 Research article
 Open access
 Published:
Including nonconcurrent control patients in the analysis of platform trials: is it worth it?
BMC Medical Research Methodology volumeÂ 20, ArticleÂ number:Â 165 (2020)
Abstract
Background
Platform trials allow adding new experimental treatments to an ongoing trial. This feature is attractive to practitioners due to improved efficiency. Nevertheless, the operating characteristics of a trial that adds arms have not been wellstudied. One controversy is whether just the concurrent control data (i.e. of patients who are recruited after a new arm is added) should be used in the analysis of the newly added treatment(s), or all control data (i.e. nonconcurrent and concurrent).
Methods
We investigate the benefits and drawbacks of using nonconcurrent control data within a twostage setting. We perform simulation studies to explore the impact of a linear and a step trend on the inference of the trial. We compare several analysis approaches when one includes all the control data or only concurrent control data in the analysis of the newly added treatment.
Results
When there is a positive trend and all the control data are used, the marginal power of rejecting the corresponding hypothesis and the type one error rate can be higher than the nominal value. A modelbased approach adjusting for a stage effect is equivalent to using concurrent control data; an adjustment with a linear term may not guarantee valid inference when there is a nonlinear trend.
Conclusions
If strict error rate control is required then nonconcurrent control data should not be used; otherwise it may be beneficial if the trend is sufficiently small. On the other hand, the root mean squared error of the estimated treatment effect can be improved through using nonconcurrent control data.
Background
The implementation of innovative trial designs for drug development programmes has increased in recent years [1]. A platform trial is one such innovative design that helps improve efficiency. It has a multiarm multistage design where a single control group is utilized to evaluate multiple experimental treatments for a single disease. Apart from the feature of dropping arms, a platform trial allows adding arms when new experimental treatments become available. This adds efficiency as adding arms takes less time and has a lower cost than setting up new trials from scratch.
A platform trial can be implemented through the specification of a master protocol [2â€“4]. Cohen et al [5] review the statistical aspects on the adding arm adaptation and present some trials that have adopted a platform approach. For pairwise comparisons between the effectiveness of newly added treatments and the control treatment, there are different views on utilising the control data [5]. To be more specific, the control data of a platform trial can be separated into nonconcurrent and concurrent. The former (latter) corresponds to the control data of the patients who are recruited before (after) a new experimental treatment is added. This definition applies to each newly added arm independently; the control data can be concurrent to a newly added treatment, and be nonconcurrent to another treatment that is added at a later time point.
To our knowledge, methodology and guidance for adding arms remain limited and the operating characteristics of platform trials have not been wellstudied. The literature on other trial adaptations such as dropping arms and changing treatment allocation ratio are more comprehensive [6â€“11]. In particular, many have explored the presence of time trends when more complex randomization procedures are recommended/ explored [12â€“21]. A trend may be present in a trial when there is a learning curve among the study personnel; a shift in the population baseline characteristics; and/ or the effect of the control treatment may change due to other reasons (e.g. improvement of the standard of care/ practice) [7,22].
In order to minimise the impact caused by the presence of a trend, most researchers advocate the use of concurrent control data when making inference about the newly added treatment(s) [23â€“25]. Others stipulate that nonconcurrent control data may be used when making inference about the newly added arm by adjusting for possible trends [26,27]. Here we investigate whether using nonconcurrent control data is worthwhile in platform trials. Several analysis approaches are considered. Within a twostage setting where a new treatment is added at the end of stage one [25,28], we explore the impact of i) the timing of adding a new arm ii) the sample size of the new arm and iii) the magnitude of a linear or a step trend [15] on the inference about the newly added experimental treatment. We provide recommendations about under which conditions nonconcurrent control data should be used.
Method
Consider a twostage multiarm trial that initially has K treatments and a control treatment with a total sample size of N=N_{1}+N_{2} where N_{s} is stage s sample size, s=1,2. Denote k=0 for a control treatment, k=1,...,K for the initial experimental treatments. Let n_{ks} be the sample size of treatment k in stage s with n_{k1}+n_{k2}=n_{k}. At the end of stage one, in particular after \(N_{1}=\sum _{k=0}^{K} n_{k1}\) patients have been randomized to the initial arms in stage one, a new treatment (denoted by K+1) is added to the ongoing trial with n_{K+1,2} patients. This increases the second stage sample size to N_{2}+n_{K+1,2} and the overall sample size of the trial to N+n_{K+1,2}.
Denote n_{01}/n_{0} as the timing of adding a new arm. Small n_{01}/n_{0} indicates the newly available treatment is added to the trial after a small number of patients have been randomized to the initial arms in stage one. We consider the case that all arms finish recruitment simultaneously (although as long as controls continue to be recruited while treatment K+1 is allocated patients, results apply more generally). For j=1,...,N+n_{K+1,2}, denoting the sequence of patient enrolment to the trial, let:
be the response of treatment k of subject j; Î¼_{k} be the true effect of treatment k; Ï„(j) be a trend that could be a function of the patient ordering j; Îµ_{j} be the random errors that are identically and independently normally distributed with mean 0 and variance Ïƒ^{2}; and T_{j} be the allocated treatment of subject j, with I(Â·) is an indicator function.
Let \(\bar {X}_{k.s}\) and \(\bar {X}_{k}\) denote the stage s sample mean response of treatment k and the overall mean response of treatment k. When there is no trend in the trial, i.e. Ï„(j)=0(j), the sample mean estimators \(\bar {X}_{k.s}\overset {iid}{\sim } N(\mu _{k}, \sigma ^{2}/n_{ks})\) and \(\bar {X}_{k}\overset {iid}{\sim } N(\mu _{k}, \sigma ^{2}/n_{k})\).
For illustration purpose, we consider the following when exploring the impact of a fixed trend on the inference:
linear trend:
$$\begin{array}{*{20}l} \tau(j)= \lambda \ (j 1) / (N+n_{K+1,2}1) \end{array} $$(2)stepwise trend:
$$\begin{array}{*{20}l} \tau(j)= \lambda \cdot c_{j} \end{array} $$(3)
where c_{j} is an indicator that subject j is enrolled in stage two, and Î» is a positive fixed value. Both of these trends inflate the value of the outcome variable in a sequential manner; the responses of the patients who were enrolled later are larger than those who were enrolled earlier.
Simple approach: Ztest
We first consider a simple method that is analogous to ignoring the possibility of a trend or stage effect: a standard Ztest for testing each hypothesis
H_{0k}:Î¼_{k}=Î¼_{0}  against  H_{1k}:Î¼_{k}>Î¼_{0}, 
using the sample means. When there is no trend, the (pairwise) marginal power of rejecting a null hypothesis to detect a difference in treatment effect, Î´_{k}=Î¼_{k}âˆ’Î¼_{0}, is
where Î¦ and Î¦^{âˆ’1} are the cumulative density function and the inverse of the cumulative density function of a standard normal distribution. For the newly added treatment, the power of rejecting H_{0,K+1} depends on whether all the control data (of size n_{0}) or only the stage two control data (of size n_{02}) is used in the inference. Elm et al [28] describe a ttest for the case where Ïƒ^{2} is unknown.
Modelbased approach: weighted linear regression
One way to account for the changes of a trial design is to consider a modelbased approach. Elm et al [28] consider the following weighted linear regression model. Let \(\bar {X}^{(k)}_{s}= \bar {X}_{k.s} \bar {X}_{0.s}\) be the stagewise difference in mean responses; fit a linear model to the samples of the stagewise mean differences:
where \(\eta ^{(k)}_{s}\) is normally distributed with mean zero and variance Ïƒ^{2}/n_{ks}+Ïƒ^{2}/n_{0s}. The covariance between mean responses, \(cov(\bar {X}^{(l)}_{1},\bar {X}^{(k)}_{2})=0 \forall l\) and k=1,...,K, as patient data are independent; whereas \(cov(\bar {X}^{(l)}_{s},\bar {X}^{(k)}_{s})= \sigma ^{2}/ n_{0s}, \forall l \neq k\), where l,k=1,...,K, as the data of a shared control group is used in all the pairwise comparisons. For the newly added treatment, \(cov(\bar {X}^{(K+1)}_{2}, \bar {X}^{(k)}_{1})= 0\) and \(cov(\bar {X}^{(K+1)}_{2}, \bar {X}^{(k)}_{2})= \sigma ^{2}/ n_{02}\) for k=1,...,K, when only the concurrent control data is used.
Elm et al [28] have not considered using all the control data in estimating the difference in mean responses of treatment K+1 and the control, i.e. defined as \(\bar {X}^{(K+1)'}_{2}= \bar {X}_{K+1,2} \bar {X}_{0}\). This estimate has a smaller variance than \(\bar {X}^{(K+1)}_{2}\), and \(cov(\bar {X}^{(K+1)'}_{2}, \bar {X}^{(k)}_{s})= \sigma ^{2}/ n_{0}\) for k=1,...,K, and s=1,2, as \(\bar {X}_{0}\) can be reexpressed as \(\frac {n_{01}}{ n_{0}} \bar {X}_{0.1}+ \frac {n_{02}}{ n_{0}} \bar {X}_{0.2}\).
Using weighted least squares estimation, an estimate of the vector of treatment effects relative to control, Î´=(Î´_{1},...,Î´_{K},Î´_{K+1})^{T}, can be obtained. The estimate, \(\boldsymbol {\hat {\delta }}\), is unbiased and has a multivariate normal distribution with mean Î´ when there is no trend. This joint distribution can be used to test the hypotheses, {H_{01},...,H_{0,K+1}}, in a similar way to the Dunnett test [29]. The marginal power of rejecting a hypothesis can be computed by considering the corresponding marginal distribution. Note that when K>0, the marginal power is different to the power obtained from a Ztest (or a ttest) as the joint distribution accounts for the fact that a common control group is used in the analysis. Note also that \(var(\hat {\delta }_{K+1})\) depends on the number of control patients.
We label these approaches as WLS_{all} and WLS_{s2} respectively, when all the control data and when only the concurrent control data are used in the estimation of Î´_{K+1}.
Modelbased approach: linear regression
In the context of an adaptive randomisation procedure, Villar et al. [19] considered fitting a linear regression model to adjust for the presence of a trend in their investigation where all arms start recruiting at the same time. We explore the following linear regression analyses for the trial that adds an arm at the end of stage one:
M_{a1}: \(X_{j }= \alpha + { \sum _{k=0}^{K+1} \beta _{k}\cdot I(T_{j}=k) }+ { \gamma } \cdot j + \epsilon _{j}\)
M_{a2}: \(X_{j }=\alpha + { \sum _{k=0}^{K+1} \beta _{k}\cdot I(T_{j}=k)} + { \nu } \cdot c_{j}+ \epsilon _{j}\)
where Î²_{0}=0 such that Î²_{k} represents the difference in mean responses of treatment k and the control treatment; and
M_{b1}: \(X_{j }=\alpha + {\beta _{0} \cdot I(T_{j}=0)} + \beta ^{b1}_{K+1}\cdot {I(T_{j}=K+1)} + { \gamma } \cdot j + \epsilon _{j}\)
M_{b2}: \(X_{j }=\alpha + {\beta _{0} \cdot I(T_{j}=0)} + \beta ^{b2}_{K+1}\cdot {I(T_{j}=K+1)} + { \nu } \cdot c_{j}+ \epsilon _{j}\)
where Î²_{0}=0 such that \(\beta ^{b1}_{K+1}\) and \(\beta ^{b2}_{K+1}\) represent the difference in mean responses of the newly added treatment and the control treatment. Note that models M_{a1} and M_{a2} use all data to estimates all Î²_{1},...,Î²_{K},Î²_{K+1}, whereas model M_{b1} (and M_{b2}) uses the data of the control arm (both stages) and the newly added arm to estimate \(\beta ^{b1}_{K+1}\) (and \(\beta ^{b2}_{K+1}\)) only. The parameter Î³ in M_{a1} and M_{b1} represents the effect of continuous recruitment, whereas Î½ in M_{a2} and M_{b2} represents the effect of recruitment that happens after the new treatment has been added (Elm et al [28] interpret Î½ as a stage effect). It is worth mentioning that the amount of the control data used to estimate Î²_{K+1} in M_{a2} (and respectively \(\beta ^{b2}_{K+1}\) in M_{b2}) is similar to (same as) that of using only the second stage control data, e.g. \(\bar {X}_{K+1,2}\bar {X}_{0.2}\), since the newly added treatment was not present in stage one and the stage one control data would contribute to the estimation of the stage effect and that of Î²_{1},...,Î²_{K} under such a model setup. In other words, the stage one control data contributes very little to the estimation of the difference in mean responses of the newly added treatment and the control treatment when we adjust for Î½Â·c_{j} in the analysis.
We note that the inference from M_{a1} without adjustment for the linear term in patient ordering is similar to that from WLS_{all}; and that from M_{a2} is similar to WLS_{s2}. This is because the joint distribution of \(\{\hat {\beta }_{1},..., \hat {\beta }_{K+1}\}\) is equivalent to the corresponding joint distribution of \(\boldsymbol {\hat {\delta }}\). The subtle difference is the former approach has one more parameter (i.e. the intercept Î±) than the weighted regression model, which reflects the mean response of the reference group assuming the model is true. On the other hand, testing \(\beta ^{b1}_{K+1}\) and \(\beta ^{b2}_{K+1}\) respectively are similar to considering an independent Ztest (or a ttest when assuming unknown variance) for the hypothesis of H_{0,K+1}.
Metrics
We explore the above analysis approaches and consider
bias of estimators;
the type one error rate;
the marginal power of rejecting a false null hypothesis;
the root mean squared error,
$${}\text{rMSE}=\sqrt{ \text{bias of an estimate}^{2}+ \text{variance of an estimate} };$$
when including or excluding the stage one control data in the analysis of the newly added treatment. As we assume Ïƒ^{2} is known, the presence of a trend would affect the accuracy of the estimate but not the variance of an estimate. We include the following metric to explore the reduction in the variance of the estimate:
where BoS stands for borrowing of strength [30], V_{0} and V_{a} correspond to the variance of an estimate that is computed with only the concurrent control data and that with all the control data respectively. Note that BoS increases as V_{a} decreases. This measure is similar to the Rsquared that measures how close the data are to the fitted regression model. Here BoSâ‰ˆ0 indicates that the variability of the response data around its mean cannot be further explained by including nonconcurrent control data; whereas BoSâ‰ˆ1 reflects that using all the control data results in a perfect estimate that has negligible variability. One may interpret BoS as an indicator of the benefit of including nonconcurrent control data in the analysis of the newly added treatments. Small values show that including the nonconcurrent control data provides little benefit to the estimation of the parameters in terms of precision gain. High value is desirable especially when the estimates of the variance are unbiased and are not affected by the presence of a trend.
In the next section, we consider K=1 and a new arm is added at the end of stage one. Let the effect size of the difference in treatment effects be Î´_{1}=Î´_{2}=0.15 under the alternative scenario, and Î´_{1}=Î´_{2}=0 under the null scenario, with Ïƒ^{2}=1,n_{11}=n_{01}=N_{1}/2, and n_{12}=n_{02}=N_{2}/2. We choose n_{0}=n_{1}=550 to obtain 80% power and 5% type one error rate of rejecting H_{01} using the Ztest.
We conduct simulation studies to examine the impact of a trend with the timing of adding a new treatment, n_{01}/n_{0}={0.25,0.5,0.75}, and the sample size of the new arm, n_{22}={n_{02},n_{0},2n_{0}}. Following (1), we adjust the generated responses post randomization with Î»={0.02,0.04,0.06,0.08} in the linear trend (2) and the stepwise trend (3) respectively. These values indicate a trend of Î»Ã—100% of standard deviation. Each scenario is replicated 100 000 times. We illustrate n_{22}=2n_{0} as one of the options to explore the marginal benefit of having more patients (than necessary) in the newly added arm while keeping the size of the control arm the same as the initial plan.
Results
Analytical power and BoS when there is no trend
We now compare BoS and the power of rejecting H_{02} when there is no trend and either all, or only concurrent control data are used in the inference.
We first demonstrate the differences between the independent Ztest (computed with only concurrent control data) and the modelbased approach, WLS_{all} and WLS_{s2}, in terms of power. For ease of presentation, we omit the comparisons to the presented linear regression models as they make adjustment for a trend (see the first row of plots in Figs. 2 and 3 for the corresponding marginal power). The left plot in Fig. 1 shows power curves when the individual Ztest (red lines) and WLS_{s2}, i.e. the approach with the bivariate normal distribution of \((\hat {\delta }_{1}, \hat {\delta }_{2})\) (blue lines), are used respectively for testing H_{02}, both using only the concurrent control responses. The timing of adding the new arm to the ongoing trial is reflected on the xaxis; small values indicate the arm is added after a small number of patients have been randomized to the initial treatment arms. Different lines correspond to having different n_{22}.
As expected, the power decreases with n_{01}/n_{0} when excluding stage one control responses in the inference about the newly added treatment. Comparing the red lines to the blue lines in the left plot of Fig. 1, the power of the individual Ztest is lower than the marginal power obtained from WLS_{s2} given the same n_{22}. When n_{22}=n_{02}, the difference between the power curves computed by the two approaches is less than 6% (shown by comparing the dashed lines). When n_{22}â‰¥n_{0}, the magnitudes of the differences between the power curves are larger especially for large n_{01}/n_{0}; e.g. compare the dotteddashed lines at n_{01}/n_{0}=0.8, the marginal power obtained from WLS_{s2} is 15% more than that computed from the individual Ztest. This finding highlights that considering the joint outcomes of the hypotheses through the joint distribution of the parameters is more efficient than using the individual Ztest when only the concurrent control data is used. This is because the former accounts for using a shared control group in the inference, whereas the Ztest uses the control data independently for each hypothesis test as if they were from several trials. Besides that, compare the (vertical) difference between the dotteddashed line (n_{22}=2n_{0}) and the dotted line (n_{22}=n_{0}), either red or blue, we see that the magnitude of the difference in the power decreases as n_{01}/n_{0} increases. This indicates that the marginal benefit of having a larger sample size in one arm than necessary can be small, depending on when the new arm is added to the trial.
For WLS_{all} with n_{22}=n_{02}, the power curve coincides with the red dotted line, i.e. the individual Ztest using only the concurrent control data and with n_{22}=n_{0}. This is not surprising as the same number of patients are used in the inference, albeit the number of patients randomized to the newly added arm is different for the two comparators (i.e. WLS_{all} considers n_{02} responses of the new treatment and n_{0} responses of the control treatment; the individual Ztest uses n_{0} responses of the new treatment and n_{02} responses of the stage two control treatment). For WLS_{all} with n_{22}=n_{0} and n_{22}=2n_{0}, the marginal power of rejecting H_{02} is 0.8 and 0.89 respectively across all n_{01}/n_{0}, which is close to the respective power obtained from Ztest using all the control data.
We now consider the BoS values that are computed with \(var(\hat {\delta }_{2})\) obtained from WLS_{all} and WLS_{s2} respectively. The right plot in Fig. 1 shows that when all the control data is included in the estimation, BoS increases with n_{01}/n_{0}. When n_{22}=n_{02}, BoS is close to 30% when the new treatment is added at n_{01}/n_{0}=0.9. When n_{22} has the same number as n_{0}, we see that BoS becomes even larger. However, the marginal return of having large n_{22} than necessary can be less notable: comparing the dotteddashed line to the dotted line shows that the magnitudes of the differences between the BoS perhaps are less impressive compared to the likely cost of enrolling an extra n_{0} patients for the new arm.
Bias of the estimate when there is a trend
We now present the simulation findings of the inference about the newly added treatment when there is a trend.
Consider the estimated difference in mean responses of the newly added treatment and the control treatment. Table 1 shows the maximum absolute bias (and median absolute bias) in the parameter estimations when there is a trend. Each row corresponds to a trend with a value of Î» for all the scenarios with different combination of n_{01}/n_{0}={0.25,0.5,0.75} and n_{22}={n_{02},n_{0},2n_{0}}. The values with an order of magnitude of 4 are set to zero, as we also obtain such a small number for the unbiased estimators when there is no trend in our simulation, which is due to the Monte Carlo simulation error.
When there is a positive trend in the simulation, we find that the estimates obtained from WLS_{s2},M_{a2} and M_{b2} respectively are unbiased. For those obtained from WLS_{all}, the magnitude of the bias is larger when there is a step trend than when there is a linear trend. This is because the presence of a positive trend inflates the estimate of the overall mean responses of the control treatment while the mean responses of the newly added treatment only have a trend that is similar to the mean responses of the second stage control treatment.
For those obtained from M_{a1} and M_{b1}, the estimated parameter is almost unbiased when there is a linear trend but is biased when there is a step trend. This shows that adjustment with a linear term may not always yield accurate estimate especially when the structure of a trend is not linear.
Marginal power in the presence of a trend
The first row of plots in Figs. 2 and 3 show the marginal power of rejecting H_{02} when there is no trend; other rows of plots correspond to different magnitudes of Î» when there is a linear and a step trend respectively. The xaxis of the plots indicates the value of n_{22} and n_{01}/n_{0}. Each column of plots has a different range of yaxis due to having different values of n_{22} when the new arm is added to the ongoing trial. As expected, the marginal power obtained from WLS_{s2} is similar to that from M_{a2}: the blue line (with Ã—) superimposes the red line (with redâ–³).
We see that the marginal power obtained from WLS_{all} is the highest and from M_{b2} is the lowest among all the comparators. However, the marginal power obtained from WLS_{all} is inflated when there is a positive trend. From our simulation, we find that relative difference in the inflated power to the noninflated power is larger when n_{22}=550 compared to when n_{22}=1100. The reason could be when n_{22} is small (relative to the fixed n_{01}), stage one control responses deviate more on average from the responses in stage two, which leads to a larger bias when using all the control data and hence increases the average number of rejected hypotheses.
On the other hand, the marginal power obtained from M_{a1} and M_{b1} respectively when there is a linear trend remains the same as when there is no trend (i.e. compare the black lines and cyan lines respectively across the row of plots in Fig. 2); but it is inflated when there is a step trend. The magnitude of the inflation increases with the values, Î», of the step trend (see Fig. 3).
We find that the marginal power obtained from WLS_{s2},M_{a2} and M_{b2} respectively remains unchanged when there is a trend. This is not surprising as the impact of a positive trend in the mean responses of the newly added arm and the second stage control arm are of similar magnitude, which is being removed when the difference in mean responses is considered.
Type one error rate in the presence of a trend
Figures 4 and 5 show the type one error rate of rejecting H_{02} in the presence of a linear trend and of a step trend respectively, when Î´_{1}=Î´_{2}=0. A subtle difference in the type one error rate obtained from WLS_{s2} and M_{b2} respectively can be seen in the plots.
We observe that the presence of a step trend would lead to a larger inflation in the type one error rate obtained from WLS_{all} than when there is a linear trend with the same Î», i.e. from comparing the green line (with green+ sign) in Fig. 4 to those in Fig. 5. Besides, the magnitude of inflation is the largest when both the values of n_{01}/n_{0} and Î» are the largest. Having larger n_{22} also causes more inflation in the type one error rate given the same n_{01}/n_{0}. This may be because, given the same magnitude of the bias, large n_{22} makes the denominator of the test statistics smaller, and hence increases the number of hypothesis rejections.
For the type one error rate obtained from M_{a1} and M_{b1}, we see a similar finding to that of the marginal power. That is, when there is a linear trend, the type one error rate is maintained within the Monte Carlo simulation error, whereas it is inflated when there is a step trend (see the black line and cyan line respectively in Fig. 5). We find that the type one error rate obtained from WLS_{s2},M_{a2} and M_{b2} respectively is maintained within the Monte Carlo simulation error when there is a trend.
rMSE in the presence of a trend
The plots of rMSE are presented in Additional file 1: Figures 1 and 2 in the supplement. When there is a positive trend, we find that M_{b2} has the highest rMSE, which remains the same across all the considered values of Î». This is not surprising as the least amount of the data is used. The rMSE values obtained from WLS_{s2} and M_{a2} also remain unchanged when there is a trend.
When there is a linear trend, the rMSE obtained from WLS_{all} increases with Î» while remaining smaller than that from M_{a1} and M_{b1} (which remain unchanged across the linear trends). However, when there is a step trend, the rMSE obtained from these approaches are inflated; the rMSE obtained from WLS_{all} is higher than that from M_{a1} when Î»â‰¥4%. Nevertheless, across all the considered scenarios, the rMSE obtained from M_{a1} is always smaller than the approaches that use less data, i.e. M_{a2},M_{b1},M_{b2}, and WLS_{s2}.
These findings highlight that even though the presence of a small trend would lead to a small bias in the estimation, the quality of the estimate computed using all the control data can be better than that computed using only the concurrent control data. This does not hold for larger trends.
Inference about the initial treatment
For the inference about the initial treatment, we find that the bias, type one error rate and power of rejecting H_{01} are maintained at the same levels as those when there is no trend in the simulation. This finding is observed for all the modelbased approaches. This may be due to using the restricted randomization procedure within each stage of the trial in the simulation, which ensures that not all n_{ks} patients are being enrolled to a particular arm k during the early or the late phase of the trial. As a result, the average impact of a trend on the stagewise mean responses of the initial treatment and of the control treatment are similar, which are then being subtracted when the difference in the mean responses is considered. This observation is consistent with the finding of Ryeznik and Sverdlov [17] who compare different randomization procedures in the context of the standard ANOVA Ftest.
Discussion
When there is no trend, the advantages of including the nonconcurrent control data are greater when the new arm is added at a later time point. Our simulation studies show that given the same magnitude of Î»>0, the presence of a step trend affects the validity of the inference more than that of a linear trend when all the control data are used in the analysis of the new arm. We also examine that using a linear regression model adjusting for a linear term may lead to spurious findings when there is a nonlinear trend, though the rMSE of the corresponding estimate can be lower than other approaches that use only the concurrent control data. This represents a caveat for including nonconcurrent control data in the analysis plan as it is impossible to know the structure of a trend at the planning stage of a trial. After data collection, one can only test the presence of a trend using some quality control techniques such as control charts and CUSUM analysis, see for example a review by Noyez [31]. The analogy between clinical trials and industrial processes is becoming clearer especially when the trial has a long enrolment period. [32â€“38]
This work has highlighted the impact of a trend when the variance of the observations is known. We emphasise that not all methods assume known variance. We conjecture that the presented results would hold when an estimated variance is unbiased, i.e. the estimate is robust to the presence of a trend. Otherwise, the over or underestimated variance would cause the test statistics to be biased and hence provide a spurious result [28]. We also note that when there is a negative trend, i.e. Î»<0, the type one error and the marginal power will be deflated, i.e. smaller than the nominal value. This is a direct consequence of a trend that causes responses to be smaller than the true values as the trial progresses.
All things considered, we would recommend that at the design stage of phase III trials, sample size calculations assume only concurrent control information will be used. When the magnitude of a trend is small, we find including the nonconcurrent control data in the analysis can improve the efficiency of estimating the parameter of interest. However there could be an inflation (or deflation) in the type one error rate and the marginal power of testing the corresponding hypothesis unless the positive (or negative) trend has a magnitude of less than 0.5% of the standard deviation (based on simulation results not presented here). This finding is similar to the finding of KoppSchneider et al [39] for the situation where external information is used in the inference of clinical trials: power gain is not possible when requiring a strict type one error rate control. When the recruitment to all arms do not finish simultaneously, the control data can be separated into before and after adding an arm and those after the initial treatment arms finish recruitment. In this case, the presence of a trend across the three stages can be tested before utilising all the control data to increase the precision of the inference.
We acknowledge that an inflated power is not an issue if we are confident that the intervention is effective. However, using the trial result for other purposes may lead to negative consequences since an inflated power could mean that the estimated effect size is likely to be higher than it should. For example, doing a costeffectiveness analysis using the estimated effect size may overestimate the value of the intervention due to the presence of a positive trend. Future work could review real trials that add arms and explore the presence of trends and their structure in the data. One can also investigate the potential of using nonconcurrent control data for the situations where i) there are more than one treatment arm being added to the ongoing trial; ii) there are more than twostages; and iii) adaptive randomization procedures are considered.
Conclusion
Platform trials can potentially speed up the drug development processes. The feature of continuous recruitment to the control arm may increase the precision of the inference about the newly added interventions in some situations; or reduce the required sample size for the evaluation of the newly added interventions at the cost of having a less stringent control of the error rates. In light of the presence of a potential trend, it is wise to compare the results of including and excluding nonconcurrent control data in the analysis of the newly added interventions.
Availability of data and materials
Not applicable.
Abbreviations
 BoS:

borrowing of strength
 rMSE:

root mean squared error
References
Bothwell LE, Avorn J, Khan NF, Kesselheim AS. Adaptive design clinical trials: a review of the literature and ClinicalTrials.gov. BMJ Open. 2018; 8(2):018320. https://doi.org/10.1136/bmjopen2017018320.
Woodcock J, LaVange LM. Master Protocols to Study Multiple Therapies, Multiple Diseases, or Both. N Engl J Med. 2017; 377(1):62â€“70. https://doi.org/10.1056/NEJMra1510062.
Hirakawa A, Asano J, Sato H, Teramukai S. Master protocol trials in oncology: Review and new trial designs,. Contemp Clin Trials Commun. 2018; 12:1â€“8. https://doi.org/10.1016/j.conctc.2018.08.009.
Angus DC, Alexander BM, Berry S, Buxton M, Lewis R, Paoloni M, Webb SAR, Arnold S, Barker A, Berry DA, Bonten MJM, Brophy M, Butler C, Cloughesy TF, Derde LPG, Esserman LJ, Ferguson R, Fiore L, Gaffey SC, Gaziano JM, Giusti K, Goossens H, Heritier S, Hyman B, Krams M, Larholt K, LaVange LM, Lavori P, Lo AW, London AJ, Manax V, McArthur C, Oâ€™Neill G, Parmigiani G, Perlmutter J, Petzold EA, Ritchie C, Rowan KM, Seymour CW, Shapiro NI, Simeone DM, Smith B, Spellberg B, Stern AD, Trippa L, Trusheim M, Viele K, Wen PY, Woodcock J. Adaptive platform trials: definition, design, conduct and reporting considerations. Nat Rev Drug Discov. 2019:1â€“11. https://doi.org/10.1038/s4157301900343.
Cohen DR, Todd S, Gregory WM, Brown JM. Adding a treatment arm to an ongoing clinical trial: a review of methodology and practice. Trials. 2015; 16(1):179. https://doi.org/10.1186/s130630150697y.
Karrison TG, Huo D, Chappell R. A group sequential, responseadaptive design for randomized clinical trials. Control Clin Trials. 2003; 24(5):506â€“22. https://doi.org/10.1016/S01972456(03)000928.
Chow SC, Chang M, Pong A. Statistical Consideration of Adaptive Methods in Clinical Development. J Biopharm Stat. 2005; 15(4):575â€“91. https://doi.org/10.1081/BIP200062277.
Feng H, Shao J, Chow SC. Adaptive Group Sequential Test for Clinical Trials with Changing Patient Population. J Biopharm Stat. 2007; 17(6):1227â€“38. https://doi.org/10.1080/10543400701645512.
Mahajan R, Gupta K. Adaptive design clinical trials: Methodology, challenges and prospect,. Indian J Pharmacol. 2010; 42(4):201â€“7. https://doi.org/10.4103/02537613.68417.
Pallmann P, Bedding AW, ChoodariOskooei B, Dimairo M, Flight L, Hampson LV, Holmes J, Mander AP, Odondi L, Sydes MR, Villar SS, Wason JMS, Weir CJ, Wheeler GM, Yap C, Jaki T. Adaptive designs in clinical trials: why use them, and how to run and report them,. BMC Med. 2018; 16(1):29. https://doi.org/10.1186/s1291601810177.
Altman DG. Avoiding bias in trials in which allocation ratio is varied. J R Soc Med. 2018; 111(4):143â€“4. https://doi.org/10.1177/0141076818764320.
Coad DS. Sequential estimation with datadependent allocation and time trends. Seq Anal. 1991; 10(12):91â€“97. https://doi.org/10.1080/07474949108836227.
Hu F, Rosenberger WF, Zidek JV. Relevance weighted likelihood for dependent data. Metrika. 2000; 51(3):223â€“43. https://doi.org/10.1007/s001840000051.
Rosenkranz GK. The impact of randomization on the analysis of clinical trials. Stat Med. 2011; 30(30):3475â€“87. https://doi.org/10.1002/sim.4376.
Tamm M, Hilgers RD. Chronological Bias in Randomized Clinical Trials Arising from Different Types of Unobserved Time Trends. Methods Inf Med. 2014; 53(06):501â€“10. https://doi.org/10.3414/ME14010048.
Hilgers RD, Uschner D, Rosenberger WF, Heussen N. ERDO  a framework to select an appropriate randomization procedure for clinical trials. BMC Med Res Methodol. 2017; 17(1):159. https://doi.org/10.1186/s128740170428z.
Ryeznik Y, Sverdlov O. A comparative study of restricted randomization procedures for multiarm trials with equal or unequal treatment allocation ratios. Stat Med. 2018; 37(21):3056â€“77. https://doi.org/10.1002/sim.7817.
Lipsky AM, Greenland S. Confounding due to changing background risk in adaptively randomized trials,. Clin Trials. 2011; 8(4):390â€“7. https://doi.org/10.1177/1740774511406950.
Villar SS, Bowden J, Wason J. Responseadaptive designs for binary responses: How to offer patient benefit while being robust to time trends?. Pharm Stat. 2018; 17(2):182â€“97. https://doi.org/10.1002/pst.1845.
Jiang Y, Zhao W, DurkalskiMauldin V. Timetrend impact on treatment estimation in twoarm clinical trials with a binary outcome and Bayesian response adaptive randomization. J Biopharm Stat. 2019:1â€“20. https://doi.org/10.1080/10543406.2019.1607368.
Hilgers RD, Manolov M, Heussen N, Rosenberger WF. Design and analysis of stratified clinical trials in the presence of bias. Stat Methods Med Res. 2019; 096228021984614. https://doi.org/10.1177/0962280219846146.
Greenland S.Interpreting timerelated trends in effect estimates. J Chron Dis. 1987; 40:17â€“24. https://doi.org/10.1016/S00219681(87)80005X.
Wason J, Magirr D, Law M, Jaki T. Some recommendations for multiarm multistage trials,. Stat Methods Med Res. 2016; 25(2):716â€“27. https://doi.org/10.1177/0962280212465498.
Ventz S, Cellamare M, Parmigiani G, Trippa L. Adding experimental arms to platform clinical trials: randomization procedures and interim analyses. Biostatistics. 2018; 19(2):199â€“215. https://doi.org/10.1093/biostatistics/kxx030.
Lee KM, Wason J, Stallard N. To add or not to add a new treatment arm to a multiarm study: A decisiontheoretic framework. Stat Med. 2019; 38(18):8194. https://doi.org/10.1002/sim.8194.
Saville BR, Berry SM. Efficiencies of platform clinical trials: A vision of the future. Clin Trials J Soc Clin Trials. 2016; 13(3):358â€“66. https://doi.org/10.1177/1740774515626362.
Butler CC, Coenen S, Saville BR, Cook J, van der Velden A, Homes J, de Jong M, Little P, Goossens H, Beutels P, Ieven M, Francis N, Moons P, Bongard E, Verheij T. A trial like ALIC4E: why design a platform, responseadaptive, open, randomised controlled trial of antivirals for influenzalike illness?. ERJ Open Res. 2018; 4(2). https://doi.org/10.1183/23120541.000462018.
Elm JJ, Palesch YY, Koch GG, Hinson V, Ravina B, Zhao W. Flexible Analytical Methods for Adding a Treatment Arm MidStudy to an Ongoing Clinical Trial. J Biopharm Stat. 2012; 22(4):758â€“72. https://doi.org/10.1080/10543406.2010.528103.
ChoodariOskooei B, Bratton DJ, Gannon MR, Meade AM, Sydes MR, Parmar MK. Adding new experimental arms to randomised clinical trials: impact on error rates. 2019. https://doi.org/1902.05336.
Jackson D, White IR, Price M, Copas J, Riley RD. Borrowing of strength and study weights in multivariate and network metaanalysis. Stat Methods Med Res. 2017; 26(6):2853â€“68. https://doi.org/10.1177/0962280215611702.
Noyez L. Control charts, Cusum techniques and funnel plots. A review of methods for monitoring performance in healthcare. Interact Cardiovasc Thorac Surg. 2009; 9(3):494â€“9. https://doi.org/10.1510/icvts.2009.204768.
Chang WR, McLean IP. CUSUM: A tool for early feedback about performance?,. BMC Med Res Methodol. 2006; 6(1):8. https://doi.org/10.1186/1471228868.
Sibanda T, Sibanda N. The CUSUM chart method as a tool for continuous monitoring of clinical outcomes using routinely collected data. BMC Med Res Methodol. 2007; 7(1):46. https://doi.org/10.1186/14712288746.
McLaren PJ, Hart KD, Dolan JP, Hunter JG. CUSUM analysis of mortality following esophagectomy to allow for identification and intervention of quality problems. J Clin Oncol. 2017; 35(4_suppl):203â€“203. https://doi.org/10.1200/JCO.2017.35.4suppl.203.
Neuburger J, Walker K, SherlawJohnson C, van der Meulen J, Cromwell DA. Comparison of control charts for monitoring clinical performance using binary data,. BMJ Qual Saf. 2017; 26(11):919â€“28. https://doi.org/10.1136/bmjqs2016005526.
ForteaSanchis C, MartÃnezRamos D, EscrigSos J. CUSUM charts in the quality control of colon cancer lymph node analysis: a populationregistry study. World J Surg Oncol. 2018; 16(1):230. https://doi.org/10.1186/s1295701815330.
Redd D, Shao Y, Cheng Y, ZengTreitler Q. Detecting Secular Trends in Clinical Treatment through Temporal Analysis. J Med Syst. 2019; 43(3):74. https://doi.org/10.1007/s1091601911730.
ForteaSanchis C, EscrigSos J. Quality Control Techniques in Surgery: Application of Cumulative Sum (CUSUM) Charts. Cir Esp (English Edition). 2019; 97(2):65â€“70. https://doi.org/10.1016/j.cireng.2019.01.010.
KoppSchneider A, Calderazzo S, Wiesenfarth M. Power gains by using external information in clinical trials are typically not possible when requiring strict type I error control. Biom J. 2019; 201800395. https://doi.org/10.1002/bimj.201800395.
Acknowledgements
We are grateful to the referees for their helpful comments on an earlier version of this paper.
Funding
This research was funded by the Medical Research Council (grant codes MC _UU_00002/6 and MR /N028171 /1). The funder played no role in the design or conduct of the research, nor decision to publish.
Author information
Authors and Affiliations
Contributions
KL conceived and designed the research and drafted the manuscript. JW provided important intellectual input to interpretation of findings and manuscript writing. All authors critically appraised the final manuscript.
Corresponding author
Ethics declarations
Ethics approval and consent to participate
Not applicable.
Consent for publication
Not applicable.
Competing interests
The authors declare that they have no competing interests.
Additional information
Publisherâ€™s Note
Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.
Supplementary information
Additional file 1
Figures 1 and 2 show the rMSE when there is a linear and a step trend respectively in the simulation.
Rights and permissions
Open Access This article is licensed under a Creative Commons Attribution 4.0 International License, which permits use, sharing, adaptation, distribution and reproduction in any medium or format, as long as you give appropriate credit to the original author(s) and the source, provide a link to the Creative Commons licence, and indicate if changes were made. The images or other third party material in this article are included in the articleâ€™s Creative Commons licence, unless indicated otherwise in a credit line to the material. If material is not included in the articleâ€™s Creative Commons licence and your intended use is not permitted by statutory regulation or exceeds the permitted use, you will need to obtain permission directly from the copyright holder. To view a copy of this licence, visithttp://creativecommons.org/licenses/by/4.0/. The Creative Commons Public Domain Dedication waiver (http://creativecommons.org/publicdomain/zero/1.0/) applies to the data made available in this article, unless otherwise stated in a credit line to the data.
About this article
Cite this article
Lee, K.M., Wason, J. Including nonconcurrent control patients in the analysis of platform trials: is it worth it?. BMC Med Res Methodol 20, 165 (2020). https://doi.org/10.1186/s12874020010436
Received:
Accepted:
Published:
DOI: https://doi.org/10.1186/s12874020010436