# Properties of Future Lifetime Distributions and Estimation

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1 Properties of Future Lifetime Distributions and Estimation Harmanpreet Singh Kapoor and Kanchan Jain Abstract Distributional properties of continuous future lifetime of an individual aged x have been studied. The expressions for moments, coefficients of skewness and kurtosis have been derived when newborn s lifetime follows different distributions. The maximum likelihood (ML estimation of the parameters of future lifetime s distributions has been explored for whole life and term assurance contracts. Simulations have been carried out to find the ML estimates and the corresponding root mean square errors. Keywords: Future Lifetime, Whole life assurance, Term assurance, Maximum Likelihood Estimation, Censoring. 2 AMS Classification: 62P5 1. Introduction Financial and actuarial models are built around the future lifetime random variable of individuals. The distribution of future lifetime is required for statistical measurements, analysis and inference. There are two approaches for modeling this distribution. In the first approach, a theoretical statistical distribution is assumed for the future lifetime associated with a stochastic status. In the second, empirical or observed survival and mortality data are used for constructing life mortality tables [1, 3, 4, 12]. In Actuarial Science, the future lifetime of a life aged x plays a significant role. The importance of its distribution is highlighted when dealing with continuous whole life and term assurance contracts where the death benefit is paid at the moment of death. It also plays a role while studying continuous n-year temporary and n-year deferred whole life annuities. The distribution of future lifetime and its properties are utilized for finding the expected present values (Actuarial values and the variances of the present value random variables for benefits or payments in case of continuous life annuities/assurances. The force of mortality, alternatively known as failure rate/hazard rate function in reliability is defined through its density function [1, 2, 3]. Let X be the new born s age at death and f(x, F (x and S(x be the corresponding probability density function (pdf, cumulative distribution function (cdf and survival function (sf respectively. (x symbolizes an Department of Statistics, Panjab University, Chandigarh-1614 (India Department of Statistics, Panjab University, Chandigarh-1614 (India Corresponding Author Kanchan Jain

2 2 individual aged x and T (x denotes the future lifetime of (x at the inception of an insurance contract. The survival function of T (x is written as (1.1 tp x = P [T (x > t] = P (X > x + t X > x = S(x + t. S(x tp x gives the survival probability that (x survives for more than t years. For t, t q x = 1 t p x gives the distribution function of T (x and represents the probability that (x does not survive beyond age t. The force of mortality is an important and fundamental concept in modelling future lifetime and is defined as 1 f(x + t (1.2 µ x+t = lim dt + P [t < T (x t + dt T (x > t] = dt S(x + t. In reliability theory, µ x is known as failure rate or hazard rate function and T (x is termed as residual life. One can refer to [2, 5] for the notations given above. If the policies are not surrendered or do not lapse as in case of whole life assurance, then the observed future lifetimes are uncensored. The censored observations arise if the policyholders surrender the policy before the expiry of the contract. In some cases, the policy lapses due to non-payment of regular premium amount. In term assurance contract or n-year temporary policy, the death benefit is paid only if the insured dies within the term of the contract. In this case, the future lifetime of those individuals who survive the term of the policy, will not be known to the company. The same is true for an n-year temporary life annuity where annuity payments stop when an individual dies within or after n years, whichever is earlier. In literature, many authors have explored the ageing properties of T (x [11]. The distributional and ageing properties of curtate future lifetime K(x, the greatest integer less than or equal to T (x have been already studied [6]. To the best of our knowledge, the distributional properties of T (x have not been studied so far. Hence, we study the distributional forms of T (x for different distributions of X and explore the Maximum Likelihood Estimation (MLE for uncensored and censored cases. The distributions chosen for X are Gompertz, exponential, Weibull, Pareto Type I, exponentiated exponential and burr [7, 8]. Gompertz and Weibull are considered to be very good models for studying human life-length. Pareto is a particular case of Benktander Gibrat distribution [9] and Burr distribution is considered since Pareto type II is its special case. The expressions for pdfs, n th order moments, coefficients of skewness and kurtosis of different distributions of T (x are derived in Section 2. In Sections 3 and 4, we discuss the maximum likelihood (ML estimation of parameters future lifetime s distributions for uncensored and censored observations in case of whole life and n-year term assurance contracts. Section 5 consists of simulation results for ML estimation using numerical techniques. The root mean square errors (RMSEs are also reported. 2. Distributional Properties of Future Lifetime In this section, we derive the n th order moment of T (x which helps in finding its expectation, variance, coefficients of skewness and kurtosis. The probability

3 3 density function of T (x is given by (2.1 f T (x (t = d dt t p x = t p x µ x+t. The cdf of T (x is t q x = t s p x µ x+s ds [2]. If we assume that the n th order moment of T (x,that is, E[(T (x n ] exists, then E[(T (x n ] = This implies that (2.2 = E[(T (x n ] t t t t n tp x µ x+t dt y n yp x µ x+y dy + y n yp x µ x+y dy + (t n t y n yp x µ x+y dy t yp x µ x+y dy y n yp x µ x+y dy (t n yp x µ x+y dy t (t n t p x As t, LHS of (2.2 tends to zero if E[(T (x n ] exists and is finite. Hence lim t t n tp x. On the other hand, t n tp x. Therefore lim t t n tp x. This gives that lim t t n tp x =. Hence (2.3 E[(T (x n ] = t n d( t p x = nt n 1 tp x dt. For n = 1, if the expected value of T (x is assumed to exist, then E[T (x] = tp x dt as the existence of E[T (x] implies that lim t t( t p x =. ([2], Chapter 3, pp 68. E[T (x] is called the complete expectation of life and is denoted by e x. In reliability theory, this is termed as the mean residual life function. We also get ( 2 (2.4 V [T (x] = 2 t( t p x dt tp x. The expectation and variance of T (x are same as derived in [2, 3]. If µ k denotes the k th order central moment of T (x, then coefficient of skewness is γ 1 = µ 3 µ 3/2 2 (2.5 µ 3 = where µ 2 = V [T (x] and 3t 2 tp x dt 3 tp x dt ( 3 2t t p x dt + 2 tp x dt.

4 4 A measure of kurtosis is given by γ 2 = µ 4 µ 2 2 where (2.6 µ 4 = ( 4t 3 tp x dt 4 ( tp x dt 3t 2 tp x dt ( ( 2 ( t t p x dt tp x dt 3 tp x dt. Using (1.1 and (2.1, we can find the probability density function (pdf and force of mortality of T (x for different distributions of X. For x, t >, the expressions for survival probability ( t p x and force of mortality (µ x+t are listed in Table 1. Table 1: Survival function and force of mortality of T (x Distribution of X tp x µ x+t Exp(λ, λ > e λt λ Gompertz (α, λ, α >, λ > 1 e (αλx (λ t 1/logλ αλ (x+t Weibull (α, λ, α >, λ > Pareto (θ, θ > Burr (γ, θ, γ >, θ > Exponentiated Exp. (α, λ, α >, λ > e (λ(x+tα e (λxα αλ α (x + t α 1 x θ (x + t θ θ (x + t (1 + x γ θ γθ(1 + x γ θ (x + t γ 1 (1 + (x + t γ θ (1 + (x + t γ 1 (1 e λ(x+t α 1 (1 e λx α α(1 e λ(x+t α 1 λe λ(x+t 1 (1 e λ(x+t α From Table 1, we can conclude that : For X Exp (λ and all λ, t p x is independent of x, the initial age; For X Gompertz (α, λ, t p x is a decreasing function of x for all α > and λ > 1; For X Weibull (α, λ, t p x is an increasing (decreasing function of x for all λ and < α < 1 (α > 1 ; For X Pareto (θ, t p x is an increasing function of x for all θ; For X Burr (γ, θ, t p x is an increasing function of x for all θ, γ; For X Exponentiated Exp (α, λ, t p x is an increasing (decreasing function of x for all α and λ > 1 ( < λ < 1. Since t p x = P [T (x > t] is the survival function of T (x, hence it is decreasing in t for the distribution under consideration. The pdfs of future lifetime distributions of X mentioned in Table 1 can be obtained by multiplying t p x and µ x+t.

6 Pareto Type I Future Lifetime Distribution. Let X follow Pareto (θ, then the likelihood function for T(x is n θx θ L(t(x, θ =, θ >, and x >. (x + t i (x θ+1 Taking log on both sides and differentiating with respect to θ, we get ( dlog(l(t(x, θ 1 (3.1 = n dθ θ + logx log(x + t i (x. Equating (3.1 to zero and solving, the MLE of θ is found to be n ˆθ = n log(x + t. i(x x 3.3. Weibull Future Lifetime Distribution. For X following Weibull (α, λ, the pdf of T (x is given by f T (x(t(x, λ, α = αλ α (x + t(x (α 1 e λα ((x+t(x α x α, α, λ >, t(x >. The corresponding log-likelihood can be written as log(l(t(x, λ, α = log{αλ α (x + t i (x α 1 e λα ((x+t i(x α x α }. ML estimators of λ and α can be obtained by equating the derivatives of loglikelihood with respect to λ and α to zero and solving where ( = nα n λ αλα 1 (x + t i (x α nx α ; dlog(l(t(x, λ, α dλ dlog(l(t(x, λ, α dα ( 1 = n α + logλ + (λxα log(λx + log(x + t i (x (1 (λ(x + t i (x α logλ 3.4. Burr Future Lifetime Distribution. If X follows Burr (γ, θ then f T (x (t(x, θ, γ = γθ(1 + xγ θ (x + t(x γ 1 (1 + (x + t(x γ θ+1, γ > and θ > If L(t(x, θ, γ denotes the likelihood function,then (3.2 dlog(l(t(x, θ, γ dγ = n γ + nθxγ logx (1 + x γ + log(x + t i (x ( 1 (x + t i(x γ (θ (x + t i (x γ ; (λ(x + t i (x α. (3.3 dlog(l(t(x, θ, γ dθ = n θ + nlog(1 + xγ log(1 + (x + t i (x γ.

7 7 The ML estimators of θ and γ shall be obtained by equating (3.2 and (3.3 to zero and finding numerical solutions to the non-linear equations in Section Exponentiated Exponential Future Lifetime Distribution. Let X follow Exponentiated Exponential (α, λ, then the corresponding likelihood function is L(t(x, α, λ = n αλ(1 e λ(x+ti(x α 1 e λ(x+ti(x 1 (1 e λx α. The ML estimators of α and λ can be obtained using the non-linear expressions given by ( dlog(l(t(x, λ, α 1 =n dλ λ + αxe λx (1 e λx α 1 1 (1 e λx α ( (α 1e λ(x+t i(x (3.4 + (x + t i (x 1 ; 1 e λ(x+ti(x (3.5 dlog(l(t(x, λ, α dα ( 1 = n α + (1 e λx α log(1 e λx 1 (1 e λx α + log(1 e λ(x+ti(x Gompertz Future Lifetime Distribution. The pdf of T (x, when X follows Gompertz (α, λ, is Hence (3.6 (3.7 f T (x(t(x = αλ (x+t(x e L(t(x, α, λ = dlog(l(t(x, α, λ dα dlog(l(t(x, α, λ dλ αλ x (λ t(x 1 logλ α >, λ > 1, t(x and x >. αλ n x (λ ti(x 1 αλ x+ti(x e logλ is the corresponding likelihood function. = n α n = αλx 1 logλ + λ x (λ ti(x 1 ; logλ ( (λ ti(x 1 1( logλ x λti(x t i (x (x + t i (x. λ Equating (3.6 and (3.7 to zero, the MLEs of α and λ are obtained in simulations section using numerical approximation. The following section discusses the estimation of parameters of different continuous future lifetime distributions when some observations are censored in a given set of observations.

8 8 4. Estimation of parameters in censored case In insurance sector, it is possible that some insured persons do not continue the contract due to some reasons. They surrender the policy or don t pay the renewal premium after few years to draw benefit from the contract. Such individuals are considered to be censored and their time of death is not known. The time of occurrence of such events is not known in advance and is a random variable, named as censoring variable. This type of censoring is known as right censoring. For an individual aged x, let the random variable C(x be the time at which an individual either leaves the contract or stops paying premium and is randomly censored. T (x and C(x are assumed to be independent. In the subsequent discussion, we study the estimation under term assurance contract and whole life assurance contract in case of censoring Estimation for term assurance contract. We consider a term assurance contract, tenable for t years. Under this contract, an insurer pays the sum assured only if insured dies within the term of the contract. This means that at the time of inception of the policy, a policy holder aged x gets the benefit if he dies between x and x + t. Hence, the insured ones surviving at time t and those who surrender the policy due to one or the other reason are lost to further follow up by the company. Surviving persons are right censored at a predetermined time t and those who surrender the policy are randomly censored. We attempt to estimate the unknown parameters of the distributionsof future lifetime. For this type of situation, we attempt to estimate the unknown parameters of the distributions of future lifetime. Let the random variable Z be defined as T (x if T (x < C(x and T (x < t ; Z = min(t (x, C(x, t = C(x if C(x < T (x and C(x < t ; t if t < T (x and t < C(x. For i th life aged x, we take t i (x to be the future lifetime and c i (x, the censoring time. This means that the policy is either surrendered or gets lapsed. We assume that, there are n individuals aged x in the cohort and a total of m leave the cohort due to death or some other reason. Out of m, r is assumed to be the random no. of individuals that leave the cohort due to death. Hence, there will be (n m observations in the cohort at time t and the likelihood function [1] can be written as r m L(z = f T (x(t i (xḡc(x(t i (x g C(x (c i (x F T (x (c i (x (4.1 [ FT (x (t ḠC(x(t ] n m where F T (x (. : survival function of T (x, Ḡ C(x (. : survival function of C(x, g C(x (. : probability density function of C(x. In the following discussion, we explore the maximum likelihood estimation of the

9 9 parameters of exponential, Weibull, Pareto, burr and Gompertz future lifetime distributions when some observations are censored. For all the cases, the distribution of C(x is assumed to follow U(, η where the value of η depends on the percentage of censoring Exponential Future Lifetime Distribution If λ is the parameter of exponential distribution followed by X, the likelihood function for T (x using (4.1, is written as ( r L(z, λ =λ r e λ r ti(x Maximizing the loglikelihood, we have ˆλ = ( 1 t i(x η r r t i(x + m c. i(x + (n mt Weibull Future Lifetime Distribution ( e λ m Let X follow Weibull (α, λ and C(x follow U(, η, then ci(x η m r ( ( e λt 1 t n m (x. η Hence L(z, α, λ = e n(λxα dlog(l(z, α, λ dα r m ( αλ α (x + t i (x α 1 e (λ(x+ti(xα 1 t i(x η e (λ(x+ci(xα η = r α + rlogλ + n(λxα log(λx ( (e (λ(x+tα 1 t n m (x. η + (log(x + t i (x λ α (x + t i (x α log(λ(x + t i (x m (λ(x + c i (x α log(λ(x + c i (x (4.2 (n m(λ(x + t α log(λ(x + t ; (4.3 dlog(l(z, α, λ dλ = αr λ + αλα 1 (nx α (n m(x + t α αλ α 1 ( (x + t i (x α + m (x + c i (x α ML estimates of α and λ are found by solving the non-linear equations obtained as a result of equating (4.2 and (4.3 to zero. This is done using numerical approximation and corresponding RMSEs are calculated in Section 5.

10 Pareto Type I Future Lifetime Distribution If X follows Pareto (θ, θ >, the MLE of θ is r ˆθ = r log(x + t i(x nlogx + m log(x + c i(x + (n mlog(x + t Burr Future Lifetime Distribution For X Burr(γ, θ γ, θ >, the ML estimators of θ and γ can be obtained using non-linear equations given by dlog(l(z, γ, θ = r m dθ θ + nlog(1 + xγ log(1 + (x + t i (x γ log(1 + (x + c i (x γ dlog(l(z, γ, θ dγ (n mlog(1 + (x + t γ ; = r γ + nθxγ logx (1 + x γ + θ m ( 1 θ(x + ti (x γ log(x + t i (x (1 + (x + t i (x γ (x + c i (x γ log(x + c i (x (1 + (x + c i (x γ (n mθ(x + t γ log(x + t (1 + (x + t γ. The estimation of γ and θ shall be explored in Section 5 by using numerical methods Gompertz Future Lifetime Distribution If X follows Gompertz (α, λ, then using the log-likelihood function, we have (4.4 (4.5 dlog(l(z, α, λ dα dlog(l(z, α, λ dλ = nλx logλ + r α + (x + t i (xlogλ ( λx λ ti(x + logλ m λ ci(x (n m λx+t logλ. x 1 xlogλ 1 =nαλ (logλ 2 (n mαλ x+t 1 logλ(x + t 1 (logλ 2 + m x + t i (x λ αλ x+ti(x 1 logλ(x + t i(x 1 (logλ 2 αλ x+ci(x 1 logλ(x + c i(x 1 (logλ 2 ; By applying Newton-Raphson Method and using (4.4 and (4.5, the estimates are found in Section 5.

11 Estimation for whole life assurance contracts in case of censoring. In insurance, there are some policies which pay benefit till or at the time of the death of the insured, known as whole life assurance contracts. This happens provided the insured does not leave the cohort due to any other reason except death. To evaluate benefits accurately under this situation, the insurance company takes account of all the conditions under which insured may leave the contract. For example, the insured may surrender the policy at a random time giving rise to a censored observation. We take account of this censoring for estimation of parameters of the future lifetime distributions under whole life assurance contract. We write { T (x if T (x < C(x Z = min(t (x, C(x = C(x if C(x < T (x where T (x and C(x are defined in Section 4.1. The likelihood function, under random right censoring [1] is written as (4.6 L(z = r f T (x(t i (xḡc(x(t i (x n g C(x (c i (x F T (x (c i (x. In the following subsections, ML estimation of unknown parameters of different future lifetime distributions shall be discussed by assuming that C(x U(, η where the value of η depends upon the assumed percentage of censoring Exponential Future Lifetime Distribution Let X follow Exponential (λ, then using (4.6 L(z, λ = λ r e λ r r ( ti(x 1 t i(x e λ n ci(x η η n r. Differentiating the log-likelihood with respect to λ and solving the normal equation, we get ˆλ = r r t i(x + n c i(x Weibull Future Lifetime Distribution If X follows Weibull (α, λ, then considering the log likelihood and differentiating with respect to α and λ, we get (4.7 dlog(l(z, α, λ dα = r α + rlogλ + n(λxα log(λx + log(x + t i (x (x + t i (x α λ α log(λ(x + t i (x (λ(x + c i (x α log(λ(x + c i (x;

12 12 (4.8 dlog(l(z, α, λ dλ ( = αr λ αλα 1 nx α + (x + t i (x α (x + c i (x α. When (4.7 and (4.8 are equated to zero, the resulting non-linear equations can be solved using numerical methods. The results are demonstrated in Section Pareto Type I Future Lifetime Distribution If X follows Pareto (θ, θ >, then differentiation of log likelihood with respect to θ and equating the resulting expressions to zero, gives r ˆθ = r log(x + t i(x nlogx + n log(x + c i(x Burr Future Lifetime Distribution Let X follow Burr (γ, θ, γ, θ >, then (4.9 (4.1 dlog(l(z, γ, θ dθ dlog(l(z, γ, θ dγ = r θ + nlog(1 + xγ = r γ + rθxγ log(x (1 + x γ θ log(1 + (x + t i (x γ log(1 + (x + c i (x γ ; + log(x + t i (x (θ + 1 (x + t i (x γ log(x + t i (x (1 + (x + t i (x γ (x + c i (x γ log(x + c i (x (1 + (x + c i (x γ + θ(n rxγ logx (1 + x γ = r γ + nθxγ logx [ (1 + x γ + (x + t i (x γ ] log(x + t i (x 1 (θ + 1 (1 + (x + t i (x γ θ (x + c i (x γ log(x + c i (x (1 + (x + c i (x γ. In Section 5, the estimates of γ and θ are obtained by equating (4.9 and (4.1 to zero and solving the resulting equations using numerical methods.

13 Gompertz Future Lifetime Distribution If X follows Gompertz (α, λ, dlog(l(z, α, λ dα dlog(l(z, α, λ dλ = nλx logλ + r α + (x + t i (xlogλ [ ] λx λ ti(x + λ ci(x. logλ x 1 xlogλ 1 =nαλ (logλ 2 + ( x + t i (x λ αλx 1 (logλ 2 λ ti(x (logλ(x + t i (x 1 + λ ci(x (logλ(x + c i (x 1 The maximum likelihood estimates are presented in Section Simulations In this section, we generate random samples of sizes 5, 1 and 5 from different continuous future lifetime distributions and estimate their parameters, using BFGS method in R. The root mean square errors (RMSEs are evaluated to check the extent of the deviation of the estimate from the true value. The number of replications is 1 for starting age x = 2, 4 and Uncensored case for whole life assurance contract. Tables 2-5 give the estimates and the corresponding RMSEs when the data sets are generated from Weibull (α = 1.5 and λ = 2.5, Burr (θ = 3 and γ = 2, exponentiated exponential (α = 1.5 and λ =.5 and Gompertz (α =.5 and λ = 1.8 future lifetime distributions. Table 2: Estimates and RMSEs for Weibull future lifetime dist. x n = 5 n = 1 n = (.626 (.2134 (.435 (.1947 (.112 ( (.173 (.134 (.97 (.891 (.55 ( (.85 (.82 (.44 (.543 (.39 (.131 Table 3: Estimates and RMSEs for Burr future lifetime dist. x n = 5 n = 1 n = (.2532 (.243 (.2232 (.214 (.1532 ( (.1917 (.1859 (.163 (.1215 (.951 ( (.959 (.1123 (.751 (.825 (.275 (.372 ;

14 14 Table 4: Estimates and RMSEs for Exponentiated Exp. future lifetime dist. x n = 5 n = 1 n = (.2468 (.74 (.124 (.512 (.991 ( (.2215 (.76 (.985 (.511 (.925 ( (.276 (.685 (.832 (.434 (.64 (.198 Table 5: Estimates and RMSEs for Gompertz future lifetime dist. x n = 5 n = 1 n = (.117 (.3 (.8 (.2 (.34 ( (.15 (.5 (.13 (.3 (.44 ( (.256 (.2 (.17 (.8 (.78 (.2 On the basis of values in Tables 2-5, we conclude that for all future lifetime distributions under study RMSEs decrease as sample size increases, that is, the estimates get closer to the assumed values of the parameters for large samples; RMSEs decrease as initial age increases, except when X follows Gompertz distribution Censored case. As mentioned in Section 4.1, C(x, the censoring variable is assumed to follow U(, η. The parameter η for C(x is determined by assuming 3% censoring in the data, that is, by solving (5.1 P [T (x > C(x] =.3. We find the values of η for different combinations of parameters of T (x. The values are generated from the future lifetime and Uniform distributions. These generated sets of values are used to estimate the values of parameters when x = 2, 4, 6; n = 5, 1, 5 and t = 1, 2, 3, 4 and Term Life Assurance Contract Tables 6 8 display the values of η, estimates of parameters of future lifetime distributions and the corresponding RMSEs when the random samples are generated from Weibull (α = 1.5 and λ = 2.5, Burr (θ = 3 and γ = 2, and Gompertz (α =.5 and λ = 1.8 future lifetime distributions. The values in the parentheses give the RMSEs.

15 15 Table 6: Estimates and RMSEs for censored Weibull future lifetime dist. n = 5 n = 1 n = 5 t x = 2, η = (.55 (.25 (.339 (.1596 (.12 ( (.462 (.2243 (.281 (.1375 (.97 ( (.451 (.272 (.262 (.1359 (.96 ( (.415 (.1689 (.259 (.1283 (.6 ( (.399 (.1374 (.16 (.932 (.58 (.19 t x = 4, η = (.223 (.288 (.172 (.271 (.73 ( (.221 (.283 (.155 (.269 (.5 ( (.198 (.251 (.143 (.25 (.46 ( (.189 (.198 (.14 (.225 (.31 ( (.166 (.152 (.14 (.178 (.27 (.56 t x = 6, η = (.225 (.354 (.165 (.185 (.68 ( (.188 (.321 (.141 (.163 (.63 ( (.179 (.313 (.14 (.154 (.62 ( (.145 (.292 (.138 (.153 (.58 ( (.97 (.112 (.75 (.132 (.55 (.94

16 16 Table 7: Estimates and RMSEs for censored Burr future lifetime dist. n = 5 n = 1 n = 5 t x = 2, η = (.4877 (.4392 (.4562 (.4278 (.4223 ( (.4583 (.4363 (.433 (.3959 (.3686 ( (.449 (.431 (.4184 (.363 (.3455 ( (.4223 (.4246 (.3799 (.357 (.2966 ( (.4164 (.447 (.3375 (.3427 (.2114 (.2697 t x = 4, η = (.1755 (.2133 (.1284 (.1396 (.1194 ( (.1612 (.1959 (.126 (.1232 (.845 ( (.1542 (.1897 (.1243 (.189 (.749 ( (.1486 (.1751 (.1212 (.189 (.69 ( (.1468 (.1678 (.789 (.186 (.322 (.326 t x = 6, η = (.1561 (.2291 (.12 (.1775 (.394 ( (.1548 (.2113 (.952 (.1261 (.26 ( (.124 (.1948 (.925 (.11 (.122 ( (.126 (.1938 (.778 (.951 (.117 ( (.73 (.1317 (.599 (.848 (.86 (.171

17 17 Table 8: Estimates and RMSEs for censored Gompertz future lifetime dist. n = 5 n = 1 n = 5 t x = 2, η = (.1395 (.3 (.1155 (.3 (.812 ( (.725 (.3 (.575 (.3 (.268 ( (.453 (.3 (.345 (.2 (.122 ( (.296 (.3 (.23 (.2 (.65 ( (.165 (.2 (.14 (.2 (.5 (.1 t x = 4, η = (.786 (.3 (.716 (.3 (.42 ( (.366 (.3 (.333 (.3 (.128 ( (.257 (.3 (.169 (.2 (.69 ( (.143 (.2 (.11 (.2 (.52 ( (.17 (.2 (.96 (.2 (.48 (.1 t x = 6, η = (.471 (.3 (.424 (.3 (.161 ( (.33 (.3 (.228 (.3 (.16 ( (.327 (.3 (.197 (.3 (.77 ( (.324 (.3 (.247 (.3 (.84 ( (.314 (.3 (.228 (.3 (.78 (.2 From Tables 6 8, it can be observed that RMSEs decrease with an increase in sample size n, initial age x and term of the contract t Whole Life Assurance contract Tables 9 11 show the estimates and the corresponding RMSEs (in parantheses when the data sets are generated from Weibull (α = 1.5 and λ = 2.5, Burr (θ = 3 and γ = 2 and Gompertz (α =.5 and λ = 1.8 future lifetime distributions.

18 18 Table 9: Estimates and RMSEs for censored Weibull future lifetime dist. n = 5 n = 1 n = η =.12 for x = 2 (.182 (.97 (.114 (.75 (.75 ( η =.85 for x = 4 (.224 (.141 (.154 (.88 (.79 ( η =.7 for x = 6 (.264 (.1188 (.356 (.511 (.124 (.156 Table 1: Estimates and RMSEs for censored Burr future lifetime dist. n = 5 n = 1 n = η = 12.5 for x = 2 (.444 (.4117 (.4418 (.4477 (.4643 ( η = 24.5 for x = 4 (.2199 (.2688 (.262 (.1936 (.175 ( η = 36.5 for x = 6 (.2152 (.269 (.1436 (.1845 (.931 (.946 Table 11: Estimates and RMSEs for censored Gompertz future lifetime dist. n = 5 n = 1 n = η = 13 for x = 2 (.135 (.3 (.93 (.2 (.41 ( η = 75 for x = 4 (.189 (.7 (.128 (.3 (.56 ( η = 34 for x = 6 (.339 (.83 (.232 (.23 (.96 (.4 On the basis of Tables 9 11, it is concluded that as sample size increases, RMSEs decrease; as initial age increases, RMSEs increase except for Burr future lifetime distribution.

19 19 Conclusions Estimation of the parameters of future lifetime distributions is helpful in modeling the remaining lifetime of the insured. This is helpful to insurer for setting the premium in order to make the insured policy mutually advantageous. We study the distributional properties of the complete future lifetime and explore the maximum likelihood estimation of the parameters when the insured does not leave the cohort till the end of the contract or leaves at a random time. Acknowledgement The first author is grateful to University Grants Commission, Government of India, for providing financial support for this work.the authors are extremely grateful to the refrees for providing valuable guidance and comments that helped in improving the manuscript. References [1] Borowiak, D. S. and Shapiro, A. F. Financial and Actuarial Statistics: An Introduction. A Series of Textbooks and Monographs, Marcel Dekker Publication, New York, 23. [2] Bowers, N. L., Gerber, H. U., Hickman, J. C, Jones, D. A. and Nesbitt, C. J. Actuarial Mathematics, The Society of Actuaries, USA, [3] Dickson, D. C. M., Hardy, M. R. and Waters, H. R. Actuarial Mathematics for Life Contingent Risks, International Series on Actuarial Science, Cambridge University Press, UK, 29. [4] Gerber, H.U. Life Insurance Mathematics, Springer, Germany, 199. [5] International Actuarial Notation. Journal of the Institute of Actuaries, 75, , [6] Jain, K. and Kapoor, H. P. S. Ageing Properties of Curtate Future Life Time. Sri Lankan Journal of Applied Statistics, 14 (1, 55-84, 213. [7] Johnson, N. L., Balakrishnan, N. and Kotz, S. Continuous Univariate Distributions 1, Wiley Series in Probability and Statistics, John Wiley and Sons, USA, [8] Johnson, N. L., Balakrishnan, N. and Kotz, S. Continuous Univariate Distributions 2, Wiley Series in Probability and Statistics, John Wiley and Sons, USA, [9] Kleiber, C. and Kotz, S. Statistical Size Distributions in Economics and Actuarial Sciences, John Wiley and Sons, New Jersey, Hoboken, 23. [1] Klein, J. P. and Moeschberger, M. L. Survival Analysis: Techniques for Censored and Truncated Data, Second edition, Springer Verlag, USA, 23. [11] Lai, C. D. and Xie, M. Stochastic Ageing and Dependence for Reliability, Springer, USA, 26. [12] Zhu, Y. J. Actuarial Model: Life Insurance and Annuity, Series in Actuarial Science, International Press, USA, 27.

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