In the years since the Great Recession, social scientists have anticipated that economic recovery in the United States, characterized by gains in employment and median household income, would augur a reversal of declining fertility trends. However, the expected post-recession rebound in fertility rates has yet to materialize. In this study, I propose an economic explanation for why fertility rates have continued to decline regardless of improvements in conventional economic indicators. I argue that ongoing structural changes in U.S. labor markets have prolonged the financial uncertainty that leads women and couples to delay or forgo childbearing. Combining statistical and survey data with restricted-use vital registration records, I examine how cyclical and structural changes in metropolitan-area labor markets were associated with changes in total fertility rates (TFRs) across racial/ethnic groups from the early 1990s to the present day, with a particular focus on the 2006–2014 period. The findings suggest that changes in industry composition—specifically, the loss of manufacturing and other goods-producing businesses—have a larger effect on TFRs than changes in the unemployment rate for all racial/ethnic groups. Because structural changes in labor markets are more likely to be sustained over time—in contrast to unemployment rates, which fluctuate with economic cycles—further reductions in unemployment are unlikely to reverse declining fertility trends.
The United States experienced an economic recession from December 2007 to June 2009 that had manifold negative consequences on the national and global economy (Bureau of Labor Statistics 2012). Approximately 8.8 million U.S. jobs disappeared, and the U.S. unemployment rate rose to a peak of 9.5 % in 2009 (Goodman and Mance 2011; Bureau of Labor Statistics 2012). All told, the economic damage wrought by the Great Recession resulted in the loss of 19.2 trillion dollars in U.S. household wealth (U.S. Department of the Treasury 2012).
In the decade since, demographers have leveraged the Great Recession’s adverse economic consequences to examine how economic conditions influence aggregate fertility trends (Sobotka et al. 2011). These studies have examined economic and labor market variation at the national and state level and found differential outcomes in fertility behavior based on characteristics such as race and ethnicity, socioeconomic status, marital status, and educational attainment (Cherlin et al. 2013; Currie and Schwandt 2014; Percheski and Kimbro 2014; Schneider and Hastings 2015). Researchers interpret these results as indications that women and couples temporarily postpone childbearing as a result of financial uncertainty induced by job loss and decreases in income (Balbo et al. 2013; Brauner-Otto and Geist 2018). Delays in childbearing—even for a few years—can reduce lifetime fertility rates for cohorts of women (Bongaarts and Feeney 1998; Currie and Schwandt 2014; Morgan 2003; Ryder 1980). As a result, both brief and sustained changes in macroeconomic conditions have long-term implications for population change and renewal.
Despite the importance of these previous studies, a puzzle remains concerning trends in post-recession fertility: although unemployment rates and median household income have returned to pre-recession levels, fertility rates continue to decline and have recently fallen to record lows in the United States (Hamilton et al. 2017; Martin et al. 2018). The total fertility rate (TFR) for the United States remained below replacement at about 1.73 births per woman in 2018 (Hamilton and Kirmeyer 2017; Hamilton et al. 2019), down from 2.12 births per woman in 2007. Meanwhile, the national unemployment rate returned to its pre-recession level of 4.7 % in May 2016 (Bureau of Labor Statistics 2017), and median household income surpassed pre-recession levels in 2015 (Semega et al. 2017). These patterns raise questions about the relative importance of cyclical economic conditions on fertility behavior and decision-making. If employment and earnings have improved, why do fertility rates continue to decline? Answering this question not only helps explain the contemporary puzzle of why fertility rates have deviated from their historically cyclical trend but also contributes insight into how longer-term economic processes shape U.S. fertility.
In this study, I propose a structural economic explanation for why fertility rates continue to decline in the United States regardless of improvements in conventional economic indicators since the Great Recession. I argue that ongoing structural changes in the industry composition of U.S. labor markets have prolonged the financial uncertainty that leads women and couples to delay or forgo childbearing. Since the 1980s, the U.S. labor market has experienced declines in middle-skill, middle-income jobs in manufacturing and other goods-producing industries as a result of improvements in assembly line automation and the displacement of routine production jobs offshore, among other explanations (Acemoglu and Autor 2011; Autor et al. 2006; Kalleberg 2009). This process of “labor market polarization” has reduced employment demand for jobs in the middle of the occupational skills distribution while increasing employment demand for low- and high-skill service sector positions. Figure 1 displays the declining share of employment, total annual payroll, and business establishments concentrated in goods-producing industries from 1987 to 2014. The graph shows that the share of workers employed in goods-producing industries declined from 29 % in 1987 to 14.7 % in 2014. This trend parallels the disappearance of goods-producing business establishments from U.S. labor markets. Economic explanations of demographic behavior should account for structural changes of this magnitude.
Combining statistical and survey data with restricted-use vital registration records, I comparatively examine how cyclical and structural changes in metropolitan-area labor markets were associated with changes in TFRs across racial/ethnic groups from the early 1990s to the present day, with an emphasis on the period 2006–2014. I focus on variation in TFR across racial/ethnic groups for two reasons. First, the variation in fertility trends across racial/ethnic groups over this period is particularly striking and may reflect the differential economic impacts of the Great Recession in domains ranging from job loss, wealth loss, and housing foreclosure (Hall et al. 2015; Hout and Cumberworth 2012; Pfeffer et al. 2016). Second, historical inequalities fundamentally shape the distribution of contemporary labor market opportunities (Browne 2000; Jaret et al. 2003). Because occupational distributions vary by race and ethnicity, labor market polarization should be expected to influence fertility behavior more for groups with larger shares of workers in goods-producing industries.
I focus this analysis on local geographic areas, specifically all 381 metropolitan statistical areas (MSAs) in the United States, because the distribution of labor market outcomes varies considerably across geographic areas (Levine 2012; U.S. Department of Labor 2012). Past studies on how economic conditions influence fertility behavior in the United States mostly focus on state- or national-level trends. However, within states, economic conditions and job opportunities vary substantially across metropolitan areas. Furthermore, family formation and fertility are processes that occur primarily at the local level. Social networks, which are mostly concentrated within local geographic areas, are influential in setting the norms of family and fertility decision-making decisions (Arai 2007; Balbo and Barban 2014; Bernardi and Klärner 2014).
The findings of the present study suggest that structural changes in U.S. labor markets have a larger effect on TFR than changes in unemployment rates for all racial/ethnic groups during the 2006–2014 period. Given that changes in labor market polarization are more likely to be sustained over time—in contrast to unemployment rates, which fluctuate with economic cycles—the results indicate that structural characteristics of U.S. labor markets are as important, if not more important, than short-term swings in unemployment in predicting decreased levels of TFR in recent years. The findings also indicate that Hispanic fertility is substantially more responsive to the loss of goods-producing businesses than white, black, and Asian fertility. Importantly, the findings from this study help explain why fertility rates continues to decline in the United States despite the presence of indicators of employment and income that would otherwise predict a return to pre-recession fertility levels. In addition to making a theoretical contribution to the demographic literature on macroeconomic change and aggregate fertility behavior, the results of this study indicate that further reductions in unemployment are unlikely to reverse declining fertility trends.
Cyclical Trends in Economic Conditions and Fertility Behavior
The 2007–2009 financial crisis led to immediate reductions in fertility in the United States and Europe (Hamilton and Sutton 2012; Matysiak et al. 2014). Researchers theorize a procyclical relationship between economic conditions and fertility behavior: fertility rates increase during economic expansions and decrease during economic recessions. As economic prosperity diminishes during cyclical downturns and financial uncertainty and distress increases, women and couples postpone childbearing (Balbo et al. 2013; Buckles et al. 2018; Currie and Schwandt 2014). Such a perspective is consistent with classical economic theory on fertility that regards children as investments in either future income or future satisfaction (Becker 1960). In economically developed contexts where satisfaction is the primary aim of having children, parents are more likely to afford the costs of bearing and raising children as incomes rise (Sawhill 1977). Other theoretical perspectives posit a countercyclical relationship between unemployment and fertility (Butz and Ward 1979; Ermisch 1988). As female unemployment rises during economic recessions, researchers theorize an increase in fertility because women’s time out of the labor force lowers the opportunity costs associated with having a child. However, few empirical studies have supported this theoretical perspective in contemporary U.S. or European contexts (Cherlin et al. 2013; Morgan et al. 2011).
Most of the literature on how economic conditions influence fertility behavior have used measures of unemployment and median household income to proxy changes in economic conditions. Morgan and colleagues’ (2011) analysis of state-level economic and fertility trends during the Great Recession found that larger increases in unemployment are associated with larger decreases in fertility rates. Similarly, Schneider’s (2015) analysis of county- and state-level fertility trends throughout the Great Recession found that both national and state trends in unemployment are negatively associated with general fertility rates (GFR).
Two other studies of the Great Recession observed similar procyclical patterns pertaining to changes in unemployment and income. First, Currie and Schwandt’s (2014) analysis of state-level fertility trends between 1975 and 2010 found that increased unemployment during economic recessions is associated with short-term and long-term reductions in fertility for women in their early 20s. Second, a study by Cherlin et al. (2013) likewise found variation across age groups during the Great Recession: while younger women experienced decreases in age-specific GFR, women in their 40s continued having children at the same pre-recession age-specific GFR levels. As a result, the authors argued that economic cycles differentially influence fertility behavior across age groups.
If the fertility response to the Great Recession was in fact procyclical, then the post-recession recovery period should have predicted increased fertility rates in the same way that the recessionary period predicted decreased fertility rates. However, fertility rates have yet to return to pre-recession levels. The analysis by Cherlin et al. (2013) indicated an 11 % decrease in TFR between 2007 and 2011—a drop from 2.1 to 1.9. By 2013, six years after the onset of the Great Recession, the national TFR remained below the level of population replacement (Martin et al. 2017). Only in 2014 did the TFR briefly reverse course and increase before decreasing yet more through 2018 (Hamilton and Kirmeyer 2017; Hamilton et al. 2019). This ongoing fertility decline in the United States is at odds with slow yet steady improvements in economic conditions. Nationally, median household income returned to pre-recession levels in 2015, and the unemployment rate returned to pre-recession levels in 2016 (Bureau of Labor Statistics 2017; Semega et al. 2017 ).
Structural Change in U.S. Labor Markets
Although theories on economic uncertainty and fertility behavior focus almost exclusively on changes in cyclical economic indicators, broader structural changes in U.S. labor markets might help explain the persistence of declining fertility rates over the past decade. Critically, the United States has experienced an economic restructuring that has reshaped the composition of industries that provide jobs to American workers (Kalleberg 2009). Since the 1980s, middle-skill, middle-income jobs in manufacturing and other goods-producing industries have steadily disappeared while service sector positions have rapidly expanded. Although there are many explanations for this labor market restructuring (Janoski et al. 2014), researchers have identified technological advances in assembly line automation and the offshoring of routine production jobs as major causes (Autor et al. 2006; Kalleberg 2009). This process of labor market polarization has reshaped the economic opportunities available for workers, especially for those who have less than a college education, by shifting the types of jobs available to workers away from stable middle-income positions (Autor and Dorn 2013). At the same time, employment demand has grown for low-wage positions requiring little education or training as well as for high-wage positions requiring a college degree. However, the availability of middle-skill, middle-income jobs—long a pathway to middle class economic stability for workers with only a high school degree—has steadily declined (Newman and Winston 2016).
My theoretical argument is motivated in part by recent empirical findings by Autor et al. (2017) and Kearney and Wilson (2018), who linked long- and short-term industrial changes in U.S. labor markets to changes in fertility rates. Examining the rise of manufacturing import penetration from China between two periods, 1990–2000 and 2000–2014, Autor et al. (2017) found that commuting zone–level import shocks were associated with decreased fertility rates for U.S. women in their 20s and 30s. In contrast to negative economic shocks, Kearney and Wilson (2018) found that the growth in local-area hydraulic fracturing production between 1997 and 2012—a positive economic shock—was associated with increases in birth rates for women aged 18–34. These two studies emphasize how the gain and loss of specific industries, in contrast to overall levels of employment, in local labor markets are associated with substantive changes in fertility rates. The findings from these studies also suggest that structural economic changes impact fertility behavior at the community level, influencing perceptions of anticipated growth or decline of industrial sectors.
I focus this analysis on the decade surrounding the Great Recession because economic downturns have historically played a critical role in accelerating the displacement of middle-skill jobs and goods-producing businesses from U.S. labor markets. However, the modeling strategy also accounts for long-term structural change with an additional analysis that extends back to the early 1990s. Importantly, after manufacturing and other goods-producing jobs disappear, they rarely return during post-recessional recovery periods (Abel and Deitz 2012). For instance, Jaimovich and Siu (2012), examining rebounds in job loss across occupational categories after every economic recession in the United States since the 1970s, found that an overwhelming proportion of permanent job loss in middle-skill positions since the 1980s occurred during periods of economic recession rather than periods between economic recessions. The Great Recession was no exception to this trend: between 2007 and 2009, employment in occupational categories of production, craft, and repair decreased by 17 %; employment in occupational categories of operators, fabricators, and laborers decreased by 15 % (Autor 2010). The data used in the present analysis indicate that the average percentage of goods-producing businesses in MSAs decreased from 16.66 % in 2006 to 14.24 % in 2014, a 2.42 percentage point decline (Fig. 1). Overall, economic recessions have hastened permanent structural changes in U.S. labor markets and have reduced the demand for middle-skill, middle-income jobs.
The present analysis relies on population-level data and cannot test how individual-level factors influence fertility decision-making. One plausible mechanism through which this macroeconomic relationship may operate is individual-level financial uncertainty, initiated by job displacement from manufacturing, construction, and other goods-producing industries. Workers displaced from positions in goods-producing industries are increasingly tasked with finding new work in service-providing industries, which often provide fewer hours, lower pay, and scarcer benefits (Bureau of Labor Statistics and U.S. Department of Labor 2016; Janoski et al. 2014; Kalleberg 2009).1 Valletta and van der List (2015), for instance, found structural changes in industry composition to contribute to involuntary part-time work in the years following the Great Recession. Reemployment for many who lost jobs during the Great Recession did not necessarily return economic security and stability to pre-recession levels (Janoski et al. 2014; Moretti 2012). This mechanism is consistent with findings on the association between work precarity and family formation and fertility (Brauner-Otto and Geist 2018; Lim 2017; Modena and Sabatini 2012; Piotrowski et al. 2015; White and Rogers 2000; Yu and Sun 2018).
Racial/Ethnic Differences in Outcomes During the Great Recession
The aggregate trend of decreased fertility in the United States during the Great Recession conceals substantial variation across racial/ethnic subgroups. First, Hispanic women experienced large decreases in fertility following the start of the Great Recession (Martin et al. 2014). Cherlin et al. (2013), for instance, found larger decreases in past-year births for Hispanic women than non-Hispanic women between 2008 and 2011. Second, black and Asian/Pacific Islander fertility rates decreased more sharply than white fertility rates but less than Hispanic fertility rates. According to national vital statistics records on fertility between 2007 and 2015 (Hamilton et al. 2017), GFR decreased 3 % for non-Hispanic whites, 10 % for non-Hispanic blacks, 10 % for Asian/Pacific Islanders, and 36 % for Hispanics.
The variation in fertility trends across race/ethnicity throughout this period is particularly striking and may reflect differential economic impacts experienced in domains such as housing foreclosure (Hall et al. 2015; Rugh 2015), wealth loss (McKernan et al. 2014; Pfeffer et al. 2016), and job loss (Holder 2017, 2018: chapter 3; Hout and Cumberworth 2012). McKernan et al. (2014), for instance, found that although white families experienced an average decrease in family wealth of approximately 26.2 % following the Great Recession, black and Hispanic families lost 47.6 % and 44.3 %, respectively.
These differential economic consequences of the Great Recession across race/ethnicity also reflect a broader history of labor market stratification. Although black occupational and socioeconomic mobility increased throughout the middle of the twentieth century, the economic downturn of the late 1970s largely constrained upward occupational mobility in the years to follow (Bound and Dresser 1999; Pattillo 2013). This constrained upward mobility is reflected in the structure of educational and occupational networks, which act as barriers to high-level professions for black and Latino workers (Catanzarite and Aguilera 2002; Cohen and Huffman 2007; Kmec 2003; Royster 2003). Racial differences in labor market occupational distributions are also attributed to employer discrimination (Pager and Pedulla 2015). Indeed, research on occupational shifts of African American men during the Great Recession by Holder (2017) indicates that black men experienced declines in representation in high-wage and middle-wage occupational categories between 2005–2006 and 2010–2011, while non-Hispanic white men experienced only minor shifts in their representation in the same occupational categories.
Because occupational distributions vary across race/ethnicity, labor market polarization should be expected to influence fertility behavior more for groups with larger shares of workers in goods-producing industries. Figure 2 displays the percentage of workers aged 15–64 who were employed in goods-producing industries by racial/ethnic group between the years 2006 and 2014. Throughout this nine-year period, Hispanics had the highest percentage of workers employed in goods-producing industries, followed by whites. Blacks and Asians had the smallest percentage of workers in goods-producing industries.
The statistical approach used in the present study estimates separate regression models for racial/ethnic subgroups and tests for differences across these racial/ethnic subgroups. I expect that macroeconomic conditions, both cyclical and structural, will differentially influence fertility behavior across race and ethnicity. Given the variation in the level of employment in goods-producing industries across racial/ethnic groups, we should expect the loss of goods-producing businesses following the Great Recession to have a disproportionate impact on racial/ethnic groups with larger shares of workers in goods-producing industries.
Data and Methods
The present study is based on 31.5 million county-level birth certificate records between 2006 and 2014 from the restricted-use natality detail file from the National Vital Statistics System (NVSS).2 The restricted-use NVSS natality detail file provides a full enumeration of county-level births in the United States, and includes information on mother’s age, race, and Hispanic origin, which allow for the calculation of stratified racial/ethnic TFR values (National Center for Health Statistics 2016). County-level birth certificate records are aggregated up to the metropolitan area to match MSA designations defined by the Office of Management and Budget in 2013. MSA-level population counts are obtained for each year of the analysis from the National Center for Health Statistics’ bridged-race population estimates, which are derived from U.S. Census Bureau population estimates. Female population counts, stratified by five-year age groups, racial/ethnic group, and county, are aggregated up to the MSA level.
Data on MSA-level business establishments are accessed from the U.S. Census Bureau’s County Business Patterns (CBP) program. Business establishments are physical locations (1) with paid employees and (2) where business activity is conducted. Companies can have multiple business establishments with different numbers of employees. The CBP data provide a complete enumeration of U.S. business establishments by industry classification codes, which are primarily derived from Internal Revenue Service administrative records. Business establishments are coded according to the Standard Industrial Classification System (SIC) between 1990 and 1997, and according to the North American Industry Classification System (NAICS) between 1998 and 2014. I use the Bureau of Labor Statistics’ designations of NAICS domains to code businesses as either goods-producing or service-providing (Table 1).3 The domain level consists of aggregated industry supersectors. Industry supersectors within the goods-producing domain include natural resources extraction and mining, construction, and manufacturing.
MSA-level unemployment data are accessed from the Bureau of Labor Statistics’ Local Area Unemployment Statistics program (LAUS). MSA-level per capita gross domestic product (GDP) are accessed from the Bureau of Economic Analysis (BEA). Table 2 displays a full summary of measures and corresponding data sources.
Finally, I calculate metropolitan-area statistics for education, past-year migration, homeownership, and marital status from the one-year American Community Survey (ACS) between 2006 and 2014 for the overall MSA-level population as well as for racial/ethnic subgroups. ACS data are accessed through the IPUMS-USA database at the University of Minnesota (Ruggles et al. 2017). The smallest identifiable geographic unit in the one-year public-release ACS estimates are public-use microdata areas (PUMAs), which have a minimum of 100,000 residents. As a result, ACS estimates used to generate covariates are available for only MSAs with greater than 100,000 residents.4
The use of local-area population estimates in conjunction with birth records to calculate TFR generates a few implausible TFR values. Online appendix Fig. S1 displays a histogram with all TFR values generated through Eq. (1). Of 13,709 total TFR observations, 159 exceed 4.0. I remove these cases from the analytic sample to ensure that outliers are not influencing the regression results. The inclusion of these values in the models, however, does not substantively change the results.
Labor Market Polarization
I operationalize labor market polarization as a relative measure: the percentage of goods-producing business establishments in an MSA. Past research on labor market polarization has established that the goods-producing business sector contains a disproportionate share of middle-skill, middle-income jobs and therefore represents an appropriate measure of local labor market polarization (Abel and Deitz 2012). Additionally, because the availability of goods-producing jobs in a local area is a matter of supply, the distribution of business establishments—rather than the distribution of workers across occupations—is the preferred measurement of structural labor market conditions. In the online appendix (Text S1), I further discuss the conceptual and data limitations of substituting a worker-based measure for the business establishment measure.
I calculate the percentage of goods-producing businesses for each MSA by aggregating CBP business establishment data by domain classification codes for each year: either goods-producing industries or service-providing industries. I then lag this measure by one year to approximate labor market structure at the time of conception each year rather than the time of birth. As an additional robustness check, I test an absolute measure of labor market polarization: the number of goods-producing businesses in an MSA, log transformed to address its nonnormal distribution.
I measure MSA-level unemployment using a one-year lagged annual measure of the unemployment rate from the LAUS data set. The unemployment rate is calculated by dividing the number of people seeking employment by the total number of people in the civilian labor force. I test the sensitivity of this measure by estimating models that substitute the unemployment rate for (1) the employment-to-population ratio for workers ages 15–64 using ACS data, and (2) sex-specific measures of the unemployment rate and employment-to-population ratio for workers ages 15–64 using ACS data.
The analytic method used in the present analysis, fixed effects, controls for time-invariant unobserved heterogeneity that enters the specifications linearly and additively (Allison 2009). Accordingly, fixed characteristics of a metropolitan area are inherently controlled for in this analysis.5 Fixed-effects analyses, however, do not account for time-varying sources of unobserved heterogeneity. To minimize threats to causal inference, I include a set of theoretically relevant covariates that are lagged one year (Table 2).
In the full MSA analysis, covariates include year fixed effects, a logged measure of population size to adjust for annual changes in metropolitan-area population size, and per capita GDP to adjust for economic growth.
In robustness models using data from the ACS, I additionally adjust for aggregate measures of educational attainment, past-year MSA in-migration, homeownership, and marital status. These covariates are first calculated at the MSA level for adults aged 15–64, and in a further sensitivity check are recalculated for adults aged 15–64 for each racial/ethnic group.
These time-varying covariates have theoretical importance for considering how MSA-level economic and noneconomic indicators might influence race/ethnicity–specific fertility rates. First, educational attainment is a well-documented determinant of fertility (Rindfuss et al. 1996): college attendance for women is associated with lower fertility levels (e.g., Brand and Davis 2012). I account for educational attainment by adjusting for the percentage with more than a high school education. Second, births in an MSA are both to women residing in the area prior to conception and to women residing there after conception. Consequently, the TFR of a population is partially influenced by births that were conceived prior to the mothers’ entering the population. This makes adjusting for MSA in-migration important.6 I define this measure as the percentage of people in an MSA who have in-migrated to the MSA in the past year.
Third, the Great Recession followed the burst of a housing bubble in the United States, which reversed a seven-decade trend of increasing homeownership rates. Metropolitan areas experienced sizable declines in homeownership throughout the 2006–2014 period, although the extent of decline varied across metropolitan areas (Flanagan and Wilson 2013). I account for these changes by including a covariate for the homeownership rate, defined as the percentage of people who own a housing unit, either with or without a mortgage, in an MSA.
Fourth, in addition to impacting population fertility behavior directly, economic conditions might influence fertility behavior indirectly vis-à-vis changes in marriage and partnership trends, which are important proximate determinants of fertility (Bongaarts 1978; Sawhill and Venator 2015; Stevenson and Wolfers 2007). Specifically, economic downturns might alter patterns in union formation and dissolution (for a more detailed discussion, see Morgan et al. 2012). Although marriage rates declined and cohabitation rates increased throughout the years of the Great Recession, researchers mostly attribute these changes to continuations of preexisting trends (Cherlin et al. 2013; Kennedy and Fitch 2012; Morgan et al. 2012; Schneider 2017). In regard to marital union dissolution, research by Cohen (2014) suggested no association between state-level unemployment rates and increases in divorce rates.
To be thorough in the present analysis, I include a covariate in the robustness models for the percentage of people in an MSA who are never-married to adjust for changes in marital status over time. Additionally, I estimate a supplemental model in online appendix Table S2, which adjusts for MSA-level divorce rates for the population aged 15–64. Limitations in the ACS data set preclude the inclusion of cohabitation measures throughout the entire 2006–2014 period (U.S. Department of Health and Human Services 2008).
In a final set of robustness checks, I test how (1) compositional changes of the Hispanic population throughout the 2006–2014 period and (2) the differential impact of the Great Recession on women and men might influence the results.
Compositional Changes of the Hispanic Population
Declines in Hispanic TFR in the 2006–2014 period attributable to cyclical or structural economic changes might be partially explained by compositional changes of the Hispanic population throughout this period—primarily a drop in Mexico-U.S. migration throughout the 2000s and 2010s (Choi 2014; Parrado 2011; Villarreal 2014). In the present analysis, I am unable to calculate TFR values separately for U.S.-born and foreign-born Hispanics because annual disaggregated county- or metro-level population estimates for these groups are not available. Instead, I conduct a robustness check for the estimates of Hispanic fertility that adjusts for the percentage of reproductive-age women who are Mexican immigrants. The inclusion of this measure in the statistical analysis, which captures a form of recession-driven compositional change in the Hispanic population, does little to alter the results.
Differential Impact of the Great Recession on Women and Men.
The timing and magnitude of job loss during the Great Recession varied substantially for men and women (Hartmann et al. 2010). Male unemployment increased more rapidly and remained at higher levels than female unemployment between 2007 and 2009 (Sahin et al. 2010). Female labor force participation rates were initially unaffected by the recession, whereas male labor force participation rates experienced immediate declines in early 2008 (Cunningham 2018). These differential trends are explained by a larger concentration of men employed in goods-producing industries, which were more vulnerable to recessionary impacts (Cunningham 2018; Hout and Cumberworth 2012; Wood 2014). Although women’s earnings growth slightly outpaced men’s earnings growth into the post-recession period (Cunningham 2018), the recovery period (2010–2011) was more financially difficult for women, who began to experience job loss at higher rates than men. In fact, gains during the recovery period in male employment displaced existing female employment in certain service-providing industries (Kochhar 2011). To account for these gender-specific trends throughout the years of the Great Recession, which might impact fertility behavior, I substitute the overall unemployment rate for gender-specific (1) unemployment rates and (2) labor force participation rates in sensitivity analyses.
Table 3 presents the means and standard deviations of all variables over the entire period of analysis, 2006–2014. Figure 3 displays the average MSA-level TFR values for each racial/ethnic group by year. For all racial/ethnic groups, the TFR began to decline in 2008, although at different rates. Hispanic women experienced the largest decline in TFR, from approximately 2.9 in 2006 to 2.2 in 2014—a 24 % drop. All other groups experienced declines in TFR between 2006 and 2014: by 7.8 % for Asians, 7.7 % for whites, and 7.0 % for blacks.
Figure 4 displays changes in unemployment and goods-producing businesses for all 381 MSAs between 1991 and 2014. In panel a, the average MSA unemployment rate increased sharply at the start of the Great Recession in 2008 and then slowly declined after 2011. In contrast to this cyclical trend, panel b displays a sharp decline in the average share of goods-producing businesses in MSAs at the start of the Great Recession. The rate of decline decreased and leveled off toward the end of the period in 2014. Online appendix Fig. S3 disaggregates the average trend for the share of goods-producing businesses and displays the 2006–2014 drop for each MSA.
Table 4 displays the results of the separate two-way fixed-effects regressions predicting race/ethnicity–specific TFR for all 381 OMB-designated MSAs in the United States between 2006 and 2014 (online appendix Table S1 displays the full results). All four models include time fixed effects and MSA-level covariates. In 2014, the population residing in these 381 MSAs represented approximately 85 % of the U.S. population.
In Model 1, I estimate an equation that aims to reproduce previous findings in family and fertility research that demonstrates a negative association between unemployment and fertility. The results indicate a significant, negative association between the unemployment rate and TFR for non-Hispanic white women (β = –0.007; SE = 0.001), non-Hispanic black women (β = –0.012; SE = .005), non-Hispanic Asian women (β = –0.016; SE = 0.007), and Hispanic women (β = –0.036; SE = 0.005).
In Model 2, I estimate the effect of labor market polarization, measured as the percentage of goods-producing businesses in a metropolitan area, on TFR without the inclusion of the unemployment rate parameter. In these models, the coefficients are positively and significantly associated with TFR for white, black, Asian, and Hispanic women. A 1 percentage point increase in goods-producing businesses in an MSA is associated with TFR increases of 0.021 for white women (SE = 0.002), 0.022 for black women (SE = 0.008), 0.027 for Asian women (SE = 0.010), and 0.058 for Hispanic women (SE = 0.008). A comparison of coefficients across racial/ethnic groups (Eq. (3)) indicates that the effect size for the goods-producing businesses parameters is significantly larger for Hispanics than for whites, blacks, and Asians. There is no significant difference between the coefficients for the latter three groups.
The percentage of goods-producing businesses in all MSAs declined an average of 2.42 percentage points between 2006 and 2014 (Fig. 4, panel b). The estimates in Model 2 suggest that a decline of this magnitude predicts an average decline in TFR of 0.051 for white women, 0.053 for black women, 0.066 for Asian women, and 0.141 for Hispanic women throughout the entire period. To contextualize these findings, the overall decline in TFR between 2006 and 2014 was 0.136 for white women, 0.139 for black women, 0.143 for Asian women, and 0.547 for Hispanic women. Altogether, this indicates that declines in the percentage of goods-producing businesses, on average, account for 37 % of overall TFR declines for white women, 38 % of overall TFR declines for black women, 46 % of overall TFR declines for Asian women, and 26 % of overall TFR declines for Hispanic women. These calculations illustrate the average relative differential impact of metropolitan-area labor market polarization on fertility rates across population subgroups.
Because parameter estimates of the percentage of goods-producing businesses might be explained by changes in the unemployment rate, Model 3 includes both variables. In this set of models, the effect size of the percentage of goods-producing businesses coefficients remains similar to those estimated in Model 2 for all racial/ethnic groups, although their magnitude slightly diminishes. In contrast, the coefficients for the unemployment rate attenuate in both size and significance in comparison with the coefficients estimated in Model 1. For Hispanic women, the coefficient for the unemployment rate remains sizable and precisely estimated. Overall, however, these results provide evidence that structural changes in labor markets (i.e., the loss of goods-producing businesses) are more predictive of fertility decline than cyclical changes (i.e., unemployment), at least when examining fertility across racial/ethnic groups at the metropolitan level.
To examine whether labor market polarization and unemployment continued to influence fertility behavior in the recovery period, Model 4 tests whether these two variables maintained their effect sizes during the post-recession period. To estimate the post-recession parameters, I interact a post-recession dummy variable (2011–2014) with all independent and control variables. I then calculate the pre-recession/recession parameters (2006–2010) in a comparable manner. Both columns of Model 4 estimate the same equation: the only difference is whether the interacted dummy indicator is set as the pre-recession/recession period or the post-recession period.
The results from Model 4 indicate that the effect size of the percentage of goods-producing businesses coefficients remain similar and statistically significant during the post-recession period as for the pre-recession/recession period for white, Asian, and Hispanic women. Although the coefficient for the percentage of goods-producing businesses does not change across periods for white women, it slightly increases for Asian women and slightly decreases for Hispanic women. In regard to the unemployment rate, the coefficients remain nonsignificant for black and Asian women and close to 0 for white women. However, for Hispanic women, the unemployment rate coefficient slightly diminishes between the 2006–2010 period (β = –0.033) and the 2011–2014 period (β = –0.026). To summarize, the results from Model 4 demonstrate that reductions in the percentage of goods-producing businesses in an MSA contributed to declines in TFR not only throughout the recession but also during the post-recession recovery period.
Because the specifications presented earlier do not account for several unmeasured, time-varying metropolitan-area characteristics that might influence the results, I conduct a series of robustness checks to test alternate specifications of these models.
Table 5 displays a set of models that incorporate a set of theoretically relevant MSA-level covariates, including educational attainment, past-year in-migration, homeownership rates, and marital status. The number of MSAs included in this analysis for each racial/ethnic group is reduced because of ACS one-year microdata availability. When these additional covariates are added, the coefficients and standard errors for the percentage of goods-producing businesses and the unemployment rate remain approximately comparable with those in the previous full 381 MSA analysis for all models estimated earlier, with two important exceptions. First, the goods-producing businesses coefficients diminish in size and significance for black and Asian women in all models. Second, the unemployment rate coefficient for white women diminishes in size and significance.
In online appendix Table S2, I substitute the never-married rate in an MSA for the divorce rate in that MSA among the population aged 15–64. The results from these ACS models are likewise comparable with the previous ACS subset models.
To account for changes in the composition of the Hispanic population, I reestimate the ACS subset robustness model for Hispanics by swapping past-year in-migration for a measure of Mexican composition: the percentage of reproductive-age Hispanic women who are Mexican immigrants (online appendix Table S3). The inclusion of this measure in the statistical analysis, which adjusts for compositional changes in the Hispanic population, does little to alter the main results for the goods-producing and unemployment parameter estimates.
Online appendix Tables S4a and S4b display the results of a further series of robustness and sensitivity checks. I reestimate the full model, Model 3, using (1) labor market polarization, measured as the logged number of goods-producing businesses; (2) MSA-level race/ethnicity–specific covariates in place of MSA-level covariates; (3) labor market polarization, measured as the percentage of workers employed in goods-producing industries in mid-March; (4) sex-specific unemployment rates in place of the overall unemployment rate; (5) employment-to-population ratios (overall and sex-specific) in place of the overall unemployment rate; and (6) labor force participation rates (overall and sex-specific) in place of the overall unemployment rate.
The results from these alternate specifications are comparable to those in Table 4. When operationalized as the logged number of goods-producing businesses (Models 1–2, online appendix Table S4a), labor market polarization similarly explains more variation than the unemployment rate for three of the four racial/ethnic groups. Substituting race/ethnicity–specific covariates for MSA-specific covariates (Model 4, online appendix Table S4a) does not substantively change the results of the ACS subset analysis in Table 5 (Model 3). Moreover, accounting for gender-specific trends in unemployment and labor force participation, the results from these models suggest little change from the main results. In total, this battery of alternate specifications yields comparable results to the previously estimated models in Tables 4 and 5, suggesting that the findings are largely robust.
Long-Term Structural Change, 1991–2014
The prior results indicate that the annual loss of goods-producing business establishments in metropolitan areas throughout the decade spanning the Great Recession was associated with declines in TFR for white, black, Asian, and Hispanic women in the full 381 MSA analysis and for white and Hispanic women in the ACS subset analysis. However, these models do not account for the longer span of time since the 1980s, when U.S. labor markets initially began to transition away from the production of goods and toward the provision of services. To evaluate whether these effects have been present over the past several decades, I extend the annual data set back to the early 1990s to estimate a set of two-way fixed-effects models that cover the years 1991–2014.8 At the beginning of the period, 1991, goods-producing businesses accounted for an average of 18.28 % of all businesses in MSAs; by the end of the period, 2014, goods-producing businesses represented an average of 14.24 % of all businesses in MSAs—an overall 4.04 percentage point decrease (Fig. 1). Throughout this 24-year period, the average unemployment rate was 6.2 %, although cyclical economic expansions and contractions resulted in an average low of 4.2 % in 2001 and an average high of 9.6 % in 2011. Online appendix Table S5 displays the means and standard deviations of variables in the 1991–2014 analysis.
In Table 6, I present the results of these extended period models. The results are substantively comparable with the results from the 2006–2014 analysis for the percentage goods-producing coefficients (Table 4), although the effect size is largest for black women. In Model 3, a 1 point increase in the percentage of goods-producing businesses in an MSA is associated with a 0.020 increase in TFR for white women (SE = 0.001), a 0.051 increase in TFR for black women (SE = 0.004), a 0.023 increase for Asian women (SE = 0.005), and a 0.035 increase for Hispanic women (SE = 0.005). Because the percentage of goods-producing businesses in metropolitan areas declined by an average of 4.04 percentage points over this 24-year period, the estimates from Model 3 predict that changes in industry composition accounted for an average decline in TFR of 0.081 for white women, 0.206 for black women, 0.093 for Asian women, and 0.141 for Hispanic women. For MSA-level unemployment, the coefficients suggest that the relationship between the unemployment rate and racial/ethnic TFR remained procyclical throughout the entire period for white and Asian women, but not for black and Hispanic women.
Overall, the empirical results from this extended period analysis bolster the theoretical argument that ongoing structural trends in the labor market have contributed to fertility decline over the past several decades.
Discussion and Conclusion
In the years since the Great Recession, social scientists have anticipated that economic recovery in the United States, characterized by gains in employment and median household income, would augur a reversal of declining fertility trends. However, the expected post-recession rebound in fertility rates has yet to materialize. In fact, fertility rates continue to decline, with 2018 reaching an unprecedented national low for general fertility rates (Hamilton et al. 2019). The analysis presented here demonstrates how structural changes in U.S. labor markets help explain trends in declining fertility rates. Overlooked in past research on fertility behavior, labor market polarization signifies a permanent change in the financial outlook of American workers, especially for those without a college degree. As businesses, and therefore jobs, in goods-producing industries disappear, displaced workers who wish to remain in the labor force must find employment in the lower-skill service sector, which provides jobs with fewer hours, lower pay, and scarcer benefits (Bureau of Labor Statistics and U.S. Department of Labor 2016; Janoski et al. 2014; Kalleberg 2009). The “hollowing out” of the middle of the occupational income distribution results in decreased financial security for American workers as they seek to start or expand families. Importantly, after goods-producing businesses disappear from American cities—either as a result of offshoring or improvements in assembly line automation—they tend to not return (Autor and Dorn 2013).
The present analysis suggests that labor market polarization contributed to the decline in fertility rates throughout the entire decade spanning the Great Recession, including the post-recession period between 2011 and 2014. These findings hold even in models adjusting for the unemployment rate and additional economic and noneconomic covariates. As long as goods-producing businesses remain a smaller share of the overall industry composition of U.S. metropolitan areas, the models presented here indicate that fertility rates will remain below their pre-recession levels.
Although labor market polarization is found to influence TFR for all racial/ethnic groups throughout the extended period analysis, the effect size is substantially larger in magnitude for women of color—particularly black and Hispanic women—than for white women. Job loss in goods-producing industries is more devastating for nonwhite workers because racial disparities in employment exist even during economically prosperous times (Janoski et al. 2014; Moore 2010). Discrimination in hiring practices by employers has played a significant role in perpetuating these racial disparities, constructing social barriers to stable and long-term employment (e.g., Pager et al. 2009). Consequently, the search for reemployment for nonwhite workers after economic recessions has been more drawn out than for whites, entailing lengthier spells of unemployment and therefore increased periods of financial precarity (Couch and Fairlie 2010; Hout and Cumberworth 2012).
The results also suggest a more heterogenous picture regarding the procyclicality of economic cycles and fertility rates during the Great Recession. With exception of the Hispanic models, the unemployment rate parameter estimates are mostly small in size and nonsignificant once labor market polarization is included in the models. These results are robust to alternative measures of employment, including sex-specific unemployment rates, sex-specific employment-to-population ratios, and the overall employment-to-population ratio. Although this finding contrasts with earlier studies on the Great Recession, which found considerable evidence of procyclicality (e.g., Currie and Schwandt 2014; Morgan et al. 2011; Schneider 2015), this discrepancy might be explained by the extended period of the present study and the geographic level examined.
The magnitude of the coefficient estimates of labor market polarization on fertility rates is nearly three times larger for Hispanic women than for white women. Differences in occupational distributions across racial/ethnic groups might partially explain this finding because groups with a larger share of workers in goods-producing industries would be disproportionately affected by the disappearance of these businesses. In fact, 33 % of the Hispanic labor force was employed in goods-producing industries prior to the recession in 2007, whereas only 22.5 % of whites, 20 % of Asians, and 18 % of blacks were employed in goods-producing industries at that time (Fig. 2). In absolute terms, as well, there were more Hispanic workers in manufacturing, construction, and mining/extraction industries than there were black and Asian workers combined in 2007 (Bureau of Labor Statistics 2008). It is therefore likely that compositional differences account for some of the differential impact. The large share of Hispanic workers in the goods-producing sector might also explain why the unemployment rate coefficient remained sizable and significant for Hispanic fertility.
During the post-recession period, availability and usage of long-acting reversible contraceptives (LARCs) increased substantially (Branum and Jones 2015; Finer et al. 2012). At the same time, rates of unintended pregnancy declined considerably (Finer and Zolna 2016). Past research suggests that these trends contributed to declines in post-recession fertility (Fletcher and Polos 2017; Schneider and Gemmill 2016). Schneider and Gemmill (2016) found that increases in regional LARC usage between 2003 and 2014 partially accounted for state-level declines in the nonmarital fertility rate, especially for Hispanic women. To the extent that labor market restructuring changed the demand for children and the attendant demand for contraception with lower failure rates, LARC take-up is an explanation of post-recession fertility decline that is consistent with the findings presented here. However, the association between structural labor market changes and fertility dates at least to the early 1990s (Table 6), before major increases in LARC availability and usage (Branum and Jones 2015), suggesting that LARCs may be part of but not the entire story of the present findings.
An important question that follows from this analysis is whether declines in fertility rates from labor market polarization are an indicator of delayed or forgone births. One way to approach this question is to identify which age groups were more responsive to declines in goods-producing businesses. In online appendix Table S6, I estimate the effect of structural and cyclical economic changes on age-specific fertility rates, binned at five-year age intervals, between 2006 and 2014. For all racial/ethnic groups, the largest effect sizes for the goods-producing businesses coefficients tend to be for women in their 20s. Given that the effect size is small and mostly nonsignificant for women in their 30s and 40s, the results suggest that births are likely being delayed rather than entirely forgone. At the same time, the cohort of women in their 20s at the beginning of the 2006–2014 period might have different birth histories than the cohort of women who began the period in their 30s. Future research will need to determine over the next decade whether increases in fertility rates for women in their 30s are large enough to offset decreases in fertility rates for women who were in their 20s over the past decade.
The age-specific fertility rate models also suggest that fertility rates for teenage women ages 15–19 were significantly responsive to changes in structural economic conditions for both white and Hispanic women and to changes in cyclical economic conditions for white women. There was no significant effect of structural or cyclical economic conditions on age-specific fertility rates for black or Asian women. Most research examining the influence of the Great Recession on teenage fertility rates has found little empirical evidence of a significant relationship with unemployment rates (Boonstra 2014; Lindberg et al. 2016; Percheski and Kimbro 2014).
The findings presented here should be interpreted with an understanding of several limitations of the data and analytic method used. First, the present analysis uses aggregate data and is susceptible to ecologically fallacious inference; the findings should be interpreted only in terms of population-level behavior patterns, not individual-level choices. Accordingly, I cannot directly test how individual-level financial uncertainty influences fertility decision-making. Future research should examine how fertility preferences, intentions, and demand are influenced by both individual-level labor market experiences and aggregate labor market conditions. Second, the TFR measure relies on both administrative data and population estimates. Although birth counts as the numerators of the equation represent a full enumeration of all new births in U.S. MSAs between 1991 and 2014, the population counts as denominators of the equation draw on bridged-race population estimates. However, the bridged-race population estimates used in the present analysis are the same source of population estimates used by the NCHS when generating and distributing public-release vital statistics figures, including fertility rates, and are regarded as the best source of population data for this task.
Despite these limitations, this study offers an innovative economic explanation for why fertility rates in the United States continue to decline. Prevailing theoretical perspectives on how economic conditions influence fertility have primarily relied on cyclical measures of economic change. Researchers have overlooked important long-term structural changes in U.S. labor markets, particularly the decline of manufacturing and other goods-producing industries, and the coincident rise of low-paid jobs in service industries. Whereas the present analysis focuses on the impact of economic restructuring in U.S. metropolitan areas on fertility rates, future research should account for whether this relationship is similar in nonmetropolitan and rural areas, which have also experienced considerable declines in goods-producing industries (Low 2017). By proposing a structural economic explanation for why fertility rates continues to decline, this study contributes to the understanding of how macroeconomic conditions influence fertility behavior and decision-making.
The author is grateful for feedback from Jenna Nobles, Myra Marx Ferree, Marcy Carlson, Christine Schwartz, and Jennifer Laird. This research was supported by a core grant to the Center for Demography and Ecology at University of Wisconsin–Madison (P2C HD047873) as well as support from a training grant awarded to the Center for Demography and Ecology (T32 HD007014). All errors are the author’s own.
Alternatively, displaced workers often exit the labor force entirely, enroll in workforce retraining programs, or pursue additional educational degrees and certifications (Janoski et al. 2014; McConnell et al. 2016).
I additionally estimate models using a data set that extends from 1991 to 2014, which relies on 82.3 million birth certificate records.
To account for comparability issues between SIC and NAICS industry classification codes in the extended period analysis (1991–2014), I construct an analogous domain-level grouping of SIC codes for CBP data between 1991 and 1997.
Because PUMAs are nested within states rather than within counties, MSA-level estimates have errors of commission and omission in which areas outside the MSA are included or areas within the MSA are not included, respectively. Most MSAs have a combined error of less than 0.1 % or less than 4.9 %, but 44 MSAs have an error of 5.0 % to 9.9 %, and 40 have an error of 10 % to 14.9 %. In other words, 296 MSAs have less than a 5 % geographic boundary error. In this analysis, I make the plausible assumptions that (1) the outlying geographic areas of MSAs are not excessively biasing the overall MSA population averages, and (2) subpopulation averages in outlying areas of MSAs are similar whether they happen to be immediately inside or outside the MSA boundary.
I conceptualize these fixed characteristics as aspects of geography, climate, city-specific cultural norms and mores, shared history, and place-specific socioeconomic and class distributions. To be sure, these metropolitan-area characteristics do change over time; but given the relatively brief period of analysis, I make the plausible assumption that metropolitan areas maintain a fixed set of social, cultural, and built-environment characteristics.
Kothari et al. (2013) noted that geographic mobility declined throughout the years of the Great Recession. For the geographic mobility that did occur, labor migration during the Great Recession varied for low- and high-skilled workers as well as across foreign-born and non-foreign-born workers (Cadena and Kovak 2016). In-migration from Mexico, for instance, decreased as a result of economic disruption to the construction and manufacturing sectors in the United States (Calnan and Painter 2017; Villarreal 2014).
Because the labor market measures and covariates are lagged by one year to approximate economic conditions at the time of conception, the prerecession/recession period (2006–2010) aligns with economic conditions in 2005–2009, and the post-recession period (2011–2014) aligns with economic conditions in 2010–2013.
County-level CBP data on industry composition are available from 1986 onward; however, county-level LAUS data are available from 1990 onward.
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