This article/vignette provides a summary of functions in the ** gsDesign** package supporting design and evaluation of trial designs for time-to-event outcomes. We do not focus on detailed output options, but what numbers summarizing the design are based on. If you are not looking for this level of detail and just want to see how to design a fixed or group sequential design for a time-to-event endpoint, see the vignette

The following functions support use of the very straightforward Schoenfeld (1981) approximation for 2-arm trials:

`nEvents()`

: number of events to achieve power or power given number of events with no interim analysis.`zn2hr()`

,`gsHR()`

and`gsBoundSummary()`

: approximate the observed hazard ratio (HR) required to achieve a targeted Z-value for a given number of events.`hrn2z()`

: approximate Z-value corresponding to a specified HR and event count.`hrz2n()`

: approximate event count corresponding to a specified HR and Z-value.

The above functions do not directly support sample size calculations. This is done with the Lachin and Foulkes (1986) method. Functions include:

`nSurvival()`

: Sample size restricted to single enrollment rate; single analysis.`nSurv()`

: More flexible enrollment scenarios; single analysis.`gsSurv()`

: Group sequential design extension of`nSurv()`

.

Output for survival design information is supported in various formats:

`gsBoundSummary()`

: Tabular summary of a design in a data frame.`plot.gsDesign()`

: Various plot summaries of a design.`gsHR()`

: Approximate HR required to cross a bound.

We will assume a hazard ratio \(\nu < 1\) represents a benefit of experimental treatment over control. We let \(\delta = \log\nu\) denote the so-called *natural parameter* for this case. Asymptotically the distribution of the Cox model estimate \(\hat{\delta}\) under the proportional hazards assumption is \[\hat\delta\sim \hbox{Normal}(\delta=\log\nu, (1+r)^2/nr).\] Using a Cox model to estimate \(\delta\), the Wald test for \(\hbox{H_0}: \delta=0\) can be approximated with the asymptotic variance from above as:

\[Z_W\approx \frac{\sqrt {nr}}{1+r}\hat\delta=\frac{\ln(\hat\nu)\sqrt{nr}}{1+r}.\]

Also, we know that the Wald test \(Z_W\) and a standard normal version of the logrank \(Z\) are both asymptotically efficient and therefore asymptotically equivalent. We denote the *standardized effect size* as

\[\theta = \delta\sqrt r / (1+r)= \log(\nu)\sqrt r / (1+r).\] Letting \(\hat\theta = -\sqrt r/(1+r)\hat\delta\) we have \[ \hat \theta \sim \hbox{Normal}(\theta, 1/ n).\] Thus, the standardized Z version of the logrank is approximately distributed as

\[Z\sim\hbox{Normal}(\sqrt n\theta,1).\] Treatment effect favoring experimental treatment compared to control in this notation corresponds to a hazard ratio \(\nu < 1\), as well as negative values of the standardized effect \(\theta\), natural parameter \(\delta\) and standardized Z-test.

Based on the above, the power for the logrank test is approximated by

\[P[Z\le z]=\Phi(z -\sqrt n\theta)=\Phi(z- \sqrt{nr}/(1+r)\log\nu).\] Thus, assuming \(n=100\) events and \(\delta = \log\nu=-\log(.7)\), and \(r=1\) (equal randomization) we approximate power for the logrank test when \(\alpha=0.025\) as

```
<- 100
n <- .7
hr <- log(hr)
delta <- .025
alpha <- 1
r pnorm(qnorm(alpha) - sqrt(n*r)/(1+r)*delta)
#> [1] 0.4299155
```

We can compute this with `gsDesign::nEvents()`

as:

```
nEvents(n=n,alpha=alpha,hr=hr,r=r)
#> [1] 0.4299155
```

We solve for the number of events \(n\) to see how many events are required to obtain a desired power

\[1-\beta=P(Z\ge \Phi^{-1}(1-\alpha))\] with

\[n = \left(\frac{\Phi^{-1} (1-\alpha)+\Phi^{-1}(1-\beta)}{\theta}\right)^2 =\frac{(1+r)^2}{r(\log\nu)^2}\left({\Phi^{-1} (1-\alpha)+\Phi^{-1}(1-\beta)}\right)^2.\] Thus, the approximate number of events required to power for HR=0.7 with \(\alpha=0.025\) one-sided and power \(1-\beta=0.9\) is

```
<- 0.1
beta 1+r)^2/r/log(hr)^2 * ((qnorm(1 - alpha) + qnorm(1 - beta)))^2
(#> [1] 330.3779
```

which, rounding up, matches (with tabular output):

```
nEvents(hr = hr, alpha = alpha, beta = beta, r = 1, tbl = TRUE) %>%
kable()
```

hr | n | alpha | sided | beta | Power | delta | ratio | hr0 | se |
---|---|---|---|---|---|---|---|---|---|

0.7 | 331 | 0.025 | 1 | 0.1 | 0.9 | 0.1783375 | 1 | 1 | 0.1099299 |

The notation `delta`

in the above table changes the sign for the standardized treatment effect \(\theta\) in the above:

```
<- delta * sqrt(r) / (1 + r)
theta
theta#> [1] -0.1783375
```

The `se`

in the table is the estimated standard error for the log hazard ratio \(\delta=\log\hat\nu\)

```
1 + r) / sqrt(331 * r)
(#> [1] 0.1099299
```

We can create a group sequential design for the above problem either with \(\theta\) or with the fixed design sample size. The parameter `delta`

in `gsDesign()`

corresponds to standardized effect size with sign changed \(-\theta\) in notation used above and by Jennison and Turnbull (2000), while the natural parameter, \(\log(\hbox{HR})\) is in the parameter `delta1`

passed to `gsDesign()`

. The name of the effect size is specified in `deltaname`

and the parameter `logdelta = TRUE`

indicates that `delta`

input needs to be exponentiated to obtain HR in the output below. This example code can be useful in practice. We begin by passing the number of events for a fixed design in the parameter `n.fix`

(continuous, not rounded) to adapt to a group sequential design.

```
<- gsDesign(k=2,
Schoenfeld n.fix = nEvents(hr = hr, alpha = alpha, beta = beta, r = 1),
delta1 = log(hr))
%>%
Schoenfeld gsBoundSummary(deltaname = "HR", logdelta = TRUE) %>%
kable(row.names = FALSE)
```

Analysis | Value | Efficacy | Futility |
---|---|---|---|

IA 1: 50% | Z | 2.7500 | 0.4122 |

N: 173 | p (1-sided) | 0.0030 | 0.3401 |

~HR at bound | 0.6577 | 0.9391 | |

P(Cross) if HR=1 | 0.0030 | 0.6599 | |

P(Cross) if HR=0.7 | 0.3412 | 0.0269 | |

Final | Z | 1.9811 | 1.9811 |

N: 345 | p (1-sided) | 0.0238 | 0.0238 |

~HR at bound | 0.8078 | 0.8078 | |

P(Cross) if HR=1 | 0.0239 | 0.9761 | |

P(Cross) if HR=0.7 | 0.9000 | 0.1000 |

Exactly the same result can be obtained with the following, passing the standardized effect size `theta`

from above to the parameter `delta`

in `gsDesign()`

.

`<- gsDesign(k=2, delta = -theta, delta1 = log(hr)) Schoenfeld `

We noted above that the asymptotic variance for \(\hat\theta\) is \(1/n\) which corresponds to statistical information \(\mathcal I=n\) for the parameter \(\theta\). Thus, the value

```
$n.I
Schoenfeld#> [1] 172.2757 344.5514
```

corresponds both to the number of events and the statistical information for the standardized effect size \(\theta\) required to power the trial at the desired level. Note that if you plug in the natural parameter \(\delta= -\log\nu > 0\), then \(n.I\) returns the statistical information for the log hazard ratio.

```
gsDesign(k=2, delta = -log(hr))$n.I
#> [1] 43.06893 86.13786
```

The reader may wish to look above to derive the exact relationship between events and statistical information for \(\delta\).

Another application of the Schoenfeld (1981) method is to approximate boundary characteristics of a design. As noted in the introduction above, we will consider `zn2hr()`

, `gsHR()`

and `gsBoundSummary()`

to approximate the treatment effect required to cross design bounds. `zn2hr()`

is complemented by the functions `hrn2z()`

and `hrz2n()`

. We begin with the basic approximation used across all of these functions in this section and follow with a sub-section with example code to reproduce some of what is in the table above.

We return to the following equation from above:

\[Z\approx Z_W\approx \frac{\sqrt {nr}}{1+r}\hat\delta=\frac{\ln(\hat\nu)\sqrt{nr}}{1+r}.\] By fixing \(Z=z, n\) we can solve for \(\hat\nu\) from the above:

\[\hat{\nu} = \exp(z(1+r)/\sqrt{rn}).\] By fixing \(\hat\nu\) and \(z\), we can solve for the corresponding number of events required: \[ n = (z(1+r)/\log(\hat{\nu}))^2/r.\]

For our first example, we note that the event counts in `Schoenfeld`

are actually continuous numbers that are rounded up in the table:

```
$n.I
Schoenfeld#> [1] 172.2757 344.5514
```

We reproduce the approximate hazard ratios required to cross efficacy bounds using the Schoenfeld approximations above:

```
gsHR(z = Schoenfeld$upper$bound, # Z-values at bound
i = 1:2, # Analysis number
x = Schoenfeld, # Group sequential design from above
ratio = r) # Experimental/control randomization ratio
#> [1] 0.6576844 0.8077846
```

For the following examples, we assume \(r=1\).

`<- 1 r `

- Assuming a Cox model estimate \(\hat\nu\) and a corresponding event count, approximately what Z-value (p-value) does this correspond to? We use the first equation above:

```
<- .73 # observed hr
hr <- 125 # Events in analysis
events
<- log(hr) * sqrt(events*r) / (1+r)
z c(z, pnorm(z)) # Z- and p-value
#> [1] -1.75928655 0.03926443
```

We replicate the Z-value with

```
hrn2z(hr=hr, n = events, ratio = r)
#> [1] -1.759287
```

- Assuming an efficacy bound Z-value and event count, approximately what hazard ratio must be observed to cross the bound? We use the second equation above:

```
<- qnorm(.025)
z <- 120
events exp(z * (1 + r) / sqrt(r * events))
#> [1] 0.6991858
```

We can reproduce this with `zn2hr()`

by switching the sign of `z`

above; note that the default is `ratio = 1`

for all of these functions and often is not specified:

```
zn2hr(z = -z, n = events, ratio = r)
#> [1] 0.6991858
```

- Finally, if we want an observed hazard ratio \(\hat\nu = .8\) to represent a positive result, how many events would be need to observe to achieve a 1-sided p-value of 0.025? assuming 2:1 randomization? We use the third equation above:

```
<- 2
r <- .8
hr <- qnorm(.025)
z <- (z * (1 + r) / log(hr))^2 / r
events
events#> [1] 347.1683
```

This is replicated with

```
hrz2n(hr = hr, z = z, ratio = r)
#> [1] 347.1683
```

For the purpose of sample size and power for group sequential design, the Lachin and Foulkes (1986) is based on substantial evaluation not documented further here. We try to make clear here what some of the strengths and weaknesses of both the (**LachinFoules?**) method as well as its implementation in the `gsDesign::nSurv()`

(fixed design) and `gsDesign::gsSurv()`

(group sequential) functions. For historical and testing purposes, we also discuss use of the less flexible `gsDesign::nSurvival()`

function that was independently programmed and can be used for some limited validations of `gsDesign::nSurv()`

.

Some detail in specification comes With the flexibility allowed by the Lachin and Foulkes (1986) method. The model assumes

- A fixed enrollment period with piecewise constant enrollment rates
- A fixed minimum follow-up period
- Piecewise exponential failure rates for the control group
- A single, constant hazard ratio for the experimental group
- Piecewise exponential loss-to-follow-up rates
- A stratified population
- A fixed randomization ratio of experimental to control group assignment

Other than the proportional hazards assumption, this allows a great deal of flexibility in trial design assumptions. While Lachin and Foulkes (1986) adjusts the piecewise constant enrollment rates proportionately to derive a sample size, `gsDesign$nSurv()`

also enables the approach of Kim and Tsiatis (1990) which fixes enrollment rates and extends the final enrollment rate duration to power the trial; the minimum follow-up period is still assumed this approach. We do not enable the drop-in option proposed in Lachin and Foulkes (1986).

The two practical differences the Lachin and Foulkes (1986) method has from the Schoenfeld (1981) method are:

- By assuming enrollment, failure and dropout rates the method delivers sample size \(N\) as well as events required.
- The variance for the log hazard ratio \(\hat\delta\) is computed differently and both a null (\(\sigma^2_0) and alternate hypothesis (\)^2_1$) variance are incorporated through the formula \[ N = \left(\frac{\Phi^{-1}(1-\alpha)\sigma_0 + \Phi^{-1}(1-\beta)\sigma_1}{\delta}\right).\] The null hypothesis is derived by averaging the alternate hypothesis rates, weighting according to the proportion randomized in each group.

We will use the same hazard ratio 0.7 as for the Schoenfeld (1981) sample size calculations above. We assume further that the trial will enroll for a constant rate for 12 months, have a control group median of 8 months (exponential failure rate \(\lambda = \log(2)/8\)), a dropout rate of 0.001 per month, and 16 months of minimum follow-up. As before, we assume a randomization ratio \(r=1\), one-sided Type I error \(\alpha=0.025\), 90% power which is equivalent to Type II error \(\beta=0.1\).

```
<- 1 # Experimental/control randomization ratio
r <- 0.025 # 1-sided Type I error
alpha <- 0.1 # Type II error (1 - power)
beta <- 0.7 # Hazard ratio (experimental / control)
hr <- 8
controlMedian <- 0.001 # Exponential dropout rate per time unit
dropoutRate <- 12
enrollDuration <- 16 # minimum follow-up
minfup <- nSurv(lambdaC = log(2) / controlMedian,
Nlf hr = hr,
eta = dropoutRate,
T = enrollDuration + minfup, # Trial duration
minfup = minfup,
ratio = r,
alpha = alpha,
beta = beta
)cat(paste("Sample size: ", ceiling(Nlf$n), "Events: ", ceiling(Nlf$d), "\n"))
#> Sample size: 422 Events: 330
```

Recall that the Schoenfeld (1981) method recommended 331 events. The two methods tend to yield very similar event count recommendations, but not the same. Other methods will also differ slightly; see Lachin and Foulkes (1986). Sample size recommendations can vary more between methods.

We can get the same result with the `nSurvival()`

routine since only a single enrollment, failure and dropout rate is proposed for this example.

```
<- log(2) / controlMedian
lambda1 nSurvival(
lambda1 = lambda1,
lambda2 = lambda1 * hr,
Ts = enrollDuration + minfup,
Tr = enrollDuration,
eta = dropoutRate,
ratio = r,
alpha = alpha,
beta = beta
)#> Fixed design, two-arm trial with time-to-event
#> outcome (Lachin and Foulkes, 1986).
#> Study duration (fixed): Ts=28
#> Accrual duration (fixed): Tr=12
#> Uniform accrual: entry="unif"
#> Control median: log(2)/lambda1=8
#> Experimental median: log(2)/lambda2=11.4
#> Censoring median: log(2)/eta=693.1
#> Control failure rate: lambda1=0.087
#> Experimental failure rate: lambda2=0.061
#> Censoring rate: eta=0.001
#> Power: 100*(1-beta)=90%
#> Type I error (1-sided): 100*alpha=2.5%
#> Equal randomization: ratio=1
#> Sample size based on hazard ratio=0.7 (type="rr")
#> Sample size (computed): n=422
#> Events required (computed): nEvents=330
```

```
<- 2 # Total number of analyses
k <- gsSurv(k=2,
lfgs lambdaC = log(2) / controlMedian,
hr = hr,
eta = dropoutRate,
T = enrollDuration + minfup, # Trial duration
minfup = minfup,
ratio = r,
alpha = alpha,
beta = beta
)%>% gsBoundSummary() %>% kable(row.names = FALSE) lfgs
```

Analysis | Value | Efficacy | Futility |
---|---|---|---|

IA 1: 50% | Z | 2.7500 | 0.4122 |

N: 440 | p (1-sided) | 0.0030 | 0.3401 |

Events: 172 | ~HR at bound | 0.6571 | 0.9390 |

Month: 13 | P(Cross) if HR=1 | 0.0030 | 0.6599 |

P(Cross) if HR=0.7 | 0.3412 | 0.0269 | |

Final | Z | 1.9811 | 1.9811 |

N: 440 | p (1-sided) | 0.0238 | 0.0238 |

Events: 344 | ~HR at bound | 0.8074 | 0.8074 |

Month: 28 | P(Cross) if HR=1 | 0.0239 | 0.9761 |

P(Cross) if HR=0.7 | 0.9000 | 0.1000 |

Although we did not use the Schoenfeld (1981) for sample size, it is still used for the approximate HR at bound calculation above:

```
<- lfgs$n.I
events <- lfgs$upper$bound
z zn2hr(z = z, n = events) # Schoenfeld approximation to HR
#> [1] 0.6571386 0.8074431
```

There are various plots available. The approximate hazard ratios required to cross bounds again use the Schoenfeld (1981) approximation. For a **ggplot2** version of this plot, use the default `base = FALSE`

.

`plot(lfgs, pl="hr", dgt=4, base=TRUE)`

The variance calculations for the Lachin and Foulkes method are mostly determined by expected event accrual under the null and alternate hypotheses. The null hypothesis characterized above is seemingly designed so that event accrual will be similar under each hypothesis. You can see the expected events accrued at each analysis under the alternate hypothesis with:

```
::tibble(Analysis = 1:2,
tibble`Control events` = lfgs$eDC,
`Experimental events` = lfgs$eDE) %>%
kable()
```

Analysis | Control events | Experimental events |
---|---|---|

1 | 96.82001 | 74.77511 |

2 | 184.06813 | 159.12213 |

It is worth noting that if events accrue at the same rate in both the null and alternate hypothesis, then the expected duration of time to achieve the targeted events would be shortened. Keep in mind that there can be many reasons events will accrue at a different rate than in the design plan.

The expected event accrual of events over time for a design can be computed as follows:

```
<- seq(0.025, enrollDuration + minfup, .025)
Month plot(c(0,Month),
c(0, sapply(Month, function(x){nEventsIA(tIA=x, x = lfgs)})),
type = 'l', xlab="Month", ylab="Expected events",
main="Expected event accrual over time")
```

On the other hand, if you want to know the expected time to accrue 25% of the final events and what the expected enrollment accrual is at that time, you compute using:

```
<- tEventsIA(x = lfgs, timing = 0.25)
b cat(paste(" Time: ", b$T,
"\n Expected enrollment:", b$eNC + b$eNE,
"\n Expected control events:", b$eDC,
"\n Expected experimental events:", b$eDE, "\n"))
#> Time: 8.88072854965729
#> Expected enrollment: 325.066468979827
#> Expected control events: 49.0303966242417
#> Expected experimental events: 36.7672374839294
```

For expected accrual of events without a design returned by `gsDesign::gsSurv()`

, see the help file for `gsDesign::eEvents()`

.

Jennison, Christopher, and Bruce W. Turnbull. 2000. *Group Sequential Methods with Applications to Clinical Trials*. Boca Raton, FL: Chapman; Hall/CRC.

Kim, Kyungmann, and Anastasios A. Tsiatis. 1990. “Study Duration for Clinical Trials with Survival Response and Early Stopping Rule.” *Biometrics* 46: 81–92.

Lachin, John M., and Mary A. Foulkes. 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: 507–19.

Schoenfeld, David. 1981. “The Asymptotic Properties of Nonparametric Tests for Comparing Survival Distributions.” *Biometrika* 68 (1): 316–19.