# Comparing the log-rank test and RMST under nonproportional hazards

This vignette uses the simulation trio (`simdata_fast`, `analysis_fast`, and `simsummary_fast`) to compare the power of the log-rank test and the restricted mean survival time (RMST) at a single fixed analysis, under three survival patterns: proportional hazards, a delayed treatment effect, and crossing hazards. Along the way it shows the two plotting layers of the package, the design-stage scenario plot from `gen_scenario_fast` and the analysis-stage Kaplan-Meier plot from `kmcurve_fast`.

The log-rank test is the most efficient test when hazards are proportional, but its power can fall under nonproportional hazards because it weights all event times equally. RMST contrasts the area under the survival curves up to a truncation time and summarizes a difference in survival rather than a hazard ratio, so it behaves differently when the treatment effect is concentrated late or reverses over time. The three scenarios below are chosen to make those differences visible at a common sample size.

## Design and sample size

The control group has a median survival of 12 months. The sample size is set so that the proportional-hazards scenario, with a hazard ratio of 0.75, has 90% power at a one-sided 0.025 level. The events are obtained from the Schoenfeld formula and inflated to a sample size with the Lachin-Foulkes method through `gsDesign::nSurv`, given 12 months of accrual, 36 months of minimum follow-up, and a 5% annual dropout. The same sample size is then applied unchanged to all three scenarios.

    m0             <- 12
    lam0           <- log(2) / m0
    hr_ph          <- 0.75
    hr_late        <- 0.60
    hr_early_cross <- 1.40
    delay          <- 6
    alpha          <- 0.025
    power_target   <- 0.90
    accrual        <- 12
    minfup         <- 36
    study_dur      <- accrual + minfup
    dropout_annual <- 0.05
    eta            <- -log(1 - dropout_annual) / 12
    tau            <- 30

    ns <- gsDesign::nSurv(
      lambdaC = lam0, hr = hr_ph, eta = eta,
      T = study_dur, minfup = minfup,
      alpha = alpha, beta = 1 - power_target, sided = 1, ratio = 1
    )
    n_per   <- ceiling(ns$n / 2)
    n       <- c(n_per, n_per)
    n_total <- sum(n)
    a_rate  <- n_total / accrual

    data.frame(
      "Total n"       = n_total,
      "Per group"     = n_per,
      "Target events" = round(ns$d),
      check.names = FALSE
    )
    #>   Total n Per group Target events
    #> 1     616       308           506

## Scenarios

The control group is exponential throughout. The treatment group differs by scenario. Under proportional hazards the treatment hazard is 0.75 times the control hazard at all times. Under the delayed effect the two groups share the same hazard for the first 6 months and the treatment hazard drops to 0.60 times the control thereafter. Under crossing hazards the treatment group has a higher hazard for the first 6 months and a lower hazard afterwards, so the survival curves cross. The design-stage scenario plot shows the assumed survival curves and the piecewise hazard ratio for each scenario.

    scn <- gen_scenario_fast(
      scenarios = list(
        "PH" = list(
          e.hazard = list(lam0, hr_ph * lam0)
        ),
        "Delayed effect" = list(
          e.hazard = list(lam0, c(lam0, hr_late * lam0)),
          e.time   = c(0, delay, Inf)
        ),
        "Crossing" = list(
          e.hazard = list(lam0, c(hr_early_cross * lam0, hr_late * lam0)),
          e.time   = c(0, delay, Inf)
        )
      ),
      shared = list(n = n, a.time = c(0, accrual), a.rate = a_rate)
    )

    plot(scn, tmax = study_dur, mfrow = c(1, 3))

## A single simulated trial

Passing each scenario object to `simdata_fast` generates the data, here `nsim` replicates per scenario. The Kaplan-Meier plot below uses one replicate of the crossing scenario and adds the smoothed time-varying hazard ratio and the RMST shading up to the truncation time, which makes the early reversal of the effect visible in a single realized trial.

    scenarios <- scn$scenarios
    seeds     <- c(101, 102, 103)

    power_tab <- data.frame(
      Scenario = character(0), LogRank = numeric(0), RMST = numeric(0),
      stringsAsFactors = FALSE
    )
    examples <- vector("list", length(scenarios))
    names(examples) <- names(scenarios)

    for (i in seq_along(scenarios)) {
      s   <- scenarios[[i]]
      dat <- do.call(
        simdata_fast,
        c(s$args, list(nsim = nsim, d.hazard = eta, seed = seeds[i]))
      )
      res <- analysis_fast(
        dat, control = 1,
        time.looks = study_dur,
        stat = c("logrank", "rmst"),
        tau = tau, side = 1
      )
      s_lr   <- simsummary_fast(res, p.col = "logrank.p", alpha = alpha)
      s_rmst <- simsummary_fast(res, p.col = "rmst.p",    alpha = alpha)
      power_tab <- rbind(power_tab, data.frame(
        Scenario = s$label,
        LogRank  = s_lr[s_lr$look   == "overall", "cum.reject"],
        RMST     = s_rmst[s_rmst$look == "overall", "cum.reject"],
        stringsAsFactors = FALSE
      ))
      examples[[i]] <- dat[dat$sim == 1L, c("tte", "event", "group")]
    }

    ex  <- examples[["Crossing"]]
    fit <- kmcurve_fast(ex$tte, ex$event, ex$group, control = 1)
    plot(fit, hr = TRUE, rmst = TRUE, tau = tau, bw = 3)

## Power comparison

The table reports the simulated power, the proportion of the 10000 replicates in which each test rejects at the one-sided 0.025 level, for each scenario.

    knitr::kable(
      power_tab, digits = 3,
      col.names = c("Scenario", "Log-rank", "RMST"),
      caption = "Simulated power at the fixed analysis (one-sided 0.025)."
    )

Simulated power at the fixed analysis (one-sided 0.025).
Scenario | Log-rank | RMST
--- | --- | ---
PH | 0.901 | 0.839
Delayed effect | 0.952 | 0.748
Crossing | 0.435 | 0.086

Under proportional hazards the log-rank test reaches its design power and RMST is somewhat lower, as expected when the hazard ratio is constant and the log-rank test is the efficient choice. At this common sample size and this truncation time the unweighted log-rank test is at least as powerful as RMST in all three scenarios, and the gap widens under nonproportional hazards. Under the delayed effect the post-delay hazard ratio is stronger than the design value, so the log-rank test is in fact more powerful here than under proportional hazards; RMST detects the same late separation but averages it over a window that also spans the early no-difference period, so its power is markedly lower. Under crossing hazards both tests lose power because the early reversal offsets the late benefit, and RMST is affected most: the net survival difference over the window is small, so its power drops to around one tenth while the log-rank test retains moderate power. The single-trial plot above shows the mechanism, with the smoothed hazard ratio crossing one within the follow-up window, the situation in which a single hazard-ratio summary is least informative. These rankings are specific to the chosen truncation time, which the next paragraph takes up.

The truncation time for RMST is set here to 30 months, within the minimum follow-up of 36 months, so that every subject contributes over the window without extrapolation. The truncation time has a strong effect on the RMST result: a window that ends before the treatment benefit has accumulated leaves RMST little to detect, so it must be long enough to span the separation, and it should be prespecified on clinical grounds rather than tuned to the data.

## References

Lachin, J. M., & Foulkes, M. A. (1986). Evaluation of sample size and power for analyses of survival with allowance for nonuniform patient entry, losses to follow-up, noncompliance, and stratification. Biometrics, 42(3), 507-519.

Royston, P., & Parmar, M. K. B. (2013). Restricted mean survival time: an alternative to the hazard ratio for the design and analysis of randomized trials with a time-to-event outcome. BMC Medical Research Methodology, 13, 152.

Schoenfeld, D. A. (1983). Sample-size formula for the proportional-hazards regression model. Biometrics, 39(2), 499-503.

Uno, H., Claggett, B., Tian, L., et al. (2014). Moving beyond the hazard ratio in quantifying the between-group difference in survival analysis. Journal of Clinical Oncology, 32(22), 2380-2385.