The 2016 election highlighted deep divisions in the USA, and exposed unhappiness and frustration among poor and uneducated whites. The starkest marker of this unhappiness is the rise in preventable deaths and suicides among the middle aged of this cohort. In contrast, minorities have much higher levels of optimism, and their life expectancies continue to rise. Low-income respondents display the largest differences, with poor blacks by far the most optimistic, and poor rural whites the least. African Americans and Hispanics also have higher life satisfaction and lower stress incidence than poor whites. The gaps across racial groups peak in middle age, at the nadir of the U-curve of age and life satisfaction. We explored the association between our subjective well-being data and the Centers for Disease Control and Prevention (CDC) mortality data. We find that the absence of hope among less than college-educated whites matches the trends in premature mortality among 35–64-year-olds. Reported pain, reliance on disability insurance, low labor force participation, and differential levels of resilience across races all have mediating effects in the desperation-mortality associations. We also explore the role of place, and map the states associated with higher/lower indicators of well-being for these different cohorts. The matches between indicators of well-being and of mortality suggest that the former could serve as warning indicators of ill-being in the future, rather than waiting for rising mortality to sound the alarms.
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For example, the high material costs of being poor in Latin America in the 1970s, which included paying as much as 18 times more per unit of water and electricity, with inferior health outcomes (Adrianzen and Graham 1973).
Gelman and Auerbach (2016) posit that these trends are driven in part by aggregation bias at the older ages of the 45–54 cohort.
Obtained through NBER at http://www.nber.org/data/seer_u.s._county_population_data.html.
This proportion has increased very slightly over time: from 85 to 85.5% from 2010 to 2014.
GH halved the number of daily interviews to 500 in 2013, decreasing the number of weighted MSAs.
In Appendix 2, we employ alternative definitions for poor, middle-income, and rich individuals, including that from the Census Bureau. The results we obtain are quantitatively similar to those in our main specification.
This percentage and the previous one do not use Gallup’s survey weights. The corresponding shares of weighted respondents are 15 and 24%.
The age dummies have ranges that contain a similar number of observations and generally match the age brackets present in the other databases that we used, such as the CDC’s Compressed Mortality File.
The inclusion of (MSA)*(Year) interactions, month of interview dummies, other types of survey weights, and no survey weights at all, does not meaningfully change in the coefficients. See Appendix 2.
While the logit and ordered logit models are technically the appropriate specification, it has become accepted practice to use OLS in happiness regressions, for ease of interpretation, as long as the results are very close. In our case, the OLS specifications yield slightly more significant coefficients for some variables of interest, but the patterns are otherwise identical and the choice of model does not affect our conclusions. We report the logit and ordered logit results in Appendix 3.
However, we include both year and MSA dummies in every specification.
We also explored reported depression and found that it was also highest among low-income whites. A more comprehensive account, however, would require a separate study. Depression and happiness are distinct emotional states. While positive emotional states, such as happiness and smiling, tend to track closely, negative states—stress, anger, and depression—track differently, with depression the most distinct. See Stone and Mackie (2013).
We computed Figure 1 using the coefficients from column (2), the BPLA regression without BPL as a control. When we use current BPL as a control (column (3)), we get slightly lower gaps between poor blacks and poor whites and they decrease less as income increases.
The social support and anger questions were asked in 2008–2012 and 2010–2013, respectively. Therefore, the time period under consideration differs for those two cases.
We also collapsed the data at the MSA level, using MSA fixed effects to control for non time-varying MSA-specific unobservables. This reduces the significance of some variables, but the main ones hold.
While we find a sharp drop in the life satisfaction and optimism of Democrats and Independents in weeks following the 2016 election, a partial recovery seems to be underway by the end of the year (https://www.brookings.edu/blog/up-front/2017/02/02/the-trump-unhappiness-effect-nears-the-great-recession-for-many/). We do not yet have the data to test if there is a longer-term negative effect of trends since then—and plan to do so going forward. The evidence above, though, suggests that this is not a finding that is explained by short-term events.
Graham and Pettinato (2002) coined the term “happy peasants and frustrated achievers” to describe such optimistic poor individuals in many poor countries over a decade ago.
These use national-level, rather than MSA-level, survey weights.
National Center for Health Statistics. Compressed Mortality File, 2008–2015 (CD-ROM Series 20, No. 2 U) Vital Statistics Cooperative Program. Hyattsville, Maryland. 2016.
The three decades considered (35–44, 45–54, 55–64) all had similar “composite” mortality rates.
The calculation is: exp(−0.087 ∗ log (1.50)). The log represents a 50% increase and (log (1.50) = 0.40547); the product equals approximately − 0.04. The mean optimism or expected future life satisfaction is 7.86, so the change above corresponds to approximately 0.5% of this mean value.
Regression results available from the authors.
Social Security Advisory Board: http://www.ssab.gov/Disability-Chart-Book. These are not age-adjusted numbers.
We thank Henry Aaron for raising this.
For the distribution of broadband, see: https://www.broadbandmap.gov/technology.
In this case, minorities comprise only African Americans and Hispanics.
We omitted income variables as otherwise state dummies would disproportionately pick up the disadvantageous state-level aspects, such as higher costs of living (Oswald and Wu 2011).
We excluded states with less than 50 observations/year for the group in question.
The U.S. has the world’s highest per capita consumption of opioids: http://www.painpolicy.wisc.edu/country/profile/united-states-america.
40% of Medicare recipients are unaware of being on a government program (Kuziemko et al. 2015).
See, e.g., https://www.whatworkswellbeing.org/ .
When regressing the household size variable on income group (recall that Gallup’s income variable assigns respondents to income brackets, coded from 0 to 10), a coefficient of 0.080 is obtained. This would mean that, on average and imposing a linear progression, an increase of 1 in the income group is associated with an increase of 0.08 in the household size.
Steven Ruggles, Katie Genadek, Ronald Goeken, Josiah Grover, and Matthew Sobek. Integrated Public Use Microdata Series: Version 6.0 [dataset]. Minneapolis, MN: University of Minnesota, 2015.
Respondents whose reported household size is larger than 10 are dropped from the analysis (951 observations).
More precisely, 19 and 20% of the (unweighted) observations corresponded to the poor and to the rich groups, respectively. Upon application of the sampling weights, these percentages changed to 27 and 14%, respectively.
Under this specification, rich respondents are defined in the same way as in third alternative of Section a). The results do not meaningfully change if the rich are classified under the criterion used for the base specification (i.e., the rich group corresponds to the respondents whose reported household income is above $120,000/year; these results are not displayed but are available from the authors, upon request).
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The authors are, respectively, Leo Pasvolsky Senior Fellow at the Brookings Institution and College Park Professor, University of Maryland, and PhD student, University of Maryland. We thank Andrew Oswald and Eddie Lawlor, as well as Alice Rivlin, Alan Blinder, Belle Sawhill, Bill Galston, Mike O’Hanlon, Bradley Hardy and other participants at a Brookings “restoring the middle class” seminar, for very helpful comments. They also appreciate the suggestions of an anonymous reviewer. Graham acknowledges the generous support from a Robert Wood Johnson Foundation pioneer award, and Pinto from a flagship fellowship at UMD.
This study was funded by grant # 74378 from the Robert Wood Johnson Foundation.
Conflict of interest
The authors declare that they have no conflict of interest.
Responsible editor: Erdal Tekin
Appendix 1 Race-income heterogeneities by year
Appendix 2 Robustness checks
Household size adjustments
One potential concern regarding the results in Table 2 is that we assign respondents to income groups based on total household income. In the GH data, household size correlates positively to household income,Footnote 36 which would introduce a bias in the estimated coefficients for the income and interaction terms. In our baseline specification, we did not adjust for household size, for two main reasons. One is the high share of missing values for this variable (25% of the observations in our baseline specifications). A second reason is that, as mentioned in Section 3, the income variable in GH is not continuous and instead assigns respondents to 1 of 11 income brackets. Adjusting the reported household income to the household size would therefore require assigning respondents an income value, based on the bracket they report. This problem is further compounded by the fact that, with a categorical income variable, incomes are inevitably top-coded, which demands further assumptions regarding how to assign income to the households in the top bracket.
We address this concern with three different strategies. In the first case, we consider only the cases of one-person households, where no adjustment is necessary. In the second alternative, we exclude those in the top income bracket (i.e., respondents reporting pre-tax household income above $120,000/year), assign every other respondent the midpoint of the income bracket they reported, and adjust reported income by household size, on a per capita basis. In the final alternative, we do not exclude any respondent. For those not in the top income bracket, we applied the adjustment described in the previous alternative. We assigned those in the top income bracket a value based on data from the American Community Survey, obtained through IPUMS (Ruggles et al. 2015).Footnote 37
Table 8 displays the results when following the first alternative. The magnitude of the indicator variables for income groups among poor white respondents increases slightly (see rows for “Poor household”). Nevertheless, the racial heterogeneities remained very stark. For instance, among the poor and holding everything else constant, African Americans score nearly 1.1 points higher on the 0–10 optimism scale than whites and are 13 percentage points less likely to have experienced stress the previous day (see rows for “Black” and “(Poor household)*(Black)”).
As mentioned above, the second and third alternatives adjust the reported pretax income for household size. This required additional assumptions. We assigned those in income brackets below the top to the midpoint. We assigned those in the top income bracket the average of households whose total pretax income exceeds $120,000/year, based on estimates using data from Ruggles et al. (2015). For every year in the 2008–2015 period, we identified households reporting pretax income above the $120,000/year threshold, computed the corresponding average income, and assigned this yearly amount to the respondents in the top bracket. We then converted all incomes into per capita amounts, by dividing the total household income by the household size.Footnote 38 Finally, we reassigned the three income categories to reflect this new per capita income variable. We specified thresholds such that we would again obtain approximately 20% of observation in the rich group, another 20% in the poor, and the remaining in the middle-income group.Footnote 39 This resulted in a maximum threshold of $12,499 per person for the poor group and in a minimum of $54,000 per person for the rich group.
The second alternative differs from the third only in the choice to exclude those assigned to the top income bracket in the GH data, which substantially reduces the number of respondents in the rich group. Table 9 displays the estimates obtained when using this alternative. As before, the race-income heterogeneities remain quantitatively large despite a slight decrease in the optimism gap: poor African Americans are now 0.83 points higher than poor whites in the optimism scale.
Table 10 displays the estimates obtained when using the third alternative, which includes the respondents in the top income bracket. As in the previous cases, this approach generates large race-income heterogeneities between African Americans and whites.
Alternative measure of poverty following US Census Bureau
Two possible objections to the robustness checks conducted in the previous subsection are that the thresholds chosen are relatively arbitrarily and that the definition of poverty used implicitly ignores any type of equivalence scale. Regarding the latter aspect, it means that the income needed for a household to be above the poverty threshold is always linearly proportional to the household size, ignoring any aspect related to its composition or the age of its members. An alternative to address both issues, then, is to use the poverty thresholds that the US Census defines every yearFootnote 40 and correspondingly classify respondents as poor.Footnote 41 Table 11 displays the results for this specification. As before, there are no meaningful differences in the race-income heterogeneities.
Exclude MSAs with smaller numbers of poor African American respondents
Another concern about the base specification results is that the results could be driven by the within-MSA variation in MSAs with very few African Americans, particularly poor ones. Table 12 displays the results obtained when running the base specification under different thresholds for the minimum number of low-income African Americans per MSA, per year. As panel A to panel C illustrate, there are again no meaningful differences in magnitude and significance levels across the different thresholds.
Include month and (MSA) × (year) dummies
A possible objection to the specification laid out in Eq. (1) is that, by using year and MSA dummies separately (i.e., without adding their interaction), we are imposing a parallel time trend on all MSAs. If the MSAs happened to follow heterogeneous time trends during the period under analysis, the absence of interaction dummies could bias our estimates. Similarly, the time of the year of the interview might be correlated with our variables of interest. Table 13 below displays the results when we include both month and (MSA) × (year) dummies. The coefficient estimates and significance are nearly unchanged, suggesting that neither factor was in fact introducing a meaningful bias into our estimates.
Robustness to use and type of survey weights
The base specification estimates in Table 2 use MSA-level weights. A possible concern is that the results may be sensitive to the type of survey weights, or simply to their use.
Table 14 below displays the results when the national-level survey weights are used. I this case, we are no longer restricted to the 196 MSAs for which we have MSA-level survey weights at some point during the 2010–2015 period, and as a result, our sample increases and encompasses nearly all the existing MSAs. The coefficient estimates for our variables of interest, however, suffer no relevant change.
Table 15 also uses this enlarged sample of respondents located in any MSA, but instead uses no weights. The differences to Table 2 are small and the coefficient estimates are often of a higher magnitude.
Appendix 3 Ordered logit and logit estimation
As mentioned in the main text, we re-estimate the main tables of the article under ordered logit and logit specifications. Tables 16 to 21 below show that our findings are robust to the choice of estimation framework.
Appendix 4 The geography of stress and pain, by race group
Figure 5 below displays the maps for stress and pain, which are not in the main text.
Figure 6 shows the corresponding boxplots of state coefficients for each of the five mapped variables. Although the geographical patterns are different for whites and minorities, the dispersion in absolute terms is similar for both groups, except for life satisfaction, where location seems to matter substantially more for minorities.
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Graham, C., Pinto, S. Unequal hopes and lives in the USA: optimism, race, place, and premature mortality. J Popul Econ 32, 665–733 (2019). https://doi.org/10.1007/s00148-018-0687-y
- Premature mortality