In the Context of Cohabitation, the National Survey of Family Growth Found That:
Census. 2008 Feb; 45(1): 129–141.
The Quality of Retrospective Data on Cohabitation
Abstract
We assess the quality of retrospective information on cohabitation by comparing data nerveless in four major U.S. family surveys: the National Survey of Families and Households and three rounds of the National Survey of Family Growth. Nosotros use event-history analysis to clarify rates of entry into cohabitation in age-flow-cohort segments captured by multiple surveys. Nosotros find consequent discrepancies among the four surveys. The pattern of differences suggests that cohabitation histories underestimate cohabitation rates in afar periods relative to rates estimated closer to the date of survey. We conclude with cautions regarding the use of retrospective data on cohabitation.
One of the defining characteristics of the field of demography is its attention to data and information quality. Demographers have used both checks of internal consistency and comparisons between different data sources to assess the accurateness of demographic measurement (e.k., Cherlin, Griffith, and McCarthy 1983; Raley, Harris, and Rindfuss 2000; Rendall et al. 1999; Swicegood, Morgan, and Rindfuss 1984; Wu, Martin, and Long 2001). When largescale surveys commencement began collecting retrospective demographic information, such as wedlock and birth histories, many demographers expressed doubt about the quality of these information. However, studies showed that in many contexts women reported births and marriages with a high level of accurateness. Thus, researchers embraced these life histories and came to rely on retrospective data for studies of family formation. As cohabitation has get more mutual, cohabiting relationships have been added to the event-history portions of major family surveys. In this paper, we compare data from four of these surveys—the National Survey of Families and Households and iii rounds of the National Survey of Family Growth—in order to assess the quality of these cohabitation data. We detect discrepancies among the four surveys consistent with the proffer that cohabitation histories underestimate cohabitation rates in distant periods relative to rates estimated closer to the date of survey.
MEASURING COHABITATION
Over the last one-half-century in the United States, cohabiting relationships between unmarried couples take become more and more mutual. Cohabitation was rare and stigmatized in the 1950s; by the turn of the twenty-first century, information technology had become accepted both every bit a precursor to marriage and equally a stand-lonely human relationship. This rapid growth spurred scientific interest in the characteristics of cohabiters and the role of cohabitation in contemporary family unit systems. Nonetheless, the collection of information on cohabitation lagged backside this interest (Casper and Cohen 2000; Smock 2000).
In the absence of directly measurement of cohabitation, researchers constructed the first national estimates of cohabitation using indirect measures based on household composition data from the U.S. census and from the Electric current Population Survey (CPS). Definitions based on the identification of "partners of the opposite sex sharing living quarters," or POSSLQ, allowed researchers to put together consequent time series that accurately reflect overall trends in cohabitation levels. However, indirect measures produce imprecise counts of cohabiters. Proposed adjustments to the measures meliorate their functioning, simply not enough to lucifer direct estimates of cohabitation (Casper and Cohen 2000).
Directly questions almost current cohabitation were introduced in the U.S. demography in 1990 and in the CPS and the Survey on Income and Program Participation (SIPP) in 1995 and 1996, respectively. Several family surveys conducted in the late 1980s also included direct measures of cohabitation, collecting retrospective cohabitation histories as well as current cohabitation status. The National Survey of Families and Households (NSFH), conducted in 1987 and 1988, was specifically designed to study nontraditional family forms, including stepfamilies, single-parent families, and cohabiting partners (among other purposes). Cohabitation status and histories were included in the 1988 wave of the National Survey of Family Growth (NSFG) and expanded in the 1995 and 2002 NSFG. More recently, surveys (notably the Delicate Families and Child Wellbeing Study) take begun to comprise more detailed questions on cohabiting relationships in guild to capture some of the variations in the meaning of cohabitation.
Because of the lack of direct historical data and the inaccuracy of indirect measures of cohabitation, researchers have relied on retrospective reports of cohabitation histories to describe the past prevalence of cohabitation (e.grand., Bumpass and Lu 2000; Smock 2000). However, the validity and reliability of retrospective data on cohabitation are in question. Bumpass and Lu (2000) reported that the NSFH and the 1995 NSFG produce similar estimates of the proportion of women e'er cohabiting and of exit rates from cohabiting relationships for the period 1980–1984. In contrast, the CPS and the SIPP, whose primary focus is labor market behavior rather than family construction, produce substantially lower estimates of cohabitation than the 1987 NSFH and the 1995 NSFG (Casper and Cohen 2000). In this article, we extend previous methodological enquiry on cohabitation past comparing information from iv major U.South. family surveys and by exploring three possible mechanisms for distortions in the reporting of cohabitation.
Recent research on cohabitation suggests that cohabiting relationships may be inherently difficult to measure. Some cohabiting unions are long-term, stable, "marriage-like" relationships, while others are temporary or on and off and may be entered into for the sake of reduced housing costs or convenience rather than as a long-term commitment (Chocolate-brown and Booth 1996; Bumpass and Lu 2000; Sassler 2004). Qualitative inquiry has shown that couples often move in together gradually, without a articulate commencement engagement, and may not have definite plans about the future of the relationship (e.g., Manning and Smock 2005). The status of a human relationship at whatever given fourth dimension may be cryptic, leading to reports of relationship start dates that differ between partners or over fourth dimension (Knab and McLanahan 2006; Teitler, Reichman, and Koball 2006).
Based on this research, we would look surveys to yield "noisy" or error-laden estimates of cohabitation prevalence. Respondents may differ in which relationships they consider cohabitations, and they may have difficulty remembering the exact start and end dates of past relationships. However, these problems do not necessarily imply that different surveys produce inconsistent estimated levels of cohabitation. If differences and inaccuracies in reporting cohabitations are random—or if biases are consistent over time and across surveys—comparable surveys should produce consequent reports of cohabitation.
Other possible sources of reporting error, which have non been fully investigated previously, could lead to differential estimates of cohabitation prevalence in unlike surveys. For example, the increasing social credence of cohabitation may increase the completeness of respondents' reporting in later on surveys. Goldscheider and Goldscheider (1994) argued that attitudes toward dwelling-leaving at the fourth dimension of the survey influence adults' reports of the age at which they left their parents' home in the past. If this mechanism applies to cohabitation likewise, later surveys would generate college estimates of cohabitation than earlier surveys, even for the same periods. The increased social credence of cohabitation in 2002 relative to 1988, for instance, may mean that cohabiting relationships that took identify in the 1980s are more likely to be reported by respondents in the 2002 NSFG than in the 1988 survey.
On the other hand, respondents may misreport or underreport events in the afar past relative to more recent events. Human memories are fallible; as time passes, people omit dates and events. Previous research has plant that events in the distant by are underreported relative to recent events, with the degree of underreporting increasing every bit the fourth dimension elapsed since the event increases (for a review of this literature, see Belli 1998; Wu et al. 2001). Unique or highly emotional events, such every bit the decease of a parent or a national disaster, and events whose dates are frequently referenced, such as marriages or birthdays, seem to exist less susceptible to this decay in reporting levels over fourth dimension (Brewer 1994; Thompson et al. 1996). Considering entry into cohabitation is often not clearly defined, and because cohabiting relationships vary in duration and importance, cohabiting relationships may not share this same resistance to omission over time.
OUTLINE OF THE ARTICLE
In this article, we pool data from four widely used national surveys that include cohabitation histories (the NSFH and Waves four–half-dozen of the NSFG) in order to determine whether data from these four sources are consistent. Our analytic approach is straightforward. We model the likelihood of starting cohabitation for unmarried women who were not cohabiting, controlling for the survey from which the observation was taken, and study the coefficients for the survey dummy variables. In order to make these comparisons, we select age-catamenia-cohort groups that are observed by more than than one survey. We take retrospective reports of cohabitation from the same nascence cohorts of women interviewed by different surveys and compare the human relationship histories generated by these retrospective reports.
We evaluate three possible sources of discrepancies: survey-specific bias, increasing social acceptability of cohabitation over time, and decreased reporting of relationships in the distant past. These mechanisms are hypothesized to produce different patterns of error. Survey-specific biases should produce stable discrepancies between surveys. Bias related to the social climate at the time the survey was administered should produce discrepancies between surveys that vary past the year of the interview. Bias related to retrieve error should produce discrepancies that vary by the time elapsed since the interview. In order to distinguish betwixt these mechanisms, we choose comparison groups to vary the combination of the time elapsed since the survey and the time the survey was administered.
We explicate our choice of analytic sample in more detail in the adjacent section, which also includes a clarification of the four surveys and of our modeling strategy. We then nowadays our results, followed by a brief discussion and conclusion.
Data AND METHODS
Overview of the Surveys
The context of the cohabitation questions in the iv surveys, likewise every bit details about sample size and characteristics, are presented in Table 1. All iv surveys measured cohabitation direct (equally opposed to using indirect measurement via household rosters) and recorded both current cohabitation status and past relationships.
Table 1.
Cohabitation Data in Four Surveys
| Survey | Sample Size | Sample Construction | Context of Cohabitation Question |
|---|---|---|---|
| 1988 NSFG | eight,450 women | Nationally representative of noninstitutionalized population; blackness women oversampled | Both formal and breezy marriages recorded Data collected on upward to 3 marriages (first, second, current/most recent) For each marriage, respondent was asked if cohabited with hubby earlier marriage Dates of cohabitation recorded for upwards to ane additional cohabitation |
| 1995 NSFG | 10,847 women | Nationally representative of noninstitutionalized population; blackness and Hispanic women oversampled | Merely formal marriages recorded Information collected on upwards to five marriages (no respondent married more five times) For each union, respondent was asked if cohabited with husband before wedlock Dates of cohabitation recorded for electric current cohabiting partner (if any) and upwards to three additional cohabitations |
| 2002 NSFG | 7,643 women (data from 4,928 men not used in this analysis) | Nationally representative of noninstitutionalized population; blacks oversampled | But formal marriages recorded Data collected on upwardly to half-dozen marriages (no respondent married more than six times) For each marriage, respondent was asked if cohabited with husband before spousal relationship Dates of cohabitation recorded for electric current cohabiting partner (if any) and upwards to eight additional cohabitations |
| NSFH | seven,790 women (information from 5,227 men not used in this analysis) | Nationally representative of population in households; blacks, Puerto Ricans, Mexican Americans, single-parent families, families with stepchildren, cohabiting couples, and recently married persons oversampled | Only formal marriages recorded Information collected on all marriages For beginning, second, and most recent marriages, respondent was asked if cohabited with spouse earlier marriage E'er-married respondents were asked about first premarital cohabitation, most recent cohabitation, and up to two cohabitations between marriages Never-married respondents were asked about kickoff cohabitation and virtually recent cohabitation All marriages and cohabitations between waves were recorded |
The NSFG series was designed to provide nationally representative estimates of pregnancy, birth rates, and contraceptive usage. In add-on to fertility information, some elements of matrimony and relationship history were collected in all vi surveys. The first moving ridge of the NSFG was fielded in 1973; subsequent surveys of independent samples of women took place in 1976, 1982, 1988, 1995, and 2002. Although there are variations in sample structure and questionnaire structure over fourth dimension, efforts were made to maximize comparability of data. Cohabitation questions were offset asked in 1982, simply these questions were limited in their usefulness. Rather than request about cohabitation as a separate relationship status, the 1982 NSFG labeled cohabitation equally "informal matrimony." Respondents reported their spousal relationship history and then were asked whether each marriage was formal or informal. Initial comparisons fabricated information technology clear that this approach leads to vast underestimates of cohabitation. Nosotros therefore begin our analysis with the 1988 NSFG. In this survey, ever-married women were explicitly asked if they had lived with their husband before getting married. These questions were repeated for each union. In addition, all respondents were asked if they had ever lived with someone whom they did not later marry, but human relationship get-go and end dates were nerveless for only 1 such cohabitation. The 1995 and 2002 surveys collected cohabitation histories for both premarital cohabitation and relationships that did non end in marriage. In 2002, men were included in the sample for the start time; here we limit our analysis to female respondents in gild to make comparisons across surveys.
The NSFH was first conducted in 1987–1988, with follow-up surveys of the original respondents in 1992–1994 and 2001–2002. Once again, nosotros limit our analysis to female respondents, although both men and women were interviewed. The survey oversampled currently cohabiting couples, families with stepchildren, and single-parent families. Cohabitation histories were collected from all respondents.
In the first wave of the NSFH, data on cohabiting relationships were nerveless separately from marriage data. The department on cohabitation began with the interviewer reading the comment, "Nowadays, many unmarried couples alive together; sometimes they eventually become married and sometimes they don't." E'er-married individuals were asked, for each matrimony, whether they lived with their spouse before getting married. They were as well asked virtually their first cohabiting relationship and about cohabiting relationships that did not atomic number 82 to matrimony. Never-married individuals were asked for the outset and terminate dates of their beginning cohabiting partnership and their electric current partnership (if any), and were asked how many other partners they had lived with. Subsequent waves of the NSFH asked nearly all cohabiting relationships that took place betwixt survey waves; respondents were too asked to correct information that they may take forgotten or misreported in before waves. Where applicative, we utilise this information to update data from the kickoff round. However, we exercise not analyze reports of cohabiting relationships that took place subsequently the first information drove in 1987. We therefore treat the NSFH as i survey carried out in 1987–1988.
In general, data collection methods in the four surveys were very similar. All interviews were conducted in person. There were no major differences in the phrasing of questions about cohabitation across surveys, and all 4 of the surveys used the terminology "living together" to ask about cohabiting relationships. The three NSFG defined a cohabiting human relationship as 1 in which the couple shares "the aforementioned usual address"; the NSFH provided no definition of cohabitation. None of the 4 surveys specified a length of time that a cohabiting relationship must final in order to exist reported. The information collected in the 1988 moving ridge of the NSFG were more than limited than those nerveless in other surveys. Offset and stop dates were nerveless for only one relationship that did not end in marriage in the 1988 wave, whereas up to four and 8 relationships were recorded in 1995 and 2002, respectively, and upwardly to two relationships were recorded in the NSFH. The 1988 NSFG may therefore underestimate cohabitation relative to the other surveys.
Analytic Approach and Construction of the Samples
Following Swicegood et al. (1984), we compare reports beyond different surveys by commencement selecting an age-period group that was observed in all four surveys. Secondary analyses treat historic period-period groups observed in at least 2 surveys. Based on the historic period range of the samples, some birth cohorts of women were eligible to be included in all 4 of the surveys. For example, women born in 1965 were 22 in 1987, 23 in 1988, 30 in 1995, and 37 in 2002, and so fell within the eligible age range for all four surveys. No individual women (that we know of) were interviewed past more than than one survey. However, because the surveys are nationally representative, they tin all be used to depict, in the aggregate, the beliefs of women born in 1965. We judge the reliability of the surveys by comparing their representations of cohorts observed by more one survey.
Each survey captures a different subset of the life experience of cohorts of women included in the survey. The 1988 NSFG, for instance, observes the 1965 birth cohort only until age 23, whereas the 2002 NSFH collects information upwards to age 37 for this same nativity cohort. Our analytic sample is therefore express by age and agenda year as well as past birth cohort.
We depict our analytic samples with reference to a Lexis diagram, shown in Effigy i. In this figure, the horizontal centrality represents calendar years, while the vertical axis represents age. Diagonal lines shown in the figure represent the life course experience of individuals or nascence cohorts who are the specified historic period during the designated agenda year. (Run across Preston, Heuveline, and Guillot 2001 for a more general description of Lexis diagrams.) This diagram runs from exact historic period eighteen to verbal historic period 45—the age range eligible to exist included in every survey. Each of the large triangles outlined in dashed lines shows age-catamenia-cohort groups observed by ane of the three waves of the NSFG. (The NSFH is omitted for the sake of visual clarity.) The right-hand edge of the triangle intersects the horizontal centrality at the year of the survey and runs from the youngest to the oldest eligible age. Each survey collects retrospective data from respondents; past events are located in the area to the left of the vertical edge. But just a subset of past events is observed by the survey: those that occurred to women who were in the eligible historic period range at the fourth dimension of the survey. The superlative (diagonal) border of the triangle represents the experiences of the oldest women in the survey and forms the upper leap of the observable events.
For this analysis, nosotros are interested in experiences that were eligible to be observed past more than one survey. Our basic question is whether eligible events were more or less likely to be reported in unlike surveys. These eligible events fall in the shaded triangles, which represent areas observed by more than than ane survey. The shaded triangle labeled Sample 1 is the intersection of all of the dashed survey triangles—that is, the age-menstruation-accomplice groups observed by all 4 surveys. (Note that the shaded triangle lies slightly below the superlative edge of the dashed triangles because the surveys were administered over a period of 8 to 15 months, rather than instantaneously as unsaid past this schematic diagram.) Sample 1, our principal analytic sample, consists of women born between Jan 1960 and December 1968, observed from age 18 until January 1987. The period covers the years 1978 to 1987. The maximum historic period in the age-period grouping is 27.
Nosotros brand no claim that cohabitation rates in this period are representative of rates across the whole time covered by the surveys. This sample is distinctive in several ways. Starting time, the women nosotros study are relatively young, and the sample is constructed such that more than of the early on experiences of these women are observed than their later experiences. This restriction is acceptable for a study of cohabitation because cohabitation is most prevalent among women in their teens and 20s. The young age of our sample does mean that the previously married women in our sample are probable to exist atypical; our findings about cohabitation amidst previously married women should be interpreted with caution. In add-on, our findings may not be generalizable to all women. For case, if cohabiting relationships among immature women are less stable, and therefore more bailiwick to recall error, discrepancies across surveys in our sample may exist larger than discrepancies among other age groups. A full consideration of the touch on of age at cohabitation on the reporting of cohabiting relationships is outside the telescopic of this commodity.
A second distinctive characteristic of the sample described above is the time elapsed between relationship feel and observation by the different surveys. Nosotros analyze experiences that take place between 1978 and 1987; women were interviewed in 1987, 1988, 1995, or 2002. We exploit this feature to distinguish betwixt possible sources of measurement mistake. The acceptability of cohabitation when the survey was administered, the time elapsed since the relationship, and fourth dimension-invariant survey furnishings could all influence the reported cohabitation rates. Based on this single sample, it is not possible to distinguish these effects on prevalence estimates. Nosotros therefore include two additional analytic samples, 1 focusing on an earlier time period than the original and ane focusing on a later menstruum. Considering of the differences in sample structure and survey design between the NSFH and the NSFG surveys, nosotros limit these additional samples to the three NSFG surveys.
Each of our secondary samples is constructed then that it has the aforementioned age construction as the original sample. Thus, nosotros compare reports from the birth cohorts of 1953–1961, from age 18 until 1980, from the 1988 and 1995 NSFG (Sample 3 in Effigy ane); and we compare reports from the birth cohorts of 1967–1975, from historic period xviii until 1994, from the 1995 and 2002 NSFG (Sample 2 in Figure 1). By comparing results from three dissimilar periods, we can appraise the plausibility of three dissimilar hypotheses. If there are no differences across surveys, or if differences are due to survey-specific furnishings, results should be the same in all samples. If the reporting of cohabiting relationships increases over time due to increased social acceptance of cohabitation, cohabitation prevalence should be college in the more recent survey for all samples. If the reporting of cohabiting relationships decreases with elapsed fourth dimension since the event because of recall error, the prevalence of cohabitation should exist lower in the more than recent survey for all samples.
Methods
We began past converting individual cohabitation and marriage histories from each survey into a file of person-months spent in each marital status (never-married and not cohabiting; previously married and not cohabiting; married; never-married and cohabiting; and previously married and cohabiting). Nosotros combined information from all iii waves of the NSFH into a single file. In the 1995 NSFG, cohabitation histories included any interruptions of cohabiting relationships in improver to first and end dates. To increase comparability between the surveys, we did not use this data in our outcome-history files merely recorded only the first date that couples moved in together and the terminal engagement they separated. Across all surveys, around 1%–2% of cohabiting relationships reported by respondents had missing start dates. Nosotros excluded these relationships from analysis but included other cohabitations and marriages reported by these women and person-months contributed by these women. All analyses were repeated excluding women with whatever missing data; results were similar to those reported here and are available from the authors on asking.
We use discrete-time issue-history analysis to predict the likelihood of entering a cohabiting relationship. We analyze all cohabiting relationships here; women who divorce or whose start cohabiting relationship dissolves return to the sample at risk. Fewer than v% of women in each survey reported multiple cohabiting relationships during the period covered by our assay. Specifically, we estimate the post-obit equation:
Here, P information technology represents the conditional probability of an private i entering into a cohabitation at time t, given that she was not cohabiting at that point. The baseline take chances of starting a cohabiting relationship is represented by α, while the β terms correspond coefficients for individual characteristics and for the survey that generated the observation. This model is non intended to be a noun model of entry into cohabitation. The chief contained variable of interest is the dummy variable for survey. If data from the four surveys are consistent, the coefficients for the survey dummy variables should be cipher. That is, the likelihood of entry into cohabitation for an private should not vary co-ordinate to the survey that collected her data. Nosotros also control for age, race, Hispanic origin, and agenda year in order to business relationship for differences in sample composition of the 4 surveys. Each of the surveys oversampled African American and Hispanic women at different rates, and the age structure of the samples differs slightly across surveys. Considering the likelihood of cohabitation varies past age and race, failing to account for differences in sample composition would atomic number 82 to misleading estimates of the differences between surveys. We tested more complex specifications of agenda time than the uncomplicated linear variable, including squared terms, log terms, and dummy variables for individual twelvemonth; but within the limited period in this analysis, alternate specifications did non affect the results.
In improver to oversampling based on race and indigenous origin, the NSFH likewise oversampled electric current cohabiters and recently married individuals. Thus, using the NSFH to calculate cohabitation rates without fully adjusting for sampling design overestimates rates of entry into cohabitation in the years immediately prior to the survey and may overestimate cohabitation rates in earlier periods if electric current cohabiters have a college propensity to cohabit. To account for sample design, we utilize sample weights for the NSFH. All models are estimated using SURVEYLOGISTIC in SAS ix.1.
Nosotros guess the basic model for three unlike subsamples as described in a higher place. We begin with the age-period group covered past all surveys, Sample 1 in Figure 1. Nosotros then repeat our analysis in both a later period (Sample ii) and an earlier period (Sample 3).
For each of these 3 subsamples, nosotros analyze cohabitation among never-married and previously married women separately. As nosotros noted earlier, the 1988 NSFG asked dates for but the beginning cohabiting human relationship that did not end in wedlock. Cohabiting relationships amongst divorced women may therefore exist disproportionately underreported in the 1988 NSFG because these relationships are less probable than relationships among never-married women to be first cohabitations. In addition, errors in the administration of the 2002 NSFG led to the failure to collect marriage end dates for some women. The sample of person-years at risk for previously married women in the 2002 NSFG is therefore unreliable. In addition to these technical issues, there are as well substantive reasons to believe that reporting differences may vary by marital status. Cohabiting relationships among divorced women tend to final longer and be more stable than relationships among never-married women (Bumpass, Sweetness, and Cherlin 1991). These relationships may therefore be perceived every bit more important by respondents, and they may be less discipline to recall mistake.
RESULTS
Tabular array 2 shows results from the first comparison, which includes data from all four surveys (Sample 1 in Figure one). Every bit we described earlier, historic period, race, Hispanic origin, and yr are included to command for differences in sampling blueprint beyond the four surveys but are non of substantive involvement. Nosotros include dummy variables for the 1995 NSFG, the 2002 NSFG, and the NSFH; observations from the 1988 NSFG are the omitted category.
Table 2.
Likelihood of Inbound Cohabitation as Measured in 4 Unlike Surveys: Discrete-Time Event-History Analysis With Logit Link
| Never-Married Women | Previously Married Women | |||||
|---|---|---|---|---|---|---|
| Coefficient | SE | t | Coefficient | SE | t | |
| Intercept | –6.xviii* | 2.45 | 2.5 | −fifteen.10* | 6.34 | 2.4 |
| Age (years) | 0.eleven | 0.22 | 0.5 | 0.63 | 0.53 | 1.2 |
| Historic period, Squared | 0.00 | 0.01 | 0.4 | –0.02 | 0.01 | ane.3 |
| African American | –0.22*** | 0.05 | four.iv | –0.66*** | 0.xix | 3.5 |
| Hispanic Origin (omitted: white non-Hispanic) | –0.xv* | 0.07 | 2.0 | –0.49** | 0.19 | ii.6 |
| Yr (omitted: NSFG 1988) | 0.00 | 0.01 | 0.0 | 0.06 | 0.04 | ane.6 |
| NSFG 1995 | 0.00 | 0.05 | 0.1 | –0.xi | 0.14 | 0.8 |
| NSFG 2002 | –0.26*** | 0.06 | 4.1 | –0.xvi | 0.17 | i.0 |
| NSFH | –0.09 | 0.07 | 1.2 | –0.12 | 0.16 | 0.viii |
| Observations (person–months) | 323,549 | 21,525 | ||||
| Events | 2,322 | 361 | ||||
| Log-Likelihood | 26,214 | 3,334 | ||||
Among never-married women, simply the coefficient for the 2002 NSFG is statistically different from zero. The fact that neither of the two other coefficients is different from null means that for the period 1978–1987 and for the specified ages and nascency cohorts, the 1988 NSFG, the 1995 NSFG, and the NSFH produce estimates of the likelihood of entering a cohabiting relationship that are equal, within the margins of sampling fault. The coefficient for the 2002 NSFG, on the other hand, is statistically significant (p < .001) and negative. That is, in this sample, the 2002 NSFG produces estimates of cohabitation for never-married women that are lower than those produced by the 1988 NSFG (and, by implication, the other surveys likewise). The coefficient for the 2002 survey is −0.26, which implies that rates of entry into cohabitation for never-married women would announced almost 23% lower using the 2002 survey than the 1988 survey (one – e −0.26 = 1 – 0.77 = 0.23). Informal life-table calculations bear witness that this reduction in the probability of entering cohabitation at all ages would reduce the proportion of women e'er cohabiting by age 27 (the upper age limit of our sample) by approximately 15%.1
The coefficients for previously married women are in the same management as for never-married women, only none of the coefficients for the survey variables is statistically meaning. Information technology is worth noting, withal, that the sample size is much smaller for previously married women than for never-married women, and standard errors are therefore larger; it may be that these surveys do not provide sufficient statistical power to observe differences in the reporting of cohabitation for previously married women for these samples.
The finding that the 2002 NSFG produces lower cohabitation rates than the other surveys in this menstruum is consistent either with a survey consequence specific to the 2002 wave of the NSFG or with a pattern of increasing omission of cohabiting relationships equally the time between the event and the survey increases. Results from the 2 additional samples provide additional relevant testify. These results (shown in Tables three and 4) are generally consistent with the hypothesis that women are more probable to omit cohabitations in the more distant by.
Table three.
Likelihood of Entering Cohabitation as Measured in the 1995 and 2002 NSFG: Detached-Fourth dimension Consequence-History Analysis With Logit Link
| Never-Married Women | Previously Married Women | |||||
|---|---|---|---|---|---|---|
| Coefficient | SE | t | Coefficient | SE | t | |
| Intercept | –sixteen.22*** | 2.98 | v.4 | –18.85 | 10.09 | 1.nine |
| Age (years) | 0.60* | 0.26 | 2.3 | 1.28 | 0.87 | one.5 |
| Age, Squared | –0.01* | 0.01 | 2.two | –0.03 | 0.02 | 1.4 |
| African American | –0.sixteen** | 0.06 | 2.half dozen | –1.28*** | 0.31 | 4.2 |
| Hispanic Origin (omitted: white not-Hispanic) | 0.03 | 0.08 | 0.four | –0.57* | 0.25 | ii.3 |
| Year (omitted: NSFG 1995) | 0.06*** | 0.01 | 4.1 | 0.01 | 0.06 | 0.1 |
| NSFG 2002 | –0.14** | 0.05 | 2.half-dozen | –0.23 | 0.17 | 1.3 |
| Observations (person-months) | 173,709 | 7,040 | ||||
| Events | i,490 | 152 | ||||
| Log-Likelihood | 17,099 | 1,434 | ||||
Tabular array four.
Likelihood of Inbound Cohabitation every bit Measured in the 1988 and 1995 NSFG: Discrete-Time Event-History Analysis With Logit Link
| Never-Married Women | Previously Married Women | |||||
|---|---|---|---|---|---|---|
| Coefficient | SE | t | Coefficient | SE | t | |
| Intercept | –xvi.00*** | iii.41 | 4.7 | –14.55 | 8.26 | 1.8 |
| Age (years) | 0.56 | 0.31 | 1.eight | 0.91 | 0.69 | 1.three |
| Historic period, Squared | –0.01 | 0.01 | 1.8 | –0.02 | 0.02 | 1.iv |
| African American | –0.11 | 0.06 | one.7 | –0.60*** | 0.18 | iii.3 |
| Hispanic Origin (omitted: white not-Hispanic) | –0.15 | 0.eleven | 1.three | –0.90** | 0.31 | 2.9 |
| Year (omitted: NSFG 1988) | 0.07*** | 0.02 | four.4 | 0.02 | 0.05 | 0.4 |
| NSFG 1995 | –0.16** | 0.06 | 2.vii | 0.17 | 0.xiv | 1.2 |
| Observations (person-months) | 196,947 | 15,678 | ||||
| Events | 1,186 | 220 | ||||
| Log-Likelihood | 14,453 | 2,281 | ||||
Table 3 compares the 1995 and 2002 NSFG (Sample two in Figure ane), using the 1995 survey as the omitted category. For both never-married women and previously married women, the coefficient for the 2002 NSFG is negative: cohabitation rates based on the more than recent survey are lower than rates based on the earlier survey, which use observations from periods closer to the survey date. The departure between the two surveys is smaller than in the previous model for never-married women and larger for previously married women. Over again, the coefficient for previously married women is non statistically different from zero. As noted earlier, at that place were errors in the collection of spousal relationship histories for some previously married women in the 2002 NSFG, rendering the results for these women suspect.
Table 4, comparing the 1988 and 1995 NSFG (Sample 3 in Effigy 1), shows mixed results. For this sample—betwixt 1971 and 1980, for a subset of age groups and birth cohorts— the 1995 NSFG produces significantly lower cohabitation rates for never-married women than the 1988 NSFG. The coefficient representing the difference between these ii surveys is −0.xvi, which translates to rates of entry into cohabitation most fifteen% lower using the later survey. For previously married women, we again observe a large divergence that is non statistically significant. In this example, cohabitation rates are college in the later on survey (1995) than in the before survey (1988). This finding may result from sampling fault, or it may be that the differences in the number of cohabitations recorded in the 1988 survey are more salient during this period than during the menses shown in Table 2.
DISCUSSION
Our results tin be summarized as follows: (one) in express comparisons, we find no statistically significant differences between the 1988 NSFG and the NSFH; (2) differences across surveys for previously married women are never statistically significant but are sometimes big and generally in the direction of lower rates in subsequently surveys; and (three) for never-married women, later surveys produce significantly lower cohabitation rates in three of four comparisons. We concentrate our discussion on this 3rd event. Given the problems with data drove in the 2002 NSFG, our results for never-married women in that survey are more trustworthy than the results for previously married women. In addition, because of the construction of our sample, our analysis is skewed toward younger women and therefore represents early cohabitation feel improve than later postmarital cohabitation feel.
Table 5 summarizes the results for never-married women from Tables 2, 3, and 4. These coefficients represent comparisons between adjacent surveys—the 1995 versus the 1988 NSFG and the 2002 versus the 1995 NSFG—for three separate periods. We find lilliputian back up for the hypothesis that the reporting of past cohabiting relationships increases as the social acceptability of cohabitation increases: three of the four coefficients are statistically significant and negative. Because nosotros detect differences betwixt the 1995 and 1988 NSFG and between the 2002 and 1995 surveys, we do not believe the discrepancies are completely attributable to survey-specific effects, although the variation in the between-survey differences implies possible biases item to one of the surveys or periods.
Table 5.
Summary of Results for Never-Married Women
| NSFG Surveys Compared | Period of Analysis | |||||
|---|---|---|---|---|---|---|
| 1971–1980 (three) | 1978–1987 (1) | 1985–1994 (ii) | ||||
| Coefficient | SE | Coefficient | SE | Coefficient | SE | |
| 1995 and 1988 | –0.sixteen** | 0.06 | 0.00 | 0.05 | ||
| 2002 and 1995 | –0.27*** | 0.06 | –0.14*** | 0.05 | ||
The preponderantly negative signs of the coefficients are most consistent with the hypothesis that women underreport relationships they had before the survey. Given the minor number of coefficients, we cannot draw firm conclusions almost the functional form of underreporting with time. It appears that the difference betwixt surveys increases as the fourth dimension elapsed since the survey increases. This design suggests that non simply do women neglect to report cohabitations every bit time passes, simply they are more than probable to omit relationships that are more distant in time.
We find differences betwixt surveys on the order of 15%–20% for periods 15–20 years before the survey. Based on these results, we urge caution in the utilize of retrospective cohabitation histories. The underreporting of early cohabiting relationships within surveys poses bug for the analysis of the relationship between early cohabitation and later life events, such as marital stability or health. Using information from more than than one survey to produce time-series estimates of cohabitation also requires care, although it might be possible to correct for cross-survey differences. Alternatively, these cantankerous-survey differences could be exploited for further methodological research.
Our findings are express to the particular subsamples we employ for assay. For instance, our enquiry overrepresents the feel of young women relative to experience at older ages. If cohabiting relationships amid young women are more than likely to be short-term and unstable than cohabiting relationships of older women, the former may be especially susceptible to omission or forgetting. In constructing our sample, our primary concern was to facilitate comparison of multiple surveys. It would be possible to carry out a more detailed comparison of two surveys that might shed lite on differential omission of cohabiting relationships to women in different age groups.
This research likewise focuses on the average differences in cohabitation rates across surveys. Differences may exist larger or smaller for particular groups of women. For example, educated women may report relationships more consistently than less educated women; cohabitations that deliquesce may be more subject to underreporting than cohabitations that lead to marriage. Farther research focusing on these possibilities could illuminate both issues with data and variations in the salience and importance of cohabitation to individuals.
Acknowledgments
Nosotros thank session participants and attendants, Larry Wu, the editors of Demography, and three anonymous reviewers for helpful comments on previous versions.
Footnotes
This research was supported past Duke University and by the National Institute of Child Health and Man Development, Grants F32 HD050032-01 (Kirschstein Postdoctoral Fellowship, Hayford) and R01 HD41042 (Morgan). An earlier version of this article was presented at the 2006 annual meetings of the American Sociological Association in Montreal, Quebec.
1.Our sample includes both women who accept previously cohabited and women who have never cohabited, and our model estimates average probabilities across these 2 groups. We constructed life tables applying these predicted probabilities to women who had never cohabited to model entry into first cohabitation.
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Source: https://www.ncbi.nlm.nih.gov/pmc/articles/PMC2722926/
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