# 2 Right Censoring and Kaplan-Meier Estimator

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1 2 Right Censoring and Kaplan-Meier Estimator In biomedical applications, especially in clinical trials, two important issues arise when studying time to event data (we will assume the event to be death. It can be any event of interest): 1. Some individuals are still alive at the end of the study or analysis so the event of interest, namely death, has not occurred. Therefore we have right censored data. 2. Length of follow-up varies due to staggered entry. So we cannot observe the event for those individuals with insufficient follow-up time. Note: It is important to distinguish calendar time and patient time Figure 2.1: Illustration of censored data x x x x x o x x o x o o Study Calendar Study starts time ends 0 Patient time (measured from entry to study) In addition to censoring because of insufficient follow-up (i.e., end of study censoring due to staggered entry), other reasons for censoring includes loss to follow-up: patients stop coming to clinic or move away. deaths from other causes: competing risks. Censoring from these types of causes may be inherently different from censoring due to staggered entry. We will discuss this in more detail later. PAGE 11

2 Censoring and differential follow-up create certain difficulties in the analysis for such data as is illustrated by the following example taken from a clinical trial of 146 patients treated after they had a myocardial infarction (MI). The data have been grouped into one year intervals and all time is measured in terms of patient time. Table 2.1: Data from a clinical trial on myocardial infarction (MI) Number alive and under Year since observation at beginning Number dying Number censored entry into study of interval during interval or withdrawn [0, 1) [1, 2) [2, 3) [3, 4) [4, 5) [5, 6) [6, 7) [7, 8) [8, 9) [9, 10) Question: Estimate the 5 year survival rate, i.e., S(5) = P [T 5]. Two naive and incorrect answers are given by 1. F (5) = P [T <5] = 2. F (5) = P [T <5] = 76 deaths in 5 years 146 individuals =52.1%, Ŝ(5) = 1 F (5) = 47.9%. 76 deaths in 5 years (withdrawn in 5 years) = 65%, Ŝ(5) = 1 F (5) = 35%. Obviously, we can observe the following PAGE 12

3 1. The first estimate would be correct if all censoring occurred after 5 years. Of cause, this was not the case leading to overly optimistic estimate (i.e., overestimates S(5)). 2. The second estimate would be correct if all individuals censored in the 5 years were censored immediately upon entering the study. This was not the case either, leading to overly pessimistic estimate (i.e., underestimates S(5)). Our clinical colleagues have suggested eliminating all individuals who are censored and use the remaining complete data. This would lead to the following estimate F (5) = P [T 5] = 76 deaths in 5 years (censored) =88.4%, Ŝ(5) = 1 F (5) = 11.6%. This is even more pessimistic than the estimate given by (2). Life-table Estimate More appropriate methods use life-table or actuarial method. The problem with the above two estimates is that they both ignore the fact that each one-year interval experienced censoring (or withdrawing). Obviously we need to take this information into account in order to reduce bias. If we can express S(5) as a function of quantities related to each interval and get a very good estimate for each quantity, then intuitively, we will get a very good estimate of S(5). By the definition of S(5), we have: S(5) = P [T 5] = P [(T 5) (T 4)] = P [T 4] P [T 5 T 4] = P [T 4] {1 P [4 T<5 T 4]} = P [T 4] q 5 = P [T 3] P [T 4 T 3] q 5 = P [T 3] {1 P [3 T<4 T 3]} q 5 = P [T 3] q 4 q 5 = = q 1 q 2 q 3 q 4 q 5 where q i =1 P [i 1 T < i T i 1],i =1, 2,..., 5. So if we can estimate q i well, then we will get a very good estimate of S(5). Note that 1 q i is the mortality rate m(x) atyear x = i 1 by our definition. PAGE 13

4 Table 2.2: Life-table estimate of S(5) assuming censoring occurred at the end of interval duration [t i 1,t i ) n(x) d(x) w(x) m(x) = d(x) n(x) 1 m(x) ŜR (t i )= (1 m(x)) [0, 1) [1, 2) [2, 3) [3, 4) [4, 5) Case 1: Let us first assume that anyone censored in an interval of time is censored at the end of that interval. Then we can estimate each q i =1 m(i 1) in the following way: d(0) Bin(n(0),m(0)) = m(0) = d(0) n(0) = =0.185, q 1 =1 m(0) = d(1) H Bin(n(1),m(1)) = m(1) = d(1) n(1) = =0.155, q 2 =1 m(1) = where H means data history (i.e, data before the second interval). The life table estimate would be computed as shown in Table 2.2. So the 5 year survival probability estimate ŜR (5) = (If the assumption that anyone censored in an interval of time is censored at the end of that interval is true, then the estimator ŜR (5) is approximately unbiased to S(5).) Of course, this estimate ŜR (5) will have variation since it was calculated from a sample. We need to estimate its variation in order to make inference on S(5) (for example, construct a 95% CI for S(5)). However, ŜR (5) is a product of 5 estimates ( q 1 q 5 ), whose variance is not easy to find. But we have log(ŝr (5)) = log( q 1 )+log( q 2 )+log( q 3 )+log( q 4 )+log( q 5 ). So if we can find out the variance of each log( q i ), we might be able to find out the variance PAGE 14

5 of log(ŝr (5)) and hence the variance of ŜR (5). For this purpose, let us first introduce a very popular method in statistics: delta method: Delta Method: If θ a N(θ, σ 2 ) then f( θ) a N(f(θ), [f (θ)] 2 σ 2 ) Proof of delta method: If σ 2 is small, θ will be close to θ with high probability. We hence can expand f( θ) aboutθ using Taylor expansion: f( θ) f(θ)+f (θ)( θ θ). We immediately get the (asymptotic) distribution of f( θ) from this expansion. Returning to our problem. Let φ i = log( q i ). Using the delta method, the variance of φ i is approximately equal to ( ) 2 var( φ 1 i )= var( q i ). q i Therefore we need to find out and estimate var( q i ). Of course, we also need to find out the covariances among φ i and φ j (i j). For this purpose, we need the following theorem: Double expectation theorem (Law of iterated conditional expectation and variance): If X and Y are any two random variables (or vectors), then E(X) = E[E(X Y )] Var(X) =Var[E(X Y )] + E[Var(X Y )] Since q i =1 m(i 1), we have var( q i ) = var( m(i 1)) PAGE 15

6 = E[var( m(i 1) H)] + var[e( m(i 1) H)] [ ] m(i 1)[1 m(i 1)] = E +var[m(i 1)] n(i 1) [ ] 1 = m(i 1)[1 m(i 1)]E, n(i 1) which can be estimated by m(i 1)[1 m(i 1)]. n(i 1) Hence the variance of φ i = log( q i ) can be approximately estimated by ( ) 2 1 m(i 1)[1 m(i 1)] q i n(i 1) = m(i 1) [1 m(i 1)]n(i 1) = d (n d)n. Now let us look at the covariances among φ i and φ j (i i). It is very amazing that they are all approximately equal to zero! For example, let us consider the covariance between φ 1 and φ 2. Since φ 1 = log( q 1 )and φ 2 = log( q 2 ), using the same argument for the delta method, we know that we only need to find out the covariance between q 1 and q 2, or equivalently, the covariance between m(0) and m(1). This can be seen from the following: E[ m(0) m(1)] = E[E[ m(0) m(1) n(0),d(0),w(0)]] = E[ m(0)e[ m(1) n(0),d(0),w(0)]] = E[ m(0)m(1)] = m(1)e[ m(0)] = m(1)m(0) = E[ m(0)]e[ m(1)]. Therefore, the covariance between m(0) and m(1) is zero. Similarly, we can show other covariances are zero. Hence, var(log(ŝr (5))) = var( φ 1 )+var( φ 2 )+var( φ 3 )+var( φ 4 )+var( φ 5 ). Let θ = log(ŝr (5)). Then ŜR (5) = e θ.so var(ŝr (5)) = (e θ ) 2 var(log(ŝr (5))) = (S(5)) 2 [var( φ 1 )+var( φ 2 )+var( φ 3 )+var( φ 4 )+var( φ 5 )], PAGE 16

7 which can be estimated by [ var(ŝr (5)) = (ŜR (5)) 2 d(0) (n(0) d(0))n(0) + + = (ŜR (5)) 2 4 d(3) (n(3) d(3))n(3) + i=0 d(1) (n(1) d(1))n(1) + ] d(4) (n(4) d(4))n(4) d(2) (n(2) d(2))n(2) d(i) [n(i) d(i)]n(i). (2.1) Case 2: Let us assume that anyone censored in an interval of time is censored right at the beginning of that interval. Then the life table estimate would be computed as shown in Table 2.3. So the 5 year survival probability estimate = (In this case, the estimator Ŝ L (5) is approximately unbiased to S(5).) The variance estimate of ŜL (5) is similar to that of ŜR (5)exceptthatweneedtochange the sample size for each mortality estimate to n w in equation (2.1). Table 2.3: Life-table estimate of S(5) assuming censoring occurred at the beginning of interval duration [t i 1,t i ) n(x) d(x) w(x) m(x) = d(x) n(x) w(x) 1 m(x) ŜL (t i )= (1 m(x)) [0, 1) [1, 2) [2, 3) [3, 4) [4, 5) The naive estimates range from 35% to 47.9% for the five year survival probability with the complete case (i.e., eliminating anyone censored) estimator giving an estimate of 11.6%. The life-table estimate ranged from 40% to 43.2% depending on whether we assume censoring occurred at the left (i.e., beginning) or right (i.e., end) of each interval. More than likely censoring occurs during the interval. Thus ŜL and ŜR are not correct. A compromise is to use the following modification: PAGE 17

8 Table 2.4: Life-table estimate of S(5) assuming censoring occurred during the interval duration [t i 1,t i ) n(x) d(x) w(x) m(x) = d(x) n(x) w(x)/2 1 m(x) ŜLT (t i )= (1 m(x)) [0, 1) [1, 2) [2, 3) [3, 4) [4, 5) That is, when calculating the mortality estimate in each interval, we use (n(x) w(x)/2) as the sample size. This number is often referred to as the effective sample size. So the 5 year survival probability estimate ŜLT (5) = 0.417, which is between ŜL =0.400 and Ŝ R = Figure 2.2: Life-table estimate of the survival probability for MI data Survival probability Time (years) Figure 2.2 shows the life-table estimate of the survival probability assuming censoring occurred during interval. Here the estimates were connected using straight lines. No special significance should be given to this. From this figure, the median survival time is estimated to PAGE 18

9 be about 3 years. The variance estimate of the life-tabble estimate ŜLT (5) is similar to equation (2.1) except that the sample size n(i) is changed to n(i) w(i)/2. That is var(ŝlt (5)) = (ŜLT (5)) 2 4 i=0 d(i) [n(i) w(i)/2 d(i)][n(i) w(i)/2]. (2.2) Of course, we can also use the above formula to calculate the variance of ŜLT (t) atother time points. For example: { } var(ŝlt (1)) = (ŜLT (1)) 2 d(0) [n(0) w(0)/2 d(0)][n(0) w(0)/2] = (146 3/2 27)(146 3/2) = = Therefore SE(ŜLT (1)) = = The calculation presented in Table 2.4 can be implemented using Proc Lifetest in SAS: options ls=72 ps=60; Data mi; input survtime number status; cards; ; proc lifetest method=life intervals=(0 to 10 by 1); time survtime*status(0); freq number; run; PAGE 19

10 Note that the number of observed events and withdrawals in [t i 1,t i ) were entered after t i 1 instead of t i. Part of the output of the above SAS program is The LIFETEST Procedure Life Table Survival Estimates Effective Conditional Interval Number Number Sample Probability [Lower, Upper) Failed Censored Size of Failure Interval Conditional Probability Standard Survival Standard Median Residual [Lower, Upper) Error Survival Failure Error Lifetime Here the numbers in the column under Conditional Probability of Failure are the estimated mortality m(x) = d(x)/(n(x) w(x)/2). The above lifetable estimation can also be implemented using R. Here is the R code: > tis <- 0:10 > ninit <- 146 > nlost <- c(3,10,10,3,3,11,5,8,1,6) > nevent <- c(27,18,21,9,1,2,3,1,2,2) > lifetab(tis, ninit, nlost, nevent) PAGE 20

11 The output from the above R function is nsubs nlost nrisk nevent surv pdf hazard se.surv NA NA se.pdf se.hazard NA NA Note: Here the numbers in the column of hazard are the estimated hazard rates at the midpoint of each interval by assuming the true survival function S(t) is a straight line in each interval. You can find an explicit expression for this estimator using the relation λ(t) = f(t) S(t), and the assumption that the true survival function S(t) is a straight line in [t i 1,t i ): S(t) =S(t i 1 )+ S(t i) S(t i 1 ) t i t i 1 (t t i 1 ), for t [t i 1,t i ). These estimates are very close to the mortality estimates we obtained before (the column under Conditional Probability of Failure in the SAS output.) Kaplan-Meier Estimator The Kaplan-Meier or product limit estimator is the limit of the life-table estimator when intervals are taken so small that only at most one distinct observation occurs within an interval. Kaplan and Meier demonstrated in a paper in JASA (1958) that this estimator is maximum likelihood estimate. PAGE 21

12 Figure 2.3: An illustrative example of Kaplan-Meier estimator x x o x o x x o x o Patient time (years) m(x) : Ŝ(t) : We will illustrate through a simple example shown in Figure 2.3 how the Kaplan-Meier estimator is constructed. By convention, the Kaplan-Meier estimate is a right continuous step function which takes jumps only at the death time. The calculation of the above KM estimate can be implemented using Proc Lifetest in SAS as follows: Data example; input survtime censcode; cards; ; Proc lifetest; PAGE 22

13 time survtime*censcode(0); run; And part of the output from the above program is The LIFETEST Procedure Product-Limit Survival Estimates Survival Standard Number Number SURVTIME Survival Failure Error Failed Left * * * * * Censored Observation The above Kaplan-Meier estimate can also be obtained using R function survfit(). The code is given in the following: > survtime <- c(4.5, 7.5, 8.5, 11.5, 13.5, 15.5, 16.5, 17.5, 19.5, 21.5) > status <- c(1, 1, 0, 1, 0, 1, 1, 0, 1, 0) > fit <- survfit(surv(survtime, status), conf.type=c("plain")) Then we can use R function summary() to see the output: > summary(fit) Call: survfit(formula = Surv(survtime, status), conf.type = c("plain")) time n.risk n.event survival std.err lower 95% CI upper 95% CI Let d(x) denote the number of deaths at time x. Generally d(x) is either zero or one, but we allow the possibility of tied survival times in which case d(x) may be greater than one.let n(x) PAGE 23

14 denote the number of individuals at risk just prior to time x; i.e., number of individuals in the sample who neither died nor were censored prior to time x. Then Kaplan-Meier estimate can be expressed as KM(t) = ( 1 d(x) ). x t n(x) Note: In the notation above, the product changes only at times x where d(x) 1;, i.e., only at times where we observed deaths. Non-informative Censoring In order that the life-table estimates give unbiased results there is an important assumption that individuals who are censored are at the same risk of subsequent failure as those who are still alive and uncensored. The risk set at any time point (the individuals still alive and uncensored) should be representative of the entire population alive at the same time. If this is the case, the censoring process is called non-informative. Statistically, if the censoring process is independent of the survival time, then we will automatically have non-informative censoring. Actually, we almost always mean independent censoring by non-informative censoring. If censoring only occurs because of staggered entry, then the assumption of non-informative censoring seems plausible. However, when censoring results from loss to follow-up or death from a competing risk, then this assumption is more suspect. If at all possible censoring from these later situations should be kept to a minimum. Greenwood s Formula for the Variance of the Life-table Estimator The derivation given below is heuristic in nature but will try to capture some of the salient feature of the more rigorous treatments given in the theoretical literature on survival analysis. For this reason, we will use some of the notation that is associated with the counting process approach to survival analysis. In fact we have seen it when we discussed the life-table estimator. PAGE 24

15 It is useful when considering the product limit estimator to partition time into many small intervals, say, with interval length equal to x where x is small. Figure 2.4: Partition of time axis Patient time x Let x denote some arbitrary time point on the grid above and define = number of individuals at risk (i.e., alive and uncensored) at time point x. = number of observed deaths occurring in [x, x + x). Recall: Previously, was denoted by n(x) and was denoted by d(x). It should be straightforward to see that w(x), the number of censored individuals in [x, x+ x), is equal to {[ Y (x + x)] }. Note: In theory, we should be able to choose x small enough so that { > 0and w(x) > 0} should never occur. In practice, however, data may not be collected in that fashion, in which case, approximations such as those given with life-table estimators may be necessary. With these definitions, the Kaplan-Meier estimator can be written as { KM(t) = 1 }, as x 0, all grid points x such that x + x t which can be modified if x is not chosen small enough to be { } LT (t) = 1, w(x)/2 all grid points x such that x + x t where LT (t) means life-table estimator. If the sample size is large and x is small, then is a small number (i.e., close to zero) and as long as x is not close to the right hand tail of the survival distribution (where may PAGE 25

16 be very small). If this is the case, then exp { } { 1 Here we used the approximation e x 1+x when x is close to zero. This approximation is exact when =0. }. Therefore, the Kaplan-Meier estimator can be approximated by { KM(t) exp } { =exp all grid points x such that x + x t here and thereafter, {x <t} means {all grid points x such that x + x t}. }, If x is taken to be small enough so that all distinct times (either death times or withdrawal times) are represented at most once in any time interval, then the estimator will be uniquely defined and will not be altered by choosing a finer partition for the grid of time points. In such a case the quantity is sometimes represented as t Basically, this estimator take the sum over all the distinct death times before time t of the number of deaths divided by the number at risk at each of those distinct death times. 2. The estimator hazard function Λ(t) = t 0 λ(x)dx. Thatis Recall that S(t) = exp( Λ(t)). is referred to as the Nelson-Aalen estimator for the cumulative Λ(t) =. By the definition of an integral, Λ(t) = t 0 λ(x)dx λ(x) x. grid points x such that x + x t PAGE 26

17 By the definition of a hazard function, λ(x) x P [x T<x+ x T x]. With independent censoring, it would seem reasonable to estimate λ(x) x, i.e., the conditional probability of dying in [x, x + x) given being alive at time x by. Therefore we obtain the Nelson-Aalen estimator Λ(t) =. We will now show how to estimate the variance of the Nelson-Aalen estimator and then show how this will be used to estimate the variance of the Kaplan-Meier estimator. time x. For a grid point x, leth(x) denote the history of all deaths and censoring occurring up to H(x) ={dn(u),w(u); for all values u on our grid of points for u<x}. Note the following 1. Conditional on H(x), we would know the value of (i.e., the number of risk at time x) and that would follow a binomial distribution denoted as H(x) Bin(,π(x)), where π(x) is the Conditional probability of an individual dying in [x, x + x) giventhat the individual was at risk at time x (i.e., π(x) =P [x T<x+ x T x]). Recall that this probability can be approximated by π(x) λ(x) x. 2. The following are standard results for a binomially distributed random variable. (a) E[ H(x)] = π(x), PAGE 27

18 (b) Var[ H(x)] = π(x)[1 π(x)], [ ] (c) E H(x) = π(x), {[ ][ ][ ] } (d) E 1 H(x) = π(x)[1 π(x)]. Consider the Nelson-Aalen estimator Λ(t) =.Wehave [ ] E[ Λ(t)] = E = [ ] E = [ [ ]] E E H(x) = π(x) λ(x) x t 0 λ(x)dx =Λ(t). Hence E[ Λ(t)] = π(x). If we take x smaller and smaller, then in the limit π(x) goestoλ(t). Namely Λ(t) is nearly unbiased to Λ(t). How to Estimate the Variance of Λ(t) The definition of variance is given by Var( Λ(t)) = E[ Λ(t) E( Λ(t))] 2 [ = E ] 2 π(x) [ { }] 2 = E π(x). Note: The square of a sum of terms is equal to the sum of the squares plus the sum of all cross product terms. So the above expectation is equal to E { } 2 π(x) + { }{ dn(x } π(x) ) Y (x x x <t ) π(x ) = [ ] 2 E π(x) + [{ }{ dn(x }] π(x) ) Y (x ) π(x ) E x x <t PAGE 28

19 We will first demonstrate that the cross product terms have expectation equal to zero. Let us take one such term and let us say, without loss of generality, that x<x. [{ }{ dn(x }] E π(x) ) Y (x ) π(x ) [ [{ }{ dn(x } ]] = E E π(x) ) Y (x ) π(x ) H(x ) Note: Conditional on H(x ),,Y(x) andπ(x) are constant since x<x. Therefore the above expectation is equal to [{ } [{ dn(x } ]] E π(x) ) E Y (x ) π(x ) H(x ) The inner conditional expectation is zero since { dn(x } ) E Y (x ) H(x ) = π(x ) by (2.c). Therefore we show that [{ }{ dn(x }] E π(x) ) Y (x ) π(x ) =0. Since the cross product terms have expectation equal to zero, this implies that Var( Λ(t)) = [ ] 2 E π(x) Using the double expectation again, we get that [ ] 2 E π(x) } 2 = E E π(x) H(x) [ [ ]] = E Var H(x) [ ] π(x)[1 π(x)] = E. Therefore, we have that Var( Λ(t)) = [ ] π(x)[1 π(x)] E. PAGE 29

20 If we wanted to estimate π(x)[1 π(x)], then using (2.d) we might think that [ ] 1 may be reasonable. In fact, we would then use as an estimate for Var( Λ(t)) the following estimator; summing the above estimator over all grid points x such that x + x t. Var( Λ(t)) = [ 1 ]. In fact, the above variance estimator is unbiased for Var( Λ(t)), which can be seen using the following argument: [ ] E 1 [ ] = E 1 { } = E E 1 H(x) = [ ] π(x)[1 π(x)] E (by (2.d)) = Var[ Λ(t)]. (double expectation again) What this last argument shows is that an unbiased estimator for Var[ Λ(t)] is given by [ 1 ]. Note: If the survival data are continuous (i.e., noties)and x is taken small enough, then would take on the values 0 or 1 only. In this case and [ ] 1 Var( Λ(t)) = = Y 2 (x), Y 2 (x), PAGE 30

21 which is also written as t 0 Y 2 (x). Remark: We proved that the Nelson-Aalen estimator π(x). We argued before that in the limit as x goes to zero, is an unbiased estimator for becomes t 0. We also argued that π(x) λ(x) x, hence as x goes to zero, then π(x) goes to t 0 λ(x)dx. These two arguments taken together imply that t 0 is an unbiased estimator of the cumulative hazard function Λ(t) = t 0 λ(x)dx, namely, E [ t 0 ] =Λ(t). Since Λ(t) = is made up of a sum of random variables that are conditionally uncorrelated, they have a martingale structure for which there exists a body of theory that enables us to show that PAGE 31

22 Λ(t) is asymptotically normal with mean Λ(t) and variance Var[ Λ(t)], which can be estimated unbiasedly by Var( Λ(t)) = [ 1 ] ; and in the case of no ties, by Var( Λ(t)) = Y 2 (x). Let us refer to the estimated standard error of Λ(t) by se[ Λ(t)] = [ 1 ] 1/2. The unbiasedness and asymptotic normality of Λ(t) aboutλ(t) allow us to form confidence intervals for Λ(t) (at time t). Specifically, the (1 α)th confidence interval for Λ(t) is given by Λ(t) ± z α/2 se( Λ(t)), where z α/2 is the (1 α/2)th quantile of a standard normal distribution. That is, the random interval [ Λ(t) z α/2 se( Λ(t)), Λ(t)+z α/2 se( Λ(t))] covers the true value Λ(t) with probability 1 α. This result could also be used to construct confidence intervals for the survival function S(t). This is seen by realizing that S(t) =e Λ(t), in which case the confidence interval is given by [e Λ(t) zα/2 se( Λ(t)), e Λ(t)+zα/2 se( Λ(t)) ], PAGE 32

23 meaning that this random interval will cover the true value S(t) with probability 1 α. An example: We will use the hypothetical data shown in Figure 2.3 to illustrate the calculation of Λ(t), Var Λ(t), and confidence intervals for Λ(t) ands(t). For illustration, let us take t = 17. Note that there are no ties in this example. So Λ(t) = t = 0 = =0.804, Var[ Λ(t)] = t Y 2 (x) = 0 Y 2 (x) = =0.145, 2 ŝe[ Λ(t)] = = So the 95% confidence interval for Λ(t) is ± = [0.0572, 1.551]. and the Nelson-Aalen estimate of S(t) is Ŝ(t) =e Λ(t) =e = The 95% confidence interval for S(t) is [e 1.551, e ]=[0.212, 0.944]. Note The above Nelson-Aalen estimate Ŝ(t) =0.448 is different from (but close to) the Kaplan-Meier estimate KM(t) = It should also be noted that above confidence interval for the survival probability S(t) is not symmetric about the estimator Ŝ(t). Another way of getting approximate confidence intervals for S(t) =e Λ(t) is by using the delta method. This method guarantees symmetric confidence intervals. Hence a (1 α)th confidence interval for f(θ) isgivenby f( θ) ± z α/2 f ( θ) σ. In our case, Λ(t) takes on the role of θ, Λ(t) takes on the role of θ, f(θ) =e θ so that S(t) =f{λ(t)}. Since f (θ) = e θ =e θ, and Ŝ(t) =e Λ(t). PAGE 33

24 Consequently, using the delta method we get Ŝ(t) a N(S(t), [S(t)] 2 Var[ Λ(t)]), and a (1 α)th confidence interval for S(t) is given by Ŝ(t) ± z α/2 {Ŝ(t) se[ Λ(t)]}. Remark: Note that [S(t)] 2 Var[ Λ(t)] is an estimate of Var[Ŝ(t)], where Ŝ(t) =exp[ Λ(t)]. Previously, we showed that the Kaplan-Meier estimator KM(t) = was well approximated by Ŝ(t) =exp[ Λ(t)]. [ 1 ] Thus a reasonable estimator of Var(KM(t)) would be to use the estimator of Var[exp( Λ(t))], or (by using the delta method) [Ŝ(t)]2 Var[ Λ(t))] = [Ŝ(t)]2 Y 2 (x). This is very close (asymptotically the same) as the estimator for the variance of the Kaplan- Meier estimator given by Greenwood. Namely Var{KM(t)} = {KM(t)} 2 [ ]. [ w(x)/2][ w(x)/2] Note: SAS uses the above formula to calculate the estimated variance for the life-table estimate of the survival function, by replacing KM(t) on both sides by LT (t). Note: The summation in the above equation can be viewed as the variance estimate for the cumulative hazard estimator defined by Λ KM (t) = log[km(t)]. Namely, Var{ Λ KM (t)} = [ w(x)/2][ w(x)/2]. In the example shown in Figure 2.3, using the delta-method approximation for getting a confidence interval with the Nelson-Aalen estimator, we get that a 95% CI for S(t) (where t=17) PAGE 34

25 is e Λ(t) ± 1.96 e Λ(t) se[ Λ(t)] = e ± 1.96 e = [0.114, 0.784]. The estimated se[ŝ(t)] = If we use the Kaplan-Meier estimator, together with Greenwood s formula for estimating the variance, to construct a 95% confidence interval for S(t), we would get [ KM(t) = 1 1 ][ 1 1 ][ 1 1 ][ 1 1 ][ 1 1 ] = { 1 Var[KM(t)] = } = ŝe[km(t)] = = Var[ Λ KM (t)] = =0.182 se[ Λ KM (t)] = Thus a 95% confidence interval for S(t) is given by KM(t) ± 1.96 ŝe[km(t)] = ± = [0.068, 0.754], which is close to the confidence interval using delta method, considering the sample size is only 10. In fact the estimated standard errors for Ŝ(t) andkm(t) using delta method and Greenwood s formula are and respectively, which agree with each other very well. Note: If we want to use R function survfit() to construct a confidence interval for S(t)with the form KM(t) ± z α/2 ŝe[km(t)],wehavetospecifytheargumentconf.type=c("plain") in survfit(). The default constructs the confidence interval for S(t) by exponentiating the confidence interval for the cumulative hazard using the Kaplan-Meier estimator. For example, a 95% CI for S(t) iskm(t) [e 1.96 se[ ΛKM (t)],e 1.96 se[ ΛKM (t)] ]=0.411 [e , [e ]= [0.178, 0.949]. Comparison of confidence intervals for S(t) 1. exponentiating the 95% CI for cumulative hazard using Nelson-Aalen estimator: [0.212, 0.944]. PAGE 35

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