximum-likelihood Probit Estimation) Effects Effects

Completed years of schooling -0.016 -0.013

[-8.31]*** [-6.65]***

Age 0.025 0.024

[34.44]*** [32.19]***

Survey Year (=1 if 2002) -0.037 -0.045

[-1.91]* [-2.23]**

Gender Roles by country and survey year 0.203 0.314

(higher values mean more "egalitarian") [1.74]* [2.55]**

Woman's individual attitudes toward gender roles -0.039

(higher values mean more "egalitarian") [-7.65]***

Observations 8,914 7,887

(1) Robust z statistics in brackets

(2) * significant at 10%; ** significant at 5%; *** significant at 1%

(3) Sample: Women between 20 and 45 years old

(4) Coefficients report the change in the probability of entering a union for an infinitesimal 

change in each independent (continuous) variable and , by default, the discrete

change in the probability for dummy variables. 

The coefficient of interest β2 is the coefficient corresponding to the country-year variable

gender roles by country and survey year (0.203). This variable is the egalitarian index

that measures average gender roles in a country in a given survey year. This coefficient is

statistically significant and has the predicted positive sign, i.e., a woman living in a more

egalitarian country has a higher probability of entering a union. In particular, a standard

deviation (one unit) increase in the egalitarian index leads to a 0.2 increase in a woman’s

probability of entering a union. The average Spanish woman, with an index of 0.02 in

2002 had about 10 percentage points lower probability of entering a union than her Swedish

counterpart, living in a country with an egalitarian index of .54. This value is just in between

the 5 to 25 percentage points reported in Table 5 for the 30 to 34 cohort.

29
5.3 Identification of Gender Roles (II)

I have argued that more egalitarian attitudes toward gender roles in a country positively

affect individual choices, in particular individual union formation probability by imposing

a social constraint that prevents potential partners from dividing the marital surplus in

an efficient manner. This results in a lower surplus for a woman, which diminishes her

probability of entering a union.

A potential problem for the previous empirical results is that average attitudes in a country

are, by definition, correlated to individual attitudes. If individual attitudes toward gender

roles were associated with a higher probability of union formation (i.e., if being a more

egalitarian woman was associated with a higher taste for forming a union) then omitting

individual attitudes would result in a bias of the β2 coefficient. The remainder of this

section takes advantage of the specific features in the ISSP data set and use subjective data,

as proposed in Manski (Manski, 1993) and (Manski, 2000), in order to identify the social

effects from individual attitudes.

Column 2 of Table 5 shows that incorporating individual attitudes toward gender roles in

the regression presented in column 1 of Table 5 does not change the sign of the coefficient on

gender roles by country and survey year. Furthermore, individual and average country gender

roles seem to go in opposite directions: Whereas, ceteris paribus, more egalitarian women

have lower probability of entering a union, women living in more egalitarian countries have,

everything else equal, a higher probability of entering a union. These two results further

support the existence of social effects on woman’s individual probability of entering a union.

The negative sign on the individual attitudes toward gender roles (−0.039) is consistent with

the theoretical model presented in Section 3, as more egalitarian women can be thought of as

having a higher disutility from the time allocated to household production. This negative sign

means that a woman’s individual attitudes toward gender roles significantly and negatively

30
affect the probability of entering a union.34 The coefficient on the social effect (0.314) on

the other hand continues to be significant and positive (with a slight increase in value as

expected). This finding is also consistent with the model presented in Section 3, as women

living in less egalitarian countries need, everything else equal, a higher share of the surplus

when entering a union.

6 Conclusion

The study of below replacement fertility characteristic of developed economies has traditionally overlooked union formation processes. Nonetheless, cross-country differences in union

formation rates are significant, especially among younger cohorts. Both declines in marriage rates and increases in cohabitation rates have followed very different trends across the

developed world. In particular, the so-called lowest-low fertility countries, like Italy, Spain

or Japan, have experienced a decline in marriage rates that have not been accompanied

by increases in cohabitation (and out-of wedlock fertility) rates characteristic of the rest of

the developed world. It becomes thus increasingly important to look at union formation

processes for the study of fertility.

Economic theories of the household and the marriage market provide an explanation for

the differences in household formation rates over time based on the evolution of female

wages. However, cross-country differences in female market human capital are unlikely to

account for the divergence in existing union formation rates across developed countries.

This paper empirically and theoretically explores the social effects associated to household

formation decisions in order to shed light onto the factors that contribute to differences in

the probabilities of entering a union across countries.

34The coefficient on individual attitudes must be interpreted with caution. Only under the assumption

that individual attitudes toward gender roles are not just reflecting attitudes toward union formation can

this coefficient be interpreted causally.

31
A stylized partial equilibrium model of union formation is developed that formally analyzes

the channel through which social effects may affect a woman’s decision to entering a union.

Social effects are rationalized as gender roles that constrain how potential partners divide

the gains from entering a union. The inability of potential partners to credibly commit to

make transfers of time, rather than private consumption, before the union is formed is at

the root of the argument. In this setting, both the non observability by third parties of

spouse’s time devoted to household production and the inexistence of credible threats for

certain household production activities leads potential partners to rely on gender roles when

making a decision on how to divide the household surplus. The model predicts that gender

roles result in inefficient household time allocations, which diminishes a woman’s gains to

entering a union. Thus, women in less egalitarian countries have, with all things equal, a

lower probability of entering a union.

Empirical results from individual level cross-country and longitudinal data support the predictions of a household formation model with social constraints upon the allocation of household time. I first provide evidence for the existence of social constraints upon de allocation of

household time by showing that after controlling for employment status and other relevant

variables household time allocation patterns have not substantially changed over time. I then

provide an identification strategy of social effects which exploits the time and cross-country

variation in the data as well as individual reported attitudes toward gender roles. This fact

allows for the identification of gender roles net of individual attitudes and other social interactions. In conclusion, imperfect commitment processes within the household that constrain

the allocation of household time to what is prescribed by gender roles and the empirical

results presented in this paper might provide the theoretical and empirical foundations for

the design of policies that encourage men’s participation in family work.

32
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37
A Ordered Probit Model Results: Allocation of Time

to Household Production

Table 6 shows the ordered logit used for constructing table 4. The sample is restricted

to those who are married or living as married and have finished school. The dependent

variables are "who does the laundry," "who takes care for the sick," "who does the grocery

shopping" and "who prepares meals." The dependent variables go from 1 to 5, 1 meaning

always woman, and 5 meaning always the man. I control for survey year, country, gender

of the respondent, work status of the respondent and the spouse and an interaction term to

control for whether the work status is related to the man or the woman.

We see that the variable "survey year" has a positive coefficient. This variable takes value 1

if the survey year is 2002, which means that there has been an increase in the probability for

higher values of the dependent variable, i.e., men are more likely (or women less likely) to

take on household production activities. When evaluating this probabilities at mean values

as in 4, we see that this has only happened for low values of the dependent variable (i.e., the

man partner continues to allocate very little time to household production).

Notice that there is also a disconnect between what men report and what women report.

Men tend to report higher values of the dependent variables than women (i.e., the dummy

that controls for the man being the respondent has a positive and significant coefficient).

38
Table 6: Ordered Logit Estimates of the Household Division of Labor

Ordered Logit

Survey Year 0.18 0.16 0.05 0.04 0.096 0.085 0.065 0.05

(5.99)*** (5.54)*** (1.81)* 1.3 (3.38)*** (2.97)*** (2.26)** (1.75)**

Man respondent 0.58 0.78 0.73 0.91 0.58 0.8 0.56 0.764

(22.92)*** (16.19)*** (29.18)*** (19.26)*** (24.40)*** (17.54)*** (23.38)*** (17.01)***

Respondent Working 0.84 0.62 0.46 0.827

(21.00)*** (16.9)*** (12.72)*** (22.57)***

Spouse Working -0.135 -0.125 -0.25 -0.153

(3.13)*** (3.08)*** (6.52)*** (3.85)***

Respondent Working -0.84 -0.69 -0.8 -0.877

(if man) (14.56)*** (12.25)*** (14.97)*** (16.17)***

Spouse Working 0.73 0.53 0.65 -0.877

(if man) (12.74)*** (9.42)*** (12.22)*** (16.17)***

Observations 24,175 24,175 22,605 22,605 24,219 24,219 24,239 24,239

Absolute value of robust z statistics in brackets

* significant at 10%; ** significant at 5%; *** significant at 1%

Dependent variable is "Who does what"?

It takes values frrom 1 (always woman) to 5 (always man).

Controls: Respondent's gender (=1 if man), survey year, 

repondent's and spouse's work status,

iteration of man dummy with respondent's and spouse's work status, 

cohort and country dummies.

Allocation of Time To Houshold Production Activities

laundry cares_for_sick shop_for_groceries meal_preparation

39
B Gender Roles and a Woman’s Union Formation Probability

Table 7 presents three different specifications of the effect of gender roles on a woman’s

probability to enter a union. The second column in each specification takes into account a

woman’s attitudes toward gender roles on top of country average attitudes. Results are qualitatively the same as in table 5, with more egalitarian women having a lower probability of

entering a union, but women living in more egalitarian countries having a higher probability.

Specification 1 measures attitudes toward gender roles using the country and woman’s egalitarian index described in Section 4, but using a wider sample size (women between 20 and

65 years old). Specifications 2 and 3 use a different measure of attitudes toward gender roles.

Average country attitudes are measure by the percentage of respondents that answered either

totally agree or agree with the statement: "A job is alright, but what women really want is

home and children." A woman’s individual attitudes are measured by a dummy which takes

the value 1 if a woman responded totally agree or agree with the statement above. Another

difference between specification 2 and 3 is the sample size considered.

40
Table 7: Gender Roles and a Woman's Union Formation Probability

Woman's Probabiltiy of Entering a Union No individual With Individual No individual With Individual No individual With Individual 

(Maximum-likelihood Probit Estimation) Effects Effects Effects Effects Effects Effects

Completed years of schooling -0.01 -0.01 -0.01 -0.009 -0.015 -0.015

[-9.17]*** [-8.58]*** [-9.09]*** [-8.50]*** [-8.14]*** [-7.64]***

Age 0.01 0.01 0.01 0.01 0.025 0.025

[32.83]*** [30.07]*** [32.86]*** [32.64]*** [34.51]*** [34.54]***

Survey Year (=1 if 2002) -0.022 -0.02 -0.027 -0.028 -0.046 -0.049

[-1.83]* [-1.65]* [-2.11]** [-2.22]** [2.21]** [-2.33]**

Gender Roles by country and survey year 0.129 0.156

(higher values mean more "egalitarian") [1.80]* [2.12]**

Percentage of people by country and survey year -0.377 -0.405 -0.516 -0.562

that agree with the statement [-1.92]* [-2.07]** [1.60] [-1.74]*

What women really want is home and children

Woman's individual attitudes toward gender roles -0.019

(higher values mean more "egalitarian") [-6.24]***

Dummy=1 if woman agrees with the statement 0.022 0.035

"What women really want is home and children" [3.55]*** [3.36]***

Observations 15,165 13,325 15,165 15,165 8,914 8,914

(1) Robust z statistics in brackets

(2) * significant at 10%; ** significant at 5%; *** significant at 1%

(3) Coefficients report the change in the probability of entering a union for an infinitesimal change in each independent (continuous) variable 

 and, by default, the discrete change in the probability for dummy variables. 

Specification 1: The sample is women between 20 and 65 years old and gender roles are measured using the principal component index.

Specification 2: The sample is women between 20 and 65 years old and gender roles are measured using answers to the statement 

"A job is all right, but what women really want is a home and children".

Specification 3: The sample is women between 20 and 45 years old and gender roles are measured using answers to the statement

"A job is all right, but what women really want is a home and children".

Specification 1 Specification 2 Specification 3

41
C Evolution of Cross-Country Attitudes toward Marriage

Table 8: Cross-Country Comparison of Attitudes Toward Marriage*

Country/

Survey Year Agree N/A Disagree Obs

CA 1994 0.31 0.37 0.32 1,313

2002

SW 1994 0.17 0.38 0.44 1,177

2002 0.14 0.39 0.47 974

NO 1994 0.16 0.33 0.50 1,910

2002 0.15 0.38 0.47 1,368

GB 1994 0.23 0.39 0.38 867

2002 0.21 0.42 0.36 1,672

NI 1994 0.25 0.39 0.36 545

2002 0.30 0.24 0.45 845

US 1994 0.43 0.32 0.24 1,270

2002 0.39 0.33 0.28 1,063

NL 1994 0.11 0.25 0.64 1,743

2002 0.22 0.37 0.41 1,110

NZ 1994 0.22 0.20 0.39 932

2002 0.21 0.43 0.35 888

IR 1994 0.32 0.22 0.46 866

2002 0.31 0.26 0.43 1,094

SP 1994 0.31 0.19 0.50 2,259

2002 0.21 0.18 0.70 2,170

AUS 1994 0.41 0.35 0.23 1,560

2002 0.43 0.33 0.24 1,239

JA 1994 0.38 0.31 0.31 1,143

2002 0.31 0.36 0.32 948

GE 1994 0.38 0.27 0.34 2,129

2002 0.32 0.32 0.36 835

IT 1994 0.30 0.31 0.39 983

2002

OS 1994 0.44 0.27 0.29 852

2002 0.33 0.32 0.35 1,770

Source: ISSP 1994 and 2002. Author's calculations

* Do you agree or disagree: 

Married people are generally happier than unmarried people?

** Countries are ordered from more to less egalitarian 

according to the egalitarian index in 1994

42
D Evolution of Cross-Country Attitudes toward Cohabitation

Table 9: Cross-Country Comparison of Attitudes Toward Cohabitation

Country/

Survey Year Agree N/A Disagree Obs

CA 1994 0.69 0.13 0.18 1,313

2002

SW 1994 0.83 0.09 0.07 1,177

2002 0.89 0.05 0.06 974

NO 1994 0.76 0.08 0.15 1,910

2002 0.90 0.08 0.10 1,368

GB 1994 0.70 0.14 0.17 867

2002 0.75 0.13 0.12 1,672

NI 1994 0.50 0.18 0.31 545

2002 0.62 0.17 0.20 845

US 1994 0.46 0.18 0.31 1,270

2002 0.50 0.18 0.20 1,063

NL 1994 0.85 0.06 0.08 1,743

2002 0.91 0.50 0.03 1,110

NZ 1994 0.61 0.16 0.23 932

2002 0.67 0.15 0.18 888

IR 1994 0.54 0.09 0.36 866

2002 0.65 0.13 0.22 1,094

SP 1994 0.67 0.09 0.23 2,259

2002 0.80 0.07 0.11 2,170

AUS 1994 0.6 0.15 0.25 1,560

2002 0.67 0.16 0.17 1,239

JA 1994 0.38 0.22 0.40 1,143

2002 0.44 0.20 0.36 948

GE 1994 0.71 0.10 0.18 2,129

2002 0.81 0.08 0.11 835

IT 1994 0.60 0.12 0.27 983

2002

OS 1994 0.64 0.10 0.25 852

2002 0.80 0.10 0.09 1,770

Source: ISSP 1994 and 2002. Author's calculations

* Do you agree or disagree: "It is all right for a couple 

to live together without intending to get married"

** Countries are ordered from more to less egalitarian 

according to the egalitarian index in 1994

43