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Life Table Suit for each x, and the rest of the computations follow Life Table suit. From any given schedule of death probabili- ties q0,q1,q2,..., the lx table is easily computed A life table is a tabular representation of central sequentially by the relation lx 1 lx(1 qx ) for x features of the distribution of a positive random 0, 1, 2,.... Much of the effort+ in= life-table− construc-= variable, say T , with an absolutely continuous tion therefore is concentrated on providing such a distribution. It may represent the lifetime of an schedule qx . individual, the failure time of a physical component, More conventional{ } methods of life-table construc- the remission time of an illness, or some other tion use grouped survival times. Suppose for sim- duration variable. In general, T is the time of plicity that the range of the lifetime T is subdivided occurrence of some event that ends individual into intervals of unit length and that the number of survival in a given status. Let its cumulative failures observed during interval x is Dx. Let the distribution function (cdf) be F (t) Pr(T t) corresponding total person-time recorded under risk = ≤ and let the corresponding survival function be of failure in the same interval be Rx . Then, if µ(t) S(t) 1 F (t), where F(0) 0. If F( ) has the is constant over interval x (the assumption of piece- = − = · probability density function (pdf) f(), then the risk wise constancy), then the death rate µ D /R is · x x x of event occurrence is measured by the hazard the maximum likelihood estimator ofˆ this= constant. µ(t) f (t)/S(t), for t where S(t) > 0. Because Relation (1) can again be used to provide an estimator of its= sensitivity to changes over time and to risk qx 1 exp( µx ), (2) differentials between population subgroups, µ(t) is a ˆ = − −ˆ centerpiece of interest in empirical investigations. and the crucial first step in the life-table computation In applications to human mortality, which is where has been achieved. Instead of (2), µ 1 1 µ is life tables originated, the time variable normally is a x 2 x often used to estimate q . This solutionˆ is+ ofˆ older person’s attained age and is denoted x. The function x vintage and may be regarded as an%& approximation' µ(x) is then called the force of mortality or death to (2). intensity (see Hazard Rate). The life-table function Two kinds of problems may arise: (i) the exact l 100 000 S(x) is called the decrement function x value of R may not be known, and (ii) the constancy and= is tabulated for integer x in complete life tables; x assumption for the hazard may be violated. in abridged life tables it is tabulated for sparser values When the exact risk time R is not known, some of x, most often for five-year intervals of age. The x approximation is often used. An Anglo-Saxon tra- radix l is selected to minimize the need for decimals 0 dition is to use the midyear population in the age in the l table; a value different from 100 000 is x interval. Alternatively, suppose that the number N sometimes chosen. Other life-table functions are the x of survivors to exact age x and the number W of expected number of deaths d l l at age x x x x 1 withdrawals (losses to follow-up) in the age interval x (i.e. between age x and age =x 1),− the+ single- + are known. What has become known as the actu- year death probability qx Pr(T x 1 T > x) = ≤ + | = arial method then consists in approximating Rx by dx /lx, and the corresponding survival probability 1 Nx 2 (Dx Wx ). If there are no withdrawals and px 1 qx . Simple integration gives − + = − Nx is known, then Dx/Nx is the maximum likelihood x 1 + estimator of qx, and this provides a suitable starting qx 1 exp µ(s) ds .(1) point for the life-table computations. = − − ! "x # For the case where only grouped data are avail- Life-table construction consists in the estima- able and the piecewise-constancy assumption for tion and tabulation of functions of this nature from the intensity function is implausible, various meth- empirical data. If ungrouped individual-level data are ods have been developed to improve on (2). For available, then the Kaplan–Meier estimator can be an overview, see Keyfitz [12]. Even if single-year used to estimate lx for all relevant x and estimators age groups are used, mortality drops too fast in the of the other life-table functions can then be com- first year of life to merit an assumption of con- puted subsequently. Alternatively, a segment of the stancy over this interval. Demographers often use Nelson–Aalen estimator can be used to estimate µ0/[1 (1 a0)µ0] to estimate q0, where a0 is some x 1 ˆ + − ˆ + x µ(s) ds; (1) can then be used to estimate qx small figure, say between 0.1 and 0.15 [2]. If it is $ 2 Life Table possible to partition the first year of life into subin- then individuals who live at widely differing ages in tervals in each of which mortality can be taken as the period of observation cannot be expected to have constant, then it is statistically more efficient essen- the same risk structure, and the period table is said tially to build up a life table for this year. This leads to reflect the patterns of a synthetic (fictitious) cohort to an estimate like q0 1 exp( i µi ), where the exposed to the risk of the period at the various ages. sum is taken over theˆ = first-year− intervals.− ˆ See Dublin et al. [5, p. 24] for an example. ( The force of mortality is sometimes represented Multiple-decrement Tables by a function h(x; θ), where θ is a vector of When two or more mutually exclusive risks operate parameters. Actuaries most often use the classical on the study population (see Competing Risks), one Gompertz–Makeham function h(x; a, b, c) a x = + may correspondingly compute a multiple-decrement bc for the force of mortality in their life tables table to reflect this. For instance, a period of sick- (see Parametric Models in Survival Analysis). ness can end in death or, alternatively, in recovery. When individual-level data are available, it would be Suppose that an integer random variable K repre- statistically most efficient to estimate the parameters sents the cause of decrement and define Fk(t) by the maximum likelihood method, but most often = Pr(T t, K k), fk(t) dFk(t)/ dt, and µk(t) they are estimated by fitting h( ; θ) to a schedule ≤ = = = · fk(t)/S(t), assuming that all Fk( ) are absolutely of death rates µx , perhaps by least squares, · continuous. Then µk( ) is the cause-specific hazard minimum chi-square{ˆ } (see Ban Estimates), or some · (intensity) for risk cause k and µ(t) k µk(t) is method of moments. This approach is called analytic the total risk of decrement at time t. For= the multiple- graduation; for its statistical theory, see [11]. One of decrement table, we define the decrement( probability many alternatives to modeling the force of mortality (k) is to let [10] qx Pr(T x 1,K k T > x) = ≤ + = | 1 t qx (x B)C 2 x A + D exp[ E(ln x ln F) ] GH . exp µ(x s)ds µk(x t)dt. p = + − − + = − + + x "0 ! "0 # (3) So far we have tacitly assumed that the data (k) come from a group of independent individuals who For given risk intensities, qx can be computed by have all been observed in parallel and whose life- numerical integration in (3). The expected number times have the same cdf. Staggered (delayed) entries of decrements at age x as a result of cause k is (k) (k) into the study population and voluntary exits (with- dx lx qx . When estimates are available for the drawals) from it are permitted provided they contain cause-specific= risk intensities, one or two columns no information about the risk in question, be it death, can therefore be added to the life table for each cause (k) (k) recurrence of a disease, or something else. Never- to include estimates of dx and possibly qx . theless, the basic idea is that of a connected cohort Several further life-table functions can be defined of individuals that is followed from some significant by formal reduction or elimination of one or more of starting point (like birth or the onset of some disease) the intensity functions in formulas like those above. and which is diminished over time due to decrements In this manner, a single-decrement life table can be (attrition) caused by the risk’s operation. In demog- computed for each cause k, depicting what the normal raphy, this corresponds to following a birth cohort life table would look like if cause k were the only through life or a marriage cohort while their mar- one that operated in the study population and if it riages last, and the ensuing tables are called cohort did so with the risk function estimated from the data. life tables. The purpose is to see the effect of the risk cause Because such tables can only be terminated at the in question without interference from other causes. end of a cohort’s life, it is more common to compute Some demographers call this abstraction the risk’s age-specific attrition rates µx from data collected for pure effect.
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