Results by Birth Cohorts, by Type of Pension, and by Residence105

weighting the different birth cohorts—is a restriction of the analysis to specific birth cohorts in the first place. In Figure 4.3, I apply this stratification and partition the data in two cohorts born between 1921 and 1930 and 1931–1940, respectively. Con-sidering that deaths occurred between 1994 and 2005, these cohorts are chosen be-cause their realized age at death of the older cohort is by construction relatively close to the age at death which can be expected, conditional on reaching retirement age. The sample is not weighted, but restricted to males with at least 25 years of contribution, and theU–shaped result survives the stratification. The difference in the level of longevity between the two cohorts is an artifact of the construction of

CHAPTER4 Non–Monotonicity in the Income–Longevity Relationship the cohorts and is due to the absence of weighting: In the younger cohort, early deaths are relatively over–sampled as compared to the older cohort.

Left panel: By birth cohort (unweighted and restricted; strategy [I]); solid: yr. of birth∈ (1920,1930], dashed:

yr. of birth∈(1930,1940]. Right panel: By type of pension (weighted and restricted; strategy [I]); solid: old–age pension, dashed: disability pension. (Data set: unweighted and restricted, weighted and restricted)

By residence (unweighted and restricted; strategy [I]); solid: west, dashed: east, dotted: abroad. (Data set: un-weighted and restricted, un-weighted and restricted)

Figure 4.3: Results by Cohorts, Type of Pension, and Residence

The results stratified along the type of pension (see also Figure 4.3) does not come as a surprise: men who die receiving an old–age pension live longer than men who die receiving a disability pension. The explanation is two–fold; first, individuals with a disability pension suffer from a shorter life expectancy as a result of their respective illness, an illness so severe to prohibit a later retirement into the regular old–age pension.

Second, an institutional factor shapes this result, because once eligible, disabil-ity pensions are transformed into old–age pensions, such that an individual receiv-ing a disability pension first, but reachreceiv-ing the age of 65 will be characterized by the outcome ’old–age pension’.¹¹ The sign of the respective coefficient in the least

11Under certain circumstances, a pensioner receiving a pension due to a reduction of his earnings capacity can already be transformed into an old–age pension at the age of 60 (only after an application of the pensioner).

CHAPTER4 Non–Monotonicity in the Income–Longevity Relationship squares regressions corroborates that old–age pensioners benefit from higher life expectancy. Altogether, theU-shape emerges for both groups.

Although the majority of observations live in West Germany, I provide estima-tion results for East Germany and abroad as well (Figure 4.3). All groups exert theU-shape, however, to a different degree. The dip is strongest for East Germany, and least pronounced for those living abroad. Further differences between the three residence groups are not conclusive; the non–parametric estimates do not disclose a clear order between all groups, while the signs of the least squares regressions for pensioners living abroad depend on the respective specifications (though the coefficients are always significant).

Interestingly, I find in the parametric estimates that people in East Germany live slightly longer, despite the fact that living standards in East Germany are ce-teris paribus lower than in West Germany.¹² Yet, this finding is compatible e.g. with von Gaudecker and Scholz (2007), who find that for some income groups, life ex-pectancy is higher in East Germany. Furthermore, official records (see Statistisches Bundesamt 2007, p. 54) show that in 2005, remaining life expectancy conditional on reaching 60 was only slightly higher in the West (20.4 years compared to 19.7 years in the East), and there are birth cohorts whose unconditional life expectancy at birth in the year was even higher in the East.¹³ In contrast to population statis-tics, the population I observe consists of pensioners of the public pension system and is conditioned on 25 years of work, and since different legislation applied for the collection of benefit claims in both parts of the country (until reunification), the observed sub–populations in East and West may differ in their relationship to the particular total populations. Although benefit claims for men are less dispersed in East Germany (15.03 in the West, compared to 11.75 in the East), the income gra-dient as graphed in Figure 4.3 is stronger: there are less income differences, but if there are differences, their impact on life expectancy is stronger in the East.

4.4.4 Results by Months in Ill–Health

I apply three distinct strategies to identify the impact of health status on the re-lationship between life–expectancy and benefit claims (see Figure 4.4). First, ap-plying strategy (I), the uni–variate analysis is stratified along the ill–health dimen-sion, hence, I estimate the relationship three times, contingent on the outcome of the stratification variable—namely, months spent in ill–health being equal to zero;

12Average income for men in the years 2002 to 2006 in the East reached only 71.0% to 75.2% of the income in West Germany, see Statistisches Bundesamt (2007, p. 523).

13E.g. the birth cohort of 1951 and 1952, covering pensioners who also appear in the data I analyze here. The life expectancy at birth in the Eastern part of Germany was 65.1 years compared to 64.6 years in the West.

CHAPTER4 Non–Monotonicity in the Income–Longevity Relationship

Left panel: Uni–variate (strategy [I]); solid: mts.= 0, dashed: mts.< 6, dotted: mts.≥6. Right panel: Multi–

variate (strategy [II]); solid: mts.= 0, dashed: mts.<6, dotted: mts.≥6. (Data set: weighted and restricted)

Figure 4.4: Results by Months in Ill–Health

smaller than six, but strictly positive; and greater or equal than six. The same groups are constructed for strategy (II).¹⁴Denote benefit claims byx_1i,x_2idenotes months in ill–health,x_3i denotes months in unemployment, andx_4i denotes years of contribution, and finally,xjiis the respective sample average over allxji. Hence, the three plots in the right panel of Figure 4.4 represent the conditional moments

m₁(x_1i | x_2i = 0, x_3i=x_3i, x_4i =x_4i)

m2(x1i | x2i ∈(0,6), x3i =x3i, x4i=x4i) (4.12) m₃(x_1i | x_2i ≥6, x_3i=x_3i, x_4i =x_4i),

The uni–variate procedure (I) produces little to no difference in the resulting relationships, and the downward–sloping area for low benefit claims survives, de-spite the stratification.

The ordering of conditional moments extracted from the array of results (strat-egy [II]) for three different groups is unique, with individuals with zero months in ill–health at the bottom, and the group with the longest spell in ill–health at the top.

This is corroborated by strategy (III), which yields either insignificant results for the influence of ill–health (WLS estimation on the complete data set) or significantly positive impact (all other specifications), meaning that individuals with rehabilita-tion spells or months in ill–health actually live longer. This result is explicitly not an artifact of individuals in bad health being more likely to claim disability pensions

14The three different groups in specification (II) match as closely as possible the groups of specifica-tion (I); a perfect match, however, is not possible, for the following reason: Applying strategy (I), the sub–groups are extractedbeforethe estimation, while in strategy (II), the sub–groups are constructed afterwards, based on a grid of 25 points over eachxji. Specifying the sub–groups on the grid after the estimation may slightly shift the cut–off limits.

CHAPTER4 Non–Monotonicity in the Income–Longevity Relationship than people in good health, which is not the case (the average number of months spent in ill–health is 3.2 for individuals with an old–age pension, and only 2.6 for individuals with a disability pension). Potential explanations for this seemingly counter–intuitive result are the following: first, means of rehabilitation are actually effective and extend life–expectancy. Second, disability pensions and times in ill–

health are substitutes, implemented for the same general reason (ill–health), but at different levels of the outcome. While months in ill–health are by definition a tem-porary means to improve the situation of ill individuals, the disability pension is more likely to be ultimate and applied for more severe degrees.¹⁵

The latter explanation is compatible with results stratified by the type of sion (Figure 4.3 and Section 4.4.3), because individuals receiving a disability pen-sion live shorter as compared to individuals with old–age penpen-sions. The former explanation is also compatible with results produced of Chapter 3, where I find that—on average—bad health increases the duration of the benefit spell (while the benefit claims–gradient is stronger for worse health). However, including health as explanatory variable in strategy (III) does not alter theU–shape, whereas it van-ishes in strategy (II).