ADVANCED STATISTICAL ANALYSIS OF EPIDEMIOLOGICAL STUDIES. Cox s regression analysis Time dependent explanatory variables


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1 ADVANCED STATISTICAL ANALYSIS OF EPIDEMIOLOGICAL STUDIES Cox s regression analysis Time dependent explanatory variables Henrik Ravn Bandim Health Project, Statens Serum Institut 4 November / 53
2 Outline Survival Data Example: Malignant Melanoma Data The Cox Model Cox in SAS Choice of TimeScale Example: GuineaBissau Data Delayed entries Time dependent explanatory variables 2 / 53
3 d i=1 exp(βx i ) j R(t i ) exp(βx j) 3 / 53
4 Survival Data Time to death or other event of interest. One timescale including a welldefined starting time timeorigin: Time from start of randomized clinical trial to death. Time from first employment to pension. Time from filling of a tooth to filling falls out. What is special about survival data? Rightskewed. No problem. CENSORING: For some we will only know a lower bound of lifetime. 4 / 53
5 Simple data Individual Times (months) 5 / 53
6 Survival and hazard function Let T be the TIME to event of interest: S(t) = P(T > t) = probability of survival to time t after entry at time 0 λ(t) = incidence, rate, or hazard Relationship: S(t) = exp ( t ) λ(s)ds = exp( Λ(t)) 0 Λ(t) is called the integrated hazard function. 6 / 53
7 λ(t) = λ S(t) = e λt Hazard rate Survival Function Time (t) Time (t) Λ(t) = λt Integrated hazard Time (t) 7 / 53
8 KaplanMeier estimate of survival function Death times t 1,..., t d (ordered). Y (t i ) = # alive just before t i. Ŝ(t) = ( 1 1 ) Y (t t i t i ) Risk sets Individual Times (months) 8 / 53
9 Survival probability Kaplan Meier survival estimate Time (months) Number at risk / 53
10 Malignant Melanoma Data In the period a total of 205 patients had their tumor removed and were followed until At the end of 1977: 57 died of mgl. mel. (status=1) 134 were still alive. (status=2) 14 died of nonrelated mgl. mel. (status=3) competing risk Purpose: Study effect on survival of sex, age, thickness of tumor, ulceration, etc / 53
11 Malignant melanoma N time status sex age year thickness ulcer / 53
12 The Cox Model The Cox model assumes that the rate for the ith individual is λ i (t) = λ 0 (t) exp(β 1 X i1 + β 2 X i β p X ip ) where β 1, β 2,..., β p are regression parameters, X i1 is the covariate value for covariate 1 for individual i, etc. Finally, λ 0 (t) is the baseline hazard. Time t is the timescale of choice, e.g. age, time since randomization, or time since operation. As formulated here the only quantity on the righthand side of the equal sign that depends on time is the baseline hazard λ 0 (t). If all covariates (X s) are zero we get λ i (t) = λ 0 (t). The interpretation of the baseline hazard is thus the hazard of a individual that have all covariates equal to zero. 12 / 53
13 The Cox model λ i (t) = λ 0 (t) exp(β 1 X i1 + β 2 X i β p X ip ) can also be written on the logscale (natural log) log(λ i (t)) = log(λ 0 (t) exp(β 1 X i1 + β 2 X i β p X ip )) The Cox model assumes that = log(λ 0 (t)) + β 1 X i1 + β 2 X i β p X ip. the effects of covariates are additive and linear on the log rate scale, just like the poisson regression. the CORNER i.e. the baseline hazard is nonparametric and depends on time, and time is thus adjusted for. We now turn to the interpretation of the regression parameters β 1, β 2,..., β p. 13 / 53
14 One binary covariate To make things more simple we only study the effect of one single binary covariate, e.g. sex on the risk of dying { 0 if individual i is a female X i = 1 if individual i is a male The Cox model is λ i (t) = λ 0 (t) exp(βx i ). With X i defined as above we get { λ 0 (t) if individual i is a female λ i (t) = λ 0 (t) exp(β) if individual i is a male 14 / 53
15 Mortality Rate Ratio Hazard Ratio If λ i (t) = { λ 0 (t) λ 0 (t) exp(β) if individual i is a female if individual i is a male then we have that the RATE RATIO (RR) between males and females is RR = λ 0(t) exp(β) = exp(β). λ 0 (t) Importantly, the ratio is independent of time, i.e. we have PROPORTIONAL HAZARDS over time. The Cox model is also called the proportional hazards model. How to estimate β? And what about baseline hazard λ 0 (t)? 15 / 53
16 Likelihood Function The baseline hazard is regarded as a nuisance and is not in general estimated, but it is possible. Let t 1,..., t d be the ordered death times It can been shown, that all we need is to find the β that maximizes the following function called Cox s partial likelihood function d exp(βx i ) L(β) = j R(t i ) exp(βx j) i=1 where R(t i ) is the RISK SET at death time t i i.e. the set of individuals being at risk of dying (under observation) just before time t i. The resulting estimate β is called the MAXIMUM LIKELIHOOD ESTIMATE of β. 16 / 53
17 Likelihood Function a closer look Death times t 1,..., t d, numbering individuals with deaths first: i = 1, 2,..., d, d + 1,..., n. with times and covariates t 1, t 2,..., t d, t d+1,..., t n. X 1, X 2,..., X d, X d+1,..., X n. At each death time we have the RISK SET: individuals alive and at risk of dying just before the death time: R(t 1 ), R(t 2 ),..., R(t d ) 17 / 53
18 Risk sets Individual Times (months) 18 / 53
19 For the Cox model λ i (t) = λ 0 (t) exp(βx i ) we use the Cox likelihood function to estimate β: L(β) = = d exp(βx i ) j R(t i ) exp(βx j) i=1 exp(βx 1 ) j R(t 1 ) exp(βx j) exp(βx 2 ) j R(t 2 ) exp(βx j) exp(βx d ) j R(t d ) exp(βx j) We index individuals in the risk sets using the letter j. Writing j R(t 1 ) exp(βx j) means summing over the individuals in the risk set for death time t 1. If we here assume that no one was censored before the first death time all individuals are in the risk set R(t 1 ) and the sum is exp(βx 1 ) + exp(βx 2 ) + + exp(βx n ). 19 / 53
20 For example for the Cox model λ i (t) = λ 0 (t) exp(β sex) Sex: 1=male, 0=female. Likelihood function: exp(β) j R(t 1 ) exp(βx j) 1 j R(t 2 ) exp(βx j) exp(β) j R(t d ) exp(βx j). If we again assume that no one was censored before the first death time all individuals are in the risk set R(t 1 ) and the sum is exp(β) exp(β) = N M exp(β) + N F, where N M and N F number of males and females respectively in R(t 1 ). The risk sets also play a crucial role in nested casecontrol studies more on this later in the course. 20 / 53
21 So far the following assumptions have been made for the Cox model The baseline hazard is assumed nonparametric, i.e. assumed to vary freely. The effects of covariates are additive and linear on the log rate scale. The ratio of the hazard rate for two subjects are constant over time. In other words, there is no interaction between the covariates and the time variable. Let us look at the Melanoma data using SAS. 21 / 53
22 Kaplan Meier survival estimates, by sex Time (years) female male What is the estimate of the RR between males and females? 22 / 53
23 Cox in SAS In SAS, proc phreg and proc tphreg can be used for estimating in the Cox model. We will use proc tphreg as this procedure can handle categorical variables much easier than proc phreg. Using proc tphreg we define the variable sex to be categorical using the class statement. For the variable sex 1 is males and 0 is females. proc tphreg data=melanom; class sex; model time*status(2,3) = sex; run; Please note, that we have two censoring codes namely 2 and 3. NB: In SAS 9.2 proc phreg now handles class variables and proc tphreg is obsolete. 23 / 53
24 Part of output from proc tphreg: Analysis of Maximum Likelihood Estimates Parameter Standard Hazard Parameter DF Estimate Error ChiSquare Pr > ChiSq Ratio sex The column Parameter Estimate is β. For a class variable SAS will automatically choose the highest number (here 1) as the reference. Thus, the rate ratio or Hazard Ratio is females compared to males. There is no estimate statement in proc (t)phreg, but a similar socalled contrast statement exists. Instead we can use the ref option in the class statement. Note also the option risklimits in the model statement which calculates the confidence interval for the hazard ratio. 24 / 53
25 proc tphreg data=melanom; class sex(ref="0"); model time*status(2,3) = sex / risklimits; run;... Analysis of Maximum Likelihood Estimates Parameter Standard Hazard 95% Hazard Ratio Parameter DF Estimate Error ChiSquare Pr > ChiSq Ratio Confidence Limits sex / 53
26 Melanoma data, thickness of tumor given by variable gtyk 1 if <2mm gtyk = 2 if 25 mm 3 if >5 mm proc tphreg data=melanom; class gtyk; model time*status(2,3) = gtyk / risklimits; run; Type 3 Tests Wald Effect DF ChiSquare Pr > ChiSq gtyk <.0001 Analysis of Maximum Likelihood Estimates Parameter Standard Hazard 95% Hazard Ratio Parameter DF Estimate Error ChiSquare Pr > ChiSq Ratio Confidence Limits gtyk < gtyk / 53
27 Melanoma data, + age in years proc tphreg data=melanom; class gtyk sex; model time*status(2,3) = gtyk sex age / risklimits; run; Type 3 Tests Wald Effect DF ChiSquare Pr > ChiSq sex gtyk <.0001 age Analysis of Maximum Likelihood Estimates Parameter Standard Hazard 95% Hazard Ratio Parameter DF Estimate Error ChiSquare Pr > ChiSq Ratio Confidence Limits sex gtyk < gtyk age / 53
28 LR = = 28.0 χ 2 2 (2 degrees of freedom) 28 / 53 Likelihood Ratio Test. proc tphreg data=melanom; class gtyk sex; model time*status(2,3) = gtyk sex; run; Model Fit Statistics Without With Criterion Covariates Covariates 2 LOG L AIC SBC proc tphreg data=melanom; class sex; model time*status(2,3) = sex; run; Model Fit Statistics Without With Criterion Covariates Covariates 2 LOG L AIC SBC
29 SAS: pvalue from chisquare test data temp; chisquare=28; df=2; p=1probchi(chisquare,df); run; proc print data=temp; run; Obs chisquare df p / 53
30 Choice of TimeScale A study may be conducted over calendar time even though the natural timescale is time since treatment Melanoma study. Cohort studies are often conducted by recruiting a random sample of the population at the start of the study and then these subjects are followed for a number of years Framingham. A natural timescale may be age rather than time in study which most often is an artificial timescale constructed by the investigators. What would timeorigin be if age was chosen as timescale? 30 / 53
31 Vaccinations in GuineaBissau Rural GuineaBissau: 5274 children under 7 months of age visited two times at home, with an interval of six months. Information about vaccination (BCG, DTP, mealses vaccine) collected at each visit and at second visit death during followup is registered. Some children moved away during followup, i.e. censored or survived until next visit, also censored. Below are some of the variable names from the bissau data. fuptime dead bcg agem Followup time in days 0 = censored, 1 = dead 1 = Yes, 2 = No Age at first visit in months 31 / 53
32 Is the risk of dying associated with vaccination? Outcome Exposure Died Survived Total BCG vaccinated 125 (3.8%) not BCG vaccinated 97 (4.9%) Total 222 (4.2%) / 53
33 proc tphreg data=bissau; class bcg; model fuptime*dead(0)=bcg / rl ; run; Testing Global Null Hypothesis: BETA=0 Test ChiSquare DF Pr > ChiSq Likelihood Ratio Score Wald Type 3 Tests Wald Effect DF ChiSquare Pr > ChiSq bcg Analysis of Maximum Likelihood Estimates Parameter Standard Hazard 95% Hazard Ratio Parameter DF Estimate Error ChiSquare Pr > ChiSq Ratio Confidence Limits bcg / 53
34 proc tphreg data=bissau; class bcg agem; model fuptime*dead(0)=bcg agem / rl ; run; Type 3 Tests Wald Effect DF ChiSquare Pr > ChiSq bcg agem Analysis of Maximum Likelihood Estimates Parameter Standard Hazard 95% Hazard Ratio Parameter DF Estimate Error ChiSquare Pr > ChiSq Ratio Confidence Limits bcg agem agem agem agem agem agem / 53
35 Delayed entries Time in study Age as time Individual 7 6 Individual Times (months) Age (months) 35 / 53
36 Subjects are only at risk at age of entry and onwards. They are not at risk in our World of analysis before age of entry! Handling of delayed entries is easily done by careful control of the RISK SET R(t i ) at death time t i in the likelihood function: L(β) = d exp(βx i ) j R(t i ) exp(βx j) i=1 Only individuals at risk and under observation is included in the risk set R(t i ) at time t i. 36 / 53
37 Delayed entries in SAS data bissau2; set bissau; outage=age+fuptime; run; proc tphreg data=bissau2; class bcg; model (age,outage)*dead(0)= bcg / rl; run; Analysis of Maximum Likelihood Estimates Parameter Standard Hazard 95% Hazard Ratio Parameter DF Estimate Error ChiSquare Pr > ChiSq Ratio Confidence Limits bcg / 53
38 Time dependent explanatory variables The Cox model can be expanded to include timevarying covariates λ i (t) = λ 0 (t) exp(βx i (t)). The likelihood function for death times t 1,..., t d becomes L(β) = d i=1 exp(βx i (t i )) j R(t i ) exp(βx j(t i )). From this we can see that we just need to know the value of the covariates at the deaths times: X i (t 1 ), X i (t 2 ),..., X i (t d ). The covariate values at any time different from a death time is not used in the likelihood function. 38 / 53
39 The most simple timevarying covariate is a binary variable that is allowed to change once during followup, e.g. new BCG vaccinations registered between visits in the Bissau data: X i (t) = { 0 if no BCG before time t 1 if BCGtime t 39 / 53
40 A child being BCGvaccinated after 3 months of followup. BCG Follow up (months) The timevarying covariate is 0 in the time interval 0 to 3 months and 1 for the rest of followup. For a child who was BCG vaccinated before first visit the timevarying covariate is one during all the followup. 40 / 53
41 Multistate Model λ 01 (t) 0 1 Unexposed Exposed λ 02 (t) 2 Dead λ 12 (t) We want to compare λ 02 (t) and λ 12 (t). The transition λ 01 (t) is not modeled here. 41 / 53
42 Instead of time of followup we will use age as timescale to illustrate the use of BCG as a timevarying covariate in the Bissau data. At visit 2 the vaccination cards were seen for the children at home and an age of BCG vaccination (bcgage) was calculated: id fuptime dead age bcg bcgage outage / 53
43 Binary timevarying covariate in SAS (I) proc tphreg data=bcg; if.<bcgage<outage then bcg_t=1; else bcg_t=0; model (age,outage)*dead(0)=bcg_t / rl ; run; Analysis of Maximum Likelihood Estimates Parameter Standard Hazard 95% Hazard Ratio Parameter DF Estimate Error ChiSquare Pr > ChiSq Ratio Confidence Limits bcg_t < / 53
44 The ifstatement if.<bcgage<outage then bcg_t=1; else bcg_t=0; is recalculated at each death time. The outage in the model statement refers to the current death times being evaluated (i.e. a t i in the likelihood). For the first death time which is t 1 = 23 days of age, the ifstatement becomes if.<bcgage<23 then bcg_t=1; else bcg_t=0; being calculated for all children at risk at age 23 days (in R(t 1 = 23)) with their individual bcgagevalues. This is a recalculation of the timevarying covariate at each death time c.f. the likelihood function. 44 / 53
45 Binary timevarying covariate in SAS (II) Splitting up persons with a changing timevarying covariate in two records: age bcgage outage bcgvacc=0 status=0 bcgvacc=1 status=dead and use delayed entries. Thus, we need to generate a new data set. 45 / 53
46 data splitbcg; set bcg; if bcgage=. or bcgage>outage then do; bcgvacc=0; entryage=age; exitage=outage; status=dead; output; end; if.<bcgage<=age then do; bcgvacc=1; entryage=age; exitage=outage; status=dead; output; end; if age<bcgage<=outage then do; bcgvacc=0; entryage=age ; exitage=bcgage; status= 0; output; bcgvacc=1; entryage=bcgage; exitage=outage; status=dead; output; end; run; id fuptime dead age bcg bcgage outage bcgvacc entryage exitage status / 53
47 proc tphreg data=splitbcg; class bcgvacc(ref="0"); model (entryage,exitage)*status(0)=bcgvacc / rl ; run; Parameter Standard Hazard 95% Hazard Ratio Parameter DF Estimate Error ChiSquare Pr > ChiSq Ratio Confidence Limits bcgvacc < / 53
48 Other timevarying covariates Effect of binary X (0,1) changes at t 0 : where λ i (t) = λ 0 (t) exp(β 1 X i + β 2 X i I (t t 0 )), I (t t 0 ) = Can be handled by method I+II. { 1 if t t 0 0 if t < t 0 Effect of binary X (0,1) decreases or increases with time: λ i (t) = λ 0 (t) exp(β 1 X i + β 2 (X i t)). Can be handled by method I or by splitting at each failure or special options. 48 / 53
49 Stanford Heart Transplant Data (p. 235) In a report (Crowley and Hu, J Amer. Statist Assoc. 1977) on the Stanford Heart Transplantation Study, patients identified as been eligible (N=103) for a heart transplant were followed until death or censorship. In total 65 received transplant during followup, whereas 38 did not. Assess whether transplanted patients survive better. On the next slide you will find the variables in the transplant data set. Here we will discuss how to analyse and at the exercises we will do some of the analyses. 49 / 53
50 Stanford Heart Transplant Data variables age cens days trans wait mismatch age (in years) at entry into the study. 0 = Censoring 1 = Dead number of days from entry to dead/censoring. 1 = if the person had a heart transplantation 0 = otherwise. number of days from entry to transplantation NB: if trans = 0 then wait = 1 1 = mismatch between HLA type in donor and patient 0 = no mismatch NB: if trans = 0 then mismatch = / 53
51 Obs age cens days trans wait mismatch / 53
52 Piecewise Constant Hazard Rate = Poisson regression Divide the time scale into K pieces and assuming piecewise constant but different hazard rates in each of the intervals. This may provide a sensible summary of many phenomena and is often used in epidemiology. λ 1 λ 2 λ 3 λ K c 0 = 0 c 1 c 2 c 3 c K 1 c K Age Thus λ(t) = λ k for t (c k 1, c k ], k = 1,..., K The intervals do not need to be of same length. We only need to keep record of the total number of deaths and the exposure time in each group. 52 / 53
53 We can further divide each interval into categories of covariates, e.g. sex (F=females, M=males): λ 1F λ 2F λ 3F λ KF λ 1M λ 2M λ 3M λ KM c 0 = 0 c 1 c 2 c 3 c K 1 c K Age Not straight forward in SAS to split the timescale, but socalled userwritten SASmacros exist. See for example: Stata use stsplit command. R packages exist (e.g. Epi Package) SPSS? 53 / 53
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