Cohabitation in Great Britain: Not for Long, But Here to Stay

Cohabitation in Great Britain: Not for Long, But Here to StayAuthor(s): John Ermisch and Marco FrancesconiSource: Journal of the Royal Statistical Society. Series A (Statistics in Society), Vol. 163, No. 2(2000), pp. 153-171Published by: Wiley for the Royal Statistical SocietyStable URL: http://www.jstor.org/stable/2680495 .

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J. R. Statist. Soc. A (2000) 163, Part 2, pp. 153-171

Cohabitation in Great Britain: not for long, but here to stay

John Ermisch and Marco Francesconi

University of Essex, Colchester, UK

[Received May 1998. Final revision June 1999]

Summary. This paper uses a new source of data to study the dramatic increase in cohabiting unions in Great Britain. It analyses, in turn, entry into first partnership, the stability of cohabiting unions and repartnering after dissolution of cohabitation. In excess of 70% of first partnerships are now cohabitations, and these last a relatively short time before being either turned into marriage or dissolved. The shift to cohabitation as the dominant mode of first partnership plays an important role in the delay of first marriage and motherhood. The paper also investigates the factors that are associated with the outcome of cohabitations.

Keywords: Cohabitation; Delayed marriage; First partnership; Union stability

1. Introduction

During the last two decades, there have been dramatic changes in patterns of live-in partnership formation and dissolution in the countries of northern and western Europe. A very important element in these changes has been the rise in cohabitation without legal marriage. For example, between the mid-1970s and the late 1980s, the percentage of women aged 20-24 years in live-in partnerships who were cohabiting rose from 11% to 49 % in France, from 57% to 78% in Sweden and from 48% to 75% in Denmark (Kiernan, 1996). In Great Britain, this percentage increased from 11% to 55% between 1980 and 1995 (Office for National Statistics (1997), Table 12.3). This paper analyses the dynamics of cohabiting unions in Britain and relates them to the delay in first marriage and motherhood among more recent cohorts. It also considers in more detail which cohabiting unions are converted into marriage and which dissolve.

In Britain, it has been difficult to study the dynamics of cohabiting unions and their relationship to other key demographic variables, including ages at first partnership, marriage and motherhood, from the registration and annual household survey data. The main source of British data on cohabitation has been the General Household Survey (GHS). The information in the GHS is, however, limited to whether respondents who were not living in a married partnership at the time of the survey were cohabiting, whether respondents who were currently or previously married had cohabited with their spouse before their marriage and durations of cohabitation for unions in progress at the time of the survey (for further details, see Haskey and Kiernan (1989) and Haskey (1992, 1995)). The National Child Development Study (NCDS) has full partnership histories, but they only pertain to the 1958 birth cohort.

Address for correspondence: John Ermisch, Institute for Social and Economic Research, University of Essex, Wivenhoe Park, Colchester, C04 3SQ, UK. E-mail: [email protected]

? 2000 Royal Statistical Society 0964-1998/00/163153



154 J. Ermisch and M. Francesconi

The analyses of partnership dynamics among the 1958 birth cohort by Berrington and Diamond (2000) are unique in incorporating cohabitation.

The analysis in this paper exploits a new source of data on British live-in partnerships. The second wave of the British Household Panel Study (BHPS) collected, during the last quarter of 1992, complete histories of all spells of marriage and cohabitation from a representative sample of 9459 adults aged 16 years and over throughout Great Britain. These included both cohabitations which preceded legal marriage and those which were not associated with any marriage. Information on cohabitation is elicited by the question

'As you know some couples live together without actually getting married. Have you ever lived with someone as a couple for three months or more?'

If the answer is yes, questions then ask how many such partnerships he or she had and the months and years at which they started and stopped living together.

As we show later, where there is overlap, results based on these data accord well with other sources, including the NCDS, GHS and registration data. We focus on people born since 1930 to reduce problems of inference created by differential mortality or recall error. In addition to these life-history data, the first five (annual) waves of the panel (1991-1995) are used to analyse the dynamics of cohabiting unions (see Appendix A for more information about the BHPS).

Fig. 1 describes the life-history data which we analyse for women. For instance, it shows that of the 3984 women in the sample 82% (3273) reported having at least one partnership, and of these 29% (964) started their first partnership as a cohabitation. Just over half of these cohabitations (504) turn into marriage whereas 29% (280) dissolve. A quarter (133) of these marriages are observed to dissolve, and so on. Although this tree diagram provides a concise description of the data, it does not control for each person's amount of exposure to risk of different transitions; for instance, many of the 711 women who are still single are very young. Nor does it differentiate between the behaviour of different birth cohorts.

The next section of the paper applies life-table methods to these data distinguished by broad cohorts and sex, and analyses, in turn, entry into first partnership, the stability of cohabiting unions and repartnering after dissolution of cohabitation. The third section examines the role played by cohabitation in the changes in the patterns of first marriage and motherhood across cohorts. In the light of the importance of dissolution of cohabitation in the creation of one-parent families headed by a never-married mother, as well as in delaying marriage and motherhood, the fourth section analyses which cohabitations are converted into marriage and which dissolve. Section 5 briefly examines childbearing within cohabiting unions, and Section 6 presents our main conclusions.

2. The formation and dissolution of cohabiting unions

2.1. First partnership: cohabitation or marriage? In this analysis, after reaching the age of 16 years, each person faces three options in each month. He or she can marry, cohabit outside marriage or remain single (i.e. without a partner). To examine change over time, the analysis is carried out for three sets of birth cohorts, all of whom were making partnership decisions in the post-war period. Table 1 shows estimates of entry into first partnership by the type of partnership, age, sex and birth cohort. These are based on maximum likelihood estimates of transition rates to marriage and cohabitation, which are assumed to be piecewise constant by single year of age (for example



Cohabitation in Great Britain 155

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Table 1. Proportions of men and women who ever had a partnership by age, birth cohort and partnership typet

Partnership type Proportions (per 1000) for the following ages (years): and birth cohort

16 20 24 28 32 36

Men Marriagel

1930-1949 0 47 473 721 809 840 1950-1962 0 62 327 476 523 536 1963-1976? 0 18 110 All 0 43 336 536 624 656

Cohabitation?? 1930-1949 0 8 29 49 67 79 1950-1962 0 42 163 278 342 371 1963-1976? 1 107 387 All 0 45 144 209 245 257

Any union* 1930-1949 0 55 502 770 876 919 1950-1962 0 104 490 754 865 907 1963-1976?? 1 125 497 All 0 88 480 745 869 913

Women Marriagel

1930-1949 1 231 700 837 875 887 1950-1962 2 228 544 625 642 649 1963-1976? 1 66 210 All 1 179 518 641 697 709

Cohabitation?? 1930-1949 0 8 33 44 53 58 1950-1962 1 87 207 270 294 307 1963-1976? 3 204 448 All 1 88 183 213 234 240

Any union* 1930-1949 1 239 733 881 928 945 1950-1962 3 315 751 895 936 956 1963-1976? 4 270 658 All 2 267 701 854 931 949

tObtained from maximum likelihood estimates of transition rates to marriage and cohabitation by single year of age. Partnership type refers to the way in which the partnership started. The standard errors of the survival curve at age 24 years range between 8 and 21 for men and between 6 and 22 for women. Excludes marrige after cohabitation.

?Estimates at age 24 years are based on people born during 1963-1968, and estimates at age 20 years are based on people born during 1963-1972. ??Whether married first partner or not. *Includes 'marriage' and 'cohabitation'.

see Lancaster (1990), pages 176-180). The survivor function implied by these estimates is the Kaplan-Meier, or product limit, estimator.

It is clear that cohabitation has become a much more important route into first partnership. By their 24th birthday, more than two-fifths of the women in the most recent cohort (1963-1976) had entered cohabitation, compared with a fifth of the previous cohort. There was a corresponding fall in the proportion of women who went directly into marriage, from 54% to 21%. By comparison, 35% of women and 40% of men in the 1958 birth cohort, surveyed in the NCDS, had cohabited in their first partnership by the time of their 33rd birthday (Berrington and Diamond, 1995), compared with 30% of women and 35% of men




for the 1950-1962 BHPS cohort. Among the 1950-1962 cohort, a third of women and two- fifths of men who ever had a partner (by age 36 years) cohabited in their first partnership. Estimates for the truncated 1963-1976 cohort, based on behaviour up to age 24 years, suggest that the proportion cohabiting in their first partnership has risen to 68% for women and 78% for men. Thus, the relative proportions marrying directly and cohabiting are reversed between the two cohorts, and cohabitation has become the dominant mode of first partnership.

Partnerships are also being postponed. The proportion of women who have ever had a partnership has declined between the latest two cohorts: 75% of women in the 1950-1962 cohort had entered a first partnership by the age of 24 years, but this had declined to 66% for the 1963-1976 cohort. This translates into a rise in the women's median age at first partnership of 1 year between the two cohorts, from 21 years and 3 months to 22 years and 3 months. Postponement among men is not evident up to the age of 24 years, which is approximately the median age in both the 1950-1962 and the 1963-1976 cohorts, but men are also postponing their first partnerships: 60% of men in the 1950-1962 cohort had partnered by age 26 years, and this has fallen to 52% for the 1963-1976 cohort.

The panel data can also be used to estimate first partnering patterns during the 1990s. The first five waves of BHPS data (1991-1995) are used in conjunction with the life-history data to estimate first transitions from the never-married, unpartnered state into a partnership. For people under the age of 35 years entering first partnerships, these data indicate that 79% of men and 71 % of women cohabit in their first partnership. These may indeed be under- estimates of the importance of cohabitation, because the panel data would miss cohabitations which started and ended between two annual waves of the panel. Although these estimates may seem high, they are not inconsistent with the seven out of 10 first marriages in the early 1990s which were preceded by the spouses' cohabitation (Haskey, 1995), nor with the truncated estimates for the 1963-1976 cohort in Table 1.

2.2. Duration of cohabiting unions How long do people live together in cohabiting unions before either marrying their partner or dissolving their union? Using the life-history data, Table 2 indicates that cohabitations are usually very short. Less than a fifth survive 5 years or more, and less than a tenth survive 10 years or more. The median duration is just under 2 years. About three-fifths of first cohabitations turn into marriage and about 30% dissolve within 10 years. We have not been able to compute exact counterparts for the 1958 birth cohort, but Berrington and Diamond (1995) found that, by age 33 years, 28% of first cohabiting unions had dissolved, 64% were converted into marriage and 8% were intact.

These same issues can be addressed with the panel data by looking at two pieces of information in the BHPS: the reported de facto marital status, which includes 'living as a couple' (not married) as one of the states, and the personal identity number of the partner.

Table 2. Stability of cohabiting unions and repartnering: survival in first cohabiting union, ment

Proportion s5Urviving by years since beginnling of union N I year 2 years 3 years 4 years 5 years 10 years

Men 68 42 27 21 17 8 897 Women 71 47 32 23 17 6 964

fThe maximum standard error in each row is 1.7.




The latter is particularly important for cohabiting unions, because a person could be living as a couple in two consecutive waves, but he or she could have changed partners between waves. Thus, people in cohabiting unions are defined as those not married but living as a couple, and their union is defined to end either if they are not living as a couple unmarried in the next wave or they have changed partners. 29% of cohabiting unions involving never-married women end each year. If the estimated transition rates were constant over time, they suggest a median duration of 2 years and less than 4% of unions lasting 10 years or more. Again, about three-fifths of the unions turn into marriages.

The small increase in median duration compared with the estimates from the retrospective data suggests longer durations of cohabitations in the 1990s than earlier. This is consistent with data on the median duration of incomplete cohabitations. Haskey and Kiernan (1989) and Kiernan and Estaugh (1993), Table 2.5, estimated median elapsed durations of cohabitation of 20 and 21 months respectively from samples from the 'stock' of never-married cohabiting women in 1986-1987 and 1989 (obtained from the GHS). Haskey (1995) estimated that this median elapsed duration increased to 29 months in 1990-1993. It should, however, be noted that the distribution of elapsed durations usually differs from the distribution of completed durations among entrants (for example see Lancaster (1990), pages 91-97), and it is the latter which is estimated in the current analysis.

2.3. Repartnering As Fig. 1 shows, if a woman's first partnership was a cohabitation which dissolved, almost all of those who repartnered cohabited in their second partnership. The estimates in Table 3 indicate that, after a cohabiting first partnership has dissolved, the median duration to the next partnership is 5 years, which is somewhat higher than we expected. But estimates from the panel data indicate similar rates of repartnering after the dissolution of cohabitation. About a third repartner within 3 years, according to both sources of data.

Cohabitation is the dominant way of repartnering after the dissolution of marriage. In our life-history data, 70% of second partnerships following dissolution of a first marriage started as cohabiting unions, and during the panel study (1991-1995) two-thirds of divorced and separated people who repartnered cohabited. Our ability to identify trends is limited by small sample sizes, but it is feasible to divide the group who dissolved first marriages into those doing so before and after 1981 (producing roughly equal sample sizes). There was an increase in the proportion of cohabitations among second partnerships between these two marital dissolution cohorts (e.g. an increase from 63% to 75% for women repartnering). This trend is broadly consistent with evidence from the GHS, which shows an increase in the percentage of women cohabiting before their second marriage from around 30% in the late 1960s to about 90% for second marriages in the early 1990s (Haskey, 1995).

Table 3. Stability of cohabiting unions and repartnering: interval between first and second cohabiting unions, ment

Proportion repartnered by years since end of first union N

I year 2 years 3 years 4 years 5 years 10 years

Men 16 27 36 44 51 65 252 Women 14 28 35 42 50 62 280

fThe maximum standard error in each row is 1.4.




800

700

600-

600

400

300

200

17 18 19 20 21 22 23 24 25 26 27 28 29 30

Age

Fig. 2. Proportions (per thousand) of women who had ever married: , 1961; --- 1966; ------ 1967; - 1968; , 1969; , 1970; 1971

3. Cohabitation and the delay in marriage and motherhood

Nine out of 10 women born in 1935 had married by their 30th birthday; this proportion is two-thirds for those born 30 years later. Fig. 2 shows (from marriage registration data) that, at each age, the proportion of women who have ever married is lower for each succeeding cohort from the 1961 birth cohort onwards (and the same is true for men). For those born in 1966, the median age at marriage is 26 years for women and more than 29 years for men. Section 2.2 has shown that, although more recent cohabitations appear to be lasting longer before dissolving or being turned into marriage, only a small percentage (about 5%) survive 10 years or more.

Thus, there appears to be a puzzle: long-term cohabitations are rare, but recent generations of young people are not marrying. The answer lies in the combination of four factors documented above: the large proportion of people who cohabit before any marriage, the time spent cohabiting, the relatively high risk that cohabitations dissolve and the time it takes to cohabit again. All these contribute to a longer time before any marriage takes place and increase the chances that a person never marries. This section argues that it is the shift to cohabitation as the dominant mode of first partnership which is the main engine for the trends in marriage patterns illustrated in Fig. 2. We focus on women, but similar conclusions are obtained for men.

Fig. 3 illustrates the sequence of partnership formation and dissolution decisions that are currently involved in producing the proportion who ever marry (we ignore mortality). The numbers in the boxes show the results of a simulation of partnership formation and dissolution behaviour up to a woman's 45th birthday based on recent partnership formation and dissolution rates. These are the rates which lie behind the partnership life-tables that were discussed in the previous section, with first-partnership rates at ages under 28 years being based on women born during 1963-1976. The assumptions for the simulations are summarized




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Fig. 3. Longer road to marriage

Table 4. Simulation assumptions

(a) First-partnership assumnptions 1. For those born during 1950-1962, actual age-specific rates are used (estimated

from BHPS life-history data) 2. For those born after 1962, actual age-specific rates are used for ages 16-27 years;

rates for the 1950-1962 cohorts are used for ages 28-44 yearst 3. For the simulation with the 1950-1962 cohort cohabitation and marriage rates

reversed, only rates for ages 16-27 years are reversedt

(b) Cohabitationi outcome and repartnering assulmptions 1. Constant (with union duration) marrige and dissolution rates for women in

cohabiting unions from the 1990s (BHPS panel data), which are very similar to those for the cohabiting unions starting in 1981 or later in the life-history data

2. Constant (with time since the first union dissolved) repartnering rate from life- history data for never-married women whose first cohabiting union dissolved

tResults are not very sensitive to different plausible assumptions for ages 28-44 years (e.g. age-specific rates 50% higher or lower). tCohabitation rates exceed marriage rates at ages 28-44 years for the 1950-1962 cohort, whereas the opposite is the case for younger ages.

in Table 4. As the broken curve in Fig. 4 illustrates in an alternative way, the simulation suggests that nearly 90% of women born in the mid-1960s may eventually marry. It is just taking a much longer time because of intervening cohabitations. Note that the median age at marriage (26 years) and the proportion ever married at age 30 years (65%) are close to those observed from registration statistics for the 1965 birth cohort.




1000 -

900

800

700

600

500

400

300-

200-

100 S,

17 18 19 20 21 22 23 24 25 26 27 28 29 30 31 32 33 34 35 36 37 38 39 40 41 42 43 44 45

Age Fig. 4. Proportion (per thousand) of women who had ever married: estimated rates for women from two groups of cohorts ( , 1950-1962 cohorts; ---, 1963 and later cohorts)

The full curve in Fig. 4 shows the results of a simulation of the series of partnership formation and dissolution decisions using the same rates for the outcomes of cohabitations and for repartnering as those used to generate the broken curve, but the rates for the first partnership are based on the experience of women born during 1950-1962, rather than those born during 1963-1976. Note that the median age at marriage (22 years) and the proportion ever married at age 30 years (83%) are close to those observed from registration statistics for the 1956 birth cohort. The difference between the two curves indicates the postponement of marriage which arises because of changes in the patterns of first partnership, both its delay and the substitution of cohabitation for direct marriage. The proportion ever marrying (by age 45 years) is about 6 percentage points lower for the later generation, as a consequence of the change in first-partnership rates at ages under 28 years.

As discussed in Section 2.1, the major change in first-partnership patterns is the approx- imate reversal of the proportions cohabiting and marrying directly in their first partnership between the two groups of birth cohorts. There was, however, also a tendency for first partnerships to occur later. To remove this latter aspect of change, a simulation which uses the first-partnership rates for the 1950-1962 cohorts, but reverses the cohabitation and marriage components of these rates (at ages below 28 years), is shown as the full curve in Fig. 5, and the broken curve again repeats the broken curve in Fig. 4. A comparison of these two curves suggests that the large increase in the proportion of women who cohabit in their first partnership may explain most of the delay in marriage. As indicated by the difference between these two curves, a later age at first partnership only accounts for a small proportion of the delay in marriage.

But a later age at first partnership appears to have played a larger role in the trend towards later marriage among cohorts born since the mid-1960s. If we take estimates of partnership rates for never-married people from the 1991-1996 BHPS as being the relevantfirst-partnership rates for these more recent cohorts, these suggest an increase in the median age at first




1000

900

800

700

600

500

400

300

200

100-

17 18 19 20 21 22 23 24 25 26 27 28 29 30 31 32 33 34 35 36 37 38 39 40 41 42 43 44 45

Age

Fig. 5. Proportion (per thousand) of women who had ever married: estimated rates for women from two groups of cohorts ( , 1950-1962 cohorts, cohabitation and marriage rates reversed; ---, 1963 and later cohorts)

partnership to 23 years for women (and 25 years for men). With the same rates for the outcomes of cohabitation and repartnering as before, but these first-partnership rates, the median age at marriage would rise to 28 years for women, only 57% would be married by age 30 years and only 82% would marry by age 45 years.

An important reason why we are interested in the delay in marriage is its implications for childbearing patterns. Registration statistics indicate that half of women born in 1964 had become mothers by their 27th birthday, which is 2 years later than the median age at first birth for the 1956 birth cohort. The fact that at least three-fifths of first births are born within marriage immediately suggests that the shift to cohabitation as the dominant form of first partnership was also strongly associated with the postponement of motherhood.

Using similar simulation methods as above, but now incorporating first-birth rates in cohabiting unions, in marriage and outside a partnership, we find that, if the rates of first entry into a cohabiting union and marriage had remained the same as for the 1950-1962 cohort, the median age at motherhood for the 1963-1976 cohort would have decreased to 25.2 years rather than increasing to 27.5 years (from 26.4 years for the 1950-1962 cohort). In other words, changes in childbearing behaviour and partnership formation and dissolution rates other than first-partnership entry rates would have reduced the median age at first birth, whereas it actually increased. Thus the delay in first partnerships and the substitution of cohabitation for direct marriage are jointly responsible for the delay in motherhood between the two synthetic cohorts. The further delay in first partnerships for cohorts born since the mid- 1960s acted to continue the increase in the median age at motherhood.

4. Outcomes of cohabiting unions: marriage or dissolution? In addition to the type of first partnership, an important determinant of the timing of first marriage and motherhood is whether and when young people convert their cohabitation into




marriage. Another reason for our interest in the outcome of cohabiting unions is that a substantial minority of them produce children. Life-table estimates from the BHPS life- history data indicate that the percentage of women starting cohabiting unions who become mothers while cohabiting was 18% for those born during 1963-1976, an increase from 9% for the 1950-1962 cohort, and about two-fifths of one-parent families headed by never- married mothers are created through the dissolution of cohabiting unions (Ermisch, 1997). We utilize the 1991-1995 BHPS panel data for original sample members to analyse whose cohabitation dissolves and whose is converted into marriage.

In addition to having a rich set of observable explanatory factors, which can vary over time in the union, a distinct advantage of the BHPS is that we can use the panel data in combination with the life-history data. The latter are used to calculate the person's age at the start of the cohabiting union and the duration of unions for unions in progress at the start of the panel study in 1991. These two demographic characteristics have often proved to be valu- able in accounting for variation between people in partnership dissolution patterns (see, for example, for some recent studies, Berrington and Diamond (1995, 2000), Lillard et al. (1995), Hoffiman and Duncan (1995) and Weiss and Willis (1997)), and it is important to know the duration of unions in the analysis which follows. We shall focus on cohabiting unions con- taining never-married women aged 50 years and under. Although we focus on women, cohabiting men in the panel study would be in unions with the women in our sample, and we include some characteristics of the male partners as explanatory variables in the analysis along with attributes of the couple. Thus, it is close to an analysis of unions.

4. 1. Bivariate associations In our sample of 694 woman-year observations, 29% end their cohabitation in each year, 18% converting it into marriage and 110% dissolving their union. Table 5 shows some bivariate associations between characteristics of the woman, her partner or the couple and what happens to the union during the following year. All characteristics are measured in the previous year, t - 1, or during the previous two years, from t - 2 to t- 1. The sample sizes vary across parts of Table 5 because of missing observations on particular variables. In addition to reporting the Pearson test for independence between two variables, we also report the Kruskal-Wallis test for independence in the parts of Table 5 in which the explanatory variable is ordered.

Women in employment are less likely to dissolve their unions and more likely to marry than those who are not in a job, but the small minority of those who are full-time students are most likely to dissolve and least likely to marry in the following year (part (a) of Table 5). Having a partner in a job also increases the chances of marriage and reduces the risk of dissolution of the union (part (b)). Thus, it is not surprising that if the couple received income support benefits, the main means-tested welfare benefit, they have a higher rate of dissolution and a lower marriage rate (part (c)).

Mothers are much less likely to marry than childless women and are slightly more likely to have their union dissolve (part (d)). Among mothers, the small minority having a child aged 5 years or more are more likely to marry but face a similar dissolution risk to those with a preschool child (part (e)). There is some variation in marriage and dissolution rates with family size among mothers: couples with more children are more likely to marry and less likely to dissolve their unions (part (f)).

The majority of never-married women in cohabiting couples are owner-occupiers. Tenants are less likely to marry than owner-occupiers and are somewhat more likely to dissolve their




Table 5. Percentages of never-married women in cohabiting unions marrying and dissolving their unions

% marrying % dissolv,ing N

(a) By woman's employment status In paid employment

employed 19.8 8.7 531 not employed 11.6 15.7 147 student 6.3 43.8 16

Pearson x2 27.88 (4 degrees of freedom)t

(b) By whether partner was in employmnent In paid employment

yes 19.1 9.7 559 no 11.9 16.3 135

Pearson x2 7.64 (2 degrees of freedom):

(c) By r-eceipt of income suppor t Received income support

yes 8.8 18.8 80 no 18.9 9.9 614

Pearson x2 9.15 (2 degrees of freedom):

(d) By whether a mother Mother

yes 10.0 11.9 210 no 21.1 10.5 484

Pearson x) 12.32 (2 degrees of freedom)?

(e) By age of youngest child (niothe7s only) Aged

less than 5 years 8.8 11.5 148 5 years or more 25.6 10.3 39

Pearson x2 8.16 (2 degrees of freedom)t; Kruskal-Wallis x2 4.02 (2 degrees of freedom)

(f) By number of dependent children Number of children

0 21.1 10.5 484 1 8.0 12.4 113 2 12.0 13.3 75 3 or more 13.6 4.6 22

Pearson x2 14.37 (6 degrees of freedom)t; Kruskal-Wallis x2 7.11 (2 degrees of freedom): (continled)

union (part (g)), which may reflect a tendency for house purchase to be a signal that a longer- term commitment has been made. Overall these associations are not, however, statistically significant at the 0.05 level. Couples who thought that their financial situation improved in the previous year (i.e. between t - 2 and t - 1) are more likely to marry and less likely to dissolve their unions relative to others (part (h)), but the patterns of marriage and dissolution did not vary significantly with perceived changes in financial situation. No statistically significant patterns by women's education level nor father's occupation group were evident (parts (i) and (j)).

Economic theory predicts that 'surprises' may have particularly large effects on the dissolution of unions (Becker et al., 1977; Weiss and Willis, 1997). It argues that traits which




Table 5. (continued)

% marrying % dissolving N

(g) By housing tenure Housing tenure

owner-occupier 20.1 9.6 438 local authority or housing association 15.0 10.2 127

tenant other tenant 12.7 16.7 126

Pearson x2 8.53 (4 degrees of freedom)

(h) By change in financial situation Financial situation

better off 20.9 9.1 263 about the same 16.4 12.8 219 worse off 15.5 11.6 207

Pearson xX 3.98 (4 degrees of freedom); Kruskal-Wallis x2 2.67 (2 degrees of freedom)

(i) By highest educational qualification Qualification

degree 15.5 10.3 97 other higher educational qualification 21.2 8.6 151 A-level 18.8 13.5 96 0-level 20.3 13.2 197 other 14.4 11.1 90 none 9.1 7.3 55

Pearson x2 10.07 (10 degrees of freedom); Kruskal-Wallis x2 1.03 (2 degrees of freedom)

(j) By father's occtupation Qualification

professional or manager 18.7 8.6 209 skilled non-manual 15.6 13.3 45 skilled manual 22.5 9.5 222 semiskilled or unskilled 13.5 11.7 111


(k) By 'SUr pr-ise' in financial situation Financial situation relative to expectation

worse 14.0 15.3 157 as expected 21.1 11.7 213 better 18.2 4.6 88

Pearson x) 9.07 (4 degrees of freedom); Kruskal-Wallis x2 5.46 (2 degrees of freedom)

tp < 0.001. lp < 0.05. ?P < 0.01.

influence the benefits of a particular union can change over time in an unpredictable manner, and such surprises can cause either partner to reconsider their original decision, which suggests that an unexpectedly good or poor financial situation may increase the dissolution of unions. This idea is implemented here by going back 2 years (to t - 2) and utilizing the question which asks whether a person expects to be better or worse off in the next year (t - 1). Their assessment of the actual change in financial situation between years t - 2 and t - 1 (which was tabulated in part (h)) is then compared with their expectations at year t - 2. For example, if the woman says that she expected to be better off at year t - 1 and ends up feeling worse off or that her financial situation has not changed, we say that she is 'worse off' relative to her expectations a 'negative surprise'. Similarly, if she expects her financial situation to




Table 6. Percentages of never-married women in cohabiting unions marrying and dissolving their unions: means, continuous variables by union outcomet

Variable % intact % married % dissolved Test for equality of means

Age at start of union 23.2 (4.8), 23.1 (4.3), 21.6 (3.9), F2,698 = 4.091 N=495 N= 123 N=76

Duration of union in years at t - 1 4.1 (3.7), 3.3 (3.5), 2.8 (2.7), F9 690 = 5.48? N=494 N=123 N=76

Household income at t - 1 1959 (1383), 2077 (1192), 1551 (870), F9689 = 4.121 N = 493 N= 123 N= 76

Partner's earnings at t - 1 1192 (841), 1270 (743), 1059 (311), F2,492 = 1.19 (employed only) N= 347 N= 99 N= 49

Own earnings at t - 1 955 (579), 964 (536), 833 (461), F2,506 = 1.01 (employed only) N= 362 N= 102 N = 45

tStandard deviations are given in parentheses. lp < 0.05. ?P < 0.01.

remain the same and ends up feeling worse off, this is also a negative surprise. Positive surprises ('better off' relative to expectations) are defined analogously. Part (k) of Table 5 indicates that negative surprises decrease the chances of marriage, whereas positive or negative surprises reduce or increase respectively the risk of dissolution of unions. Overall, the patterns are not fully consistent with the economic argument; nor are they statistically significant at the 0.05 level. Note that the sample size is substantially lower than in the other parts of Table 5, because we must have an additional year of panel data on a woman to define the surprise variable.

Table 6 shows differences in means across the three groups. Women who dissolved their union had a lower mean age at the start of the union relative to those who either married or continued to cohabit. Couples who continued to cohabit had a higher mean duration of union (as of the previous year) than those who married, and those who dissolved their union had the lowest mean duration. Women who dissolved their union had the lowest household income, whereas women who married had the highest. Women's own earnings and their partner's earnings were lowest for those dissolving their union, and these were highest for those who married, but these differences are not statistically significant at the 0.05 level.

4.2. Competing risks model All these variables were considered in an analysis of the outcome of cohabiting unions. The model that we estimate is a discrete time competing risks transition rate model, the two risks being marriage and dissolution of the union. Under the conventional independence assumptions, the log-likelihood function factors into a sum of terms, each of which is a function of the parameters of a single transition rate only. As a consequence, we can estimate each of the competing risk transition rates by treating the alternative outcome as censoring at that point (see Narendranathan and Stewart (1993), page 68). Furthermore, the resulting log-likelihood function is identical with that of a binary logit model for each type of transition (see Allison (1982), pages 74-75).

The outcome-specific transition rate for ending the cohabiting union in year t for the jth couple, pijt, is assumed to take the form

ln {pj1/(I -pijt)} = ai ln(durjt-,) + /iXjt-l, i = 1, 2 (1)




Table 7. Discrete time competing risk transition rate model for the outcome of cohabiting unionst

Variable Odds of union Odlds of dissolution macrriage

log(duration of union) at t- 1 -0.160 (1.44) -0.131 (1.45) Age at start of union -0.092 (2.28) -0.013 (0.57) (Partner in job x earnings at t - 1)/100 -0.0323 (1.78) 0.0199 (1.74) Student at t - 1 1.386 (2.82) - 1.455 (1.34) Mother at t - 1 -0.201 (0.64) -0.921 (3.36) Constant 0.382 (0.42) -1.054 (1.93)

tRatios of the coefficient to the Huber-White-sandwich robust standard errors are given in parentheses; N = 629; x2 41.70 (10 degrees of freedom).

where the couple may dissolve their union (i = 1) or may marry (i = 2) in year t, con- ditionally on survival in the union until t - 1, durj,_1 is the duration of the union at year t - 1, X1,_l is a vector of explanatory variables measured at year t - 1, including a constant term, /3 is a vector of parameters to be estimated and ai is also a parameter to be estimated. This formulation assures that the probability that the couple remain together in year t is given by (1 - pljt)(I - P2jt), as is appropriate in a discrete time competing risk model (see, for example, Narendranathan and Stewart (1993)).

The data used to estimate the model are annual observations on cohabiting couples at risk either to dissolve the union or to convert it into marriage in the following year (i.e. cohabiting at year t - 1). Some of these observations come from unions in progress at the beginning of the panel study (1991). The contribution to the likelihood function of such unions must, therefore, condition on surviving in the union up to the time of the start of the panel. Jenkins (1995) showed that, owing to 'cancelling of terms' in the conditional survivor probability, their likelihood contribution depends only on the transition rates and data for years since the beginning of the panel study, provided that the total elapsed duration of union (dul>i_-) is used for the transition rates in model (1). This convenient cancelling result does not carry over to models in which there is unmeasured couple-specific heterogeneity, nor to analogous continuous time transition models (see Lancaster (1990), chapter 8). We obtain this duration by combining the life-history data with the panel data.

The parameter estimates in Table 7 are for a relatively parsimonious model, which has excluded the other potential explanatory variables in Table 6 on the basis of likelihood ratio tests. Each partner's earnings initially enter the model in the following interactive way:

Oil empjt ? + /i2 empjt_1 * earningsjty,,

where 'emp' denotes whether or not the partner is employed and 'earnings' is his or her real monthly earnings if employed. These two variables appear together to provide more flex- ibility in the effects of being in a job and of earnings on that job. A woman's own earnings and employment never approached statistical significance, but her partner's earnings are important for the outcome of cohabiting unions. We can also accept the hypothesis that the Oil-coefficients are 0 (X2 = 0.78, 2 degrees of freedom); thus the effect of an additional pound of his earnings on the outcome is /i2 empj,-1, and the effect of being in employment is /iA2 earningsj.,-.

There are several noteworthy features of the estimates in Table 7. First, there are few significant predictors of the outcomes of cohabiting unions. Second, most of these are not known for certain at the start of the union but rather evolve during the union. Third, the




annual rates of dissolution of a union and marriage fall with the duration of the union, although not significantly. Fourth, higher earnings for a woman's partner decrease the chances that the union is dissolved and increase the chances that it is converted into marriage. Information about the partner's earnings prospects may be very limited at the time that the union was formed, and so the experience of high or low partner's earnings during the union tends to strengthen or weaken it respectively. This is consistent with the idea that new information is important for the outcome of cohabiting unions. Unions in which the woman is a full- time student are four times more likely to dissolve, but these are very rare (2.3% of person- year observations).

People who form a cohabiting union earlier in their life are likely to have searched for a shorter time and therefore are more likely to make a poor match. They should, therefore, be more likely to dissolve the union in the future. For example, the negative relationship between age at first marriage and the probability of marital dissolution found in most studies is consistent with this argument (see Becker et al. (1977) and Weiss and Willis (1997)). Age at the start of the union is indeed found to be negatively associated with dissolution of the union.

Finally, being a mother has little effect on the union dissolution rate, and it actually reduces the odds of converting the union into marriage (relative to not marrying) by 60%. This finding is consistent with Lelievre's (1993) results (in her Fig. 7.7 on page 120), using data from the 1989 GHS. When we take account of the effects of motherhood on the two competing risks together, cohabitations with children are more likely to dissolve eventually, because they are subject to the high dissolution risk of cohabitations for longer. However, mothers also tend to stay in cohabiting unions longer than non-mothers.

Of course, these women may not have become mothers within the current union. The panel data do not support exact timings of births relative to the formations of unions, but they suggest that about a quarter of mothers in cohabiting unions had their youngest child before the start of the union. If these mothers are distinguished in the multivariate analysis, they are as likely to marry (or to dissolve) as are childless women. The odds of marriage (relative to not marrying) for women who have their youngest child within the union are 67% lower than for childless women, and the union dissolution rate is also lower, but not significantly so (the coefficient for a birth within a union in the marriage odds equation is -1.128, with a standard error of 0.344, whereas the corresponding coefficient in the dissolution odds equations is -0.502, with a standard error of 0.409).

Thus, it appears that births within cohabiting unions substantially reduce the marriage rate, which ultimately leads to more of these unions dissolving than for childless women. As measured by conversion of the union into marriage, childbearing within cohabiting unions does not signal longer-term commitments, but it does signal longer cohabitations. We may be observing a selection mechanism in which couples who are favourably inclined towards marriage marry before the child arrives, and those who are not favourable towards marriage but anticipate a longer-term commitment are more likely to have a child. They are, however, still subject to a high risk of dissolution of the union, and so many of these fertile unions end up dissolving.

5. Births within cohabiting unions

In the light of this last finding, women who give birth in cohabiting unions face a relatively high risk of becoming lone mothers because the high risk that cohabitations dissolve is not mitigated for them. It is, therefore, interesting to examine who becomes a mother in cohabiting unions.




Table 8. Percentage becoming mothers in a cohabiting union

% with % unioni dissolved or N birth marriage

(a) By whether partner iwas in job In paid employment at t- 1

yes 2.8 29.6 429 no 9.1 30.9 55


(b) By sur-prise in financial situiation Change in financial situation relative to expectation

worse 7.1 29.3 99 as expected 0.7 34.9 152 better 5.3 22.8 57

Pearson x2 9.97 (4 degrees of freedom)t; Kruskal-Wallis x2 1.63 (2 degrees of freedom)

(c) By father s occupational group Father's occupation

professional or manager 1.8 26.8 164 skilled non-manual 3.1 28.1 32 skilled manual 1.9 34.6 162 semiskilled or unskilled 9.2 24.6 65

Pearson x2 12.73 (6 degrees of freedom)t

tp < 0.05.

In our sample of 484 childless woman-year observations (see Table 5, part (d)), 3.5% of never-married childless women who start the year in a cohabiting union give birth in the union during the year (and are cohabiting at the end of the year). This represents a lower bound on the birth-rate, because some of the cohabitations which were dissolved or converted into marriage during the year may have produced a child before the dissolution or conversion. Table 8 shows how this conservative estimate of the annual first-birth rate in cohabiting unions varies with some characteristics of never-married childless women and their partners. It also shows how the competing risk that the cohabiting union ends (either through dissolution or marrige) varies with these characteristics. We considered associations with a range of variables similar to those in Table 5, but those in Table 8 are the only ones which approached statistical significance. The X2-statistics which are reported in Table 8 test for independence between the three-category response variable (remain childless in the union, have a child within the union or end the cohabiting union) and each explanatory variable.

The bivariate associations indicate that a first birth within a cohabiting union is more likely when the man is not employed (part (a) of Table 8), and it is also more likely if there are surprises in the financial situation, particularly negative ones (part (b)). Young women whose fathers were in semiskilled or unskilled manual jobs are much more likely to become mothers in a cohabiting union (part (c)). All these associations suggest that cohabiting couples who are in poorer financial circumstances are more likely to have a baby. The direction and significance of the associations in parts (a) and (c) are confirmed in a multivariate setting, in which we control for the duration of unions and age at the start of the cohabiting union.

6. Conclusions

This paper has used a new source of data to study the dramatic increase in cohabitation before




marriage, the stability of such unions and the role which cohabitation plays in the delay in marriage and motherhood. Long-term cohabitations are rare. It is the shift to cohabitation as the dominant mode of first live-in partnership (more than 70% of first partnerships) which is the primary contributor to the postponement of marriage, although partnerships are also being postponed in young people's lives, particularly for more recent cohorts reaching adulthood.

Cohabiting unions last only a short time before being converted into marriage or dissolving: their median length is about 2 years. About three in five cohabitations turn into marriage and 35% dissolve within 10 years. After a first cohabitation has dissolved, the median duration to the next cohabitation is 5 years.

The results of the analysis of union dissolution and conversion into marrige are consistent with the importance of new information, particularly information on a woman's partner's earnings, acquired subsequently to the formation of the union, for the outcome of cohabiting unions. Higher partner's earnings increase the chances of marriage and reduce the risk of dissolution. In conjunction with the relatively short time spent living together in cohabiting unions before either marrying his or her partner or the union dissolves, these results suggest that cohabitation is used while waiting to resolve uncertainties, to signal economic success and as a learning experience before stronger commitments are made, i.e. it is a way of coping with uncertainty.

Childbearing within cohabiting unions has become more common (about one in six unions involving never-married women), but these unions are much less likely to be converted into marriage and more likely to dissolve eventually than childless unions are. Such dissolutions create lone parent families headed by never-married mothers. Our finding that children are more likely to be born into unions which are doing less well financially suggests a tendency for the dissolution of unions to 'select' mothers in relatively unfavourable economic circumstances into lone parenthood. This tendency is reinforced by a higher rate of dissolution of unions among couples in which the male partner's earnings are lower.

Acknowledgements

We are grateful to the Economic and Social Research Council for financial support for this research and to two referees and the Joint Editor for helpful comments on a previous version of the paper.

Appendix A: The British Household Panel Study

The first wave of the BHPS, collected in the autumn of 1991, was designed as a nationally representative sample of the population of Great Britain living in private households in 1991. The achieved wave 1 sample covered 5500 households and corresponds to a response rate of about 74% of the effective sample size. At wave 1, about 92% of eligible adults, i.e. almost 10000 individuals, provided full interviews. The same individuals are reinterviewed each successive year, and if they split off from their original households to form new households all adult members of these households are also interviewed. Similarly, children in the original households are interviewed when they reach 16 years of age. Thus, the sample remains broadly representative of the population of Britain as it changes through the 1990s. Of those interviewed in the first wave, 88% were successfully reinterviewed at wave 2 (autumn 1992), and subsequent wave-on-wave response rates have consistently been around 95% (Taylor (1997), chapter IV, section 6, Tables 18-21).

The core questionnaire elicits information about income, labour market behaviour, housing conditions, household composition, education and health at each yearly interview. Information on changes




(e.g. employment, household membership and receipt of each income source) which have occurred within the households in the period between interviews is also collected.

The second wave (1992) obtained retrospective information on complete fertility, marital, cohabitation and employment histories for all adult panel members in that year, and the third wave collected detailed job histories. Further information can be obtained at

http: //www.iser.essex.ac.uk/bhps/doc/index.html

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the Family (eds C. Jonung and I. Persson). London: Routledge. Haskey, J. (1992) Pre-marital cohabitation and the probability of subsequent divorce: analyses using new data from

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Cohabitation in Great Britain: Not for Long, But Here to Stay