 Research
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Implications of analysing timetoevent outcomes as binary in metaanalysis: empirical evidence from the Cochrane Database of Systematic Reviews
BMC Medical Research Methodology volume 22, Article number: 73 (2022)
Abstract
Background
Systematic reviews and metaanalysis of timetoevent outcomes are frequently published within the Cochrane Database of Systematic Reviews (CDSR). However, these outcomes are handled differently across metaanalyses. They can be analysed on the hazard ratio (HR) scale or can be dichotomized and analysed as binary outcomes using effect measures such as odds ratios (OR) or risk ratios (RR). We investigated the impact of reanalysing metaanalyses from the CDSR that used these different effect measures.
Methods
We extracted two types of metaanalysis data from the CDSR: either recorded in a binary form only (“binary”), or in binary form together with observed minus expected and variance statistics (“OEV”). We explored how results for timetoevent outcomes originally analysed as “binary” change when analysed using the complementary log–log (cloglog) link on a HR scale. For the data originally analysed as HRs (“OEV”), we compared these results to analysing them as binary on a HR scale using the cloglog link or using a logit link on an OR scale.
Results
The pooled HR estimates were closer to 1 than the OR estimates in the majority of metaanalyses. Important differences in betweenstudy heterogeneity between the HR and OR analyses were also observed. These changes led to discrepant conclusions between the OR and HR scales in some metaanalyses. Situations under which the cloglog link performed better than logit link and vice versa were apparent, indicating that the correct choice of the method does matter. Differences between scales arise mainly when event probability is high and may occur via differences in betweenstudy heterogeneity or via increased withinstudy standard error in the OR relative to the HR analyses.
Conclusions
We identified that dichotomising timetoevent outcomes may be adequate for low event probabilities but not for high event probabilities. In metaanalyses where only binary data are available, the complementary log–log link may be a useful alternative when analysing timetoevent outcomes as binary, however the exact conditions need further exploration. These findings provide guidance on the appropriate methodology that should be used when conducting such metaanalyses.
Background
Systematic reviews and metaanalyses of timetoevent outcomes (e.g. time to death, recurrence of symptoms, relief of pain etc.) are frequently carried out in areas such as cancer, respiratory and cardiovascular diseases, since event timings are crucial to assessing the impact of an intervention [1]. The decision on how timetoevent outcomes are handled in a particular metaanalysis largely depends on how eligible studies are reported, and is usually out of the control of the metaanalyst except if individual participant data (IPD) are available. The information extracted by systematic reviewers may include the total number of participants and events per arm, and/or the hazard ratio alongside its confidence interval, and/or the logrank observed minus expected statistic (“OE”) and its variance (“V”) (which are useful alternative statistics if a hazard ratio is not directly reported [1]). Timetoevent data can be analysed using the effect measure of hazard ratio (HR), or can be dichotomised and analysed as binary using effect measures such as the odds ratio (OR) or risk ratio (RR) [2]. Although HR is considered the most appropriate scale for analysis of timetoevent data, in practice OR and RR are frequently used instead due to the following reasons: unavailability of individual participant data (IPD); limitations on how these outcomes are reported in individual trial reports; lack of familiarity in handling timetoevent outcomes for metaanalysis; difficulties in understanding the methods of analysing such data without a statistician; limited available training for the majority of systematic reviewers and metaanalysts who perform such analyses [3].
In the past, research was conducted comparing the differences between the OR using logistic regression models and the HR using proportional hazard (PH) models within individual studies. Green and Symons [4] showed that logistic and Cox PH models produce similar results when the event is rare and for shorter followup times under a constant hazard rate. Ingram and Kleinman [5] added that important differences among the methods occur in the presence of varying censoring rates and length of followup. However, it has not been established yet how such results transfer to the context of an aggregate data metaanalysis for which summary data is extracted from trial reports. Further, in this context it is of interest to examine potential alternatives such as the use of the complementary log–log link, which may reduce the difference in the results between the two effect measures used. The overall metaanalytic estimate can be affected due to changes to the weighting allocated to each study, and therefore changes to the results can be unpredictable. We aimed to carry out an empirical “metaepidemiological” study using survival metaanalysis data from the Cochrane Database of Systematic Reviews (CDSR) (Issue 1, 2008) to explore the implications of analysing timetoevent outcomes as binary in metaanalysis. We assessed the importance of extracting suitable data such as the “OE” and “V” statistics rather than binary summaries to perform such analyses; in the occasion where binary data were available we examined whether the use of alternative methodology such as the complementary log–log link (cloglog), proven to facilitate interpretation of the results on a HR scale [6, 7] can minimise the error we may observe in the results. We assess only the differences between the OR and the HR, as the RR, according to the literature [8,9,10,11], is placed in between these measures and therefore, we expect to capture any bias within these extremes. We perform these analyses under both two and onestage models.
The rest of the paper is set out as follows. In the methods section, we describe the dataset we used and the statistical models that we applied. In the results, we present descriptive statistics of the database and then we describe the results obtained from reanalysing the data originally analysed as binary on an HR scale and from reanalysing the data originally analysed using “OE” and “V” data on an OR scale. These results are followed by a discussion exploring the strengths and limitations of our findings, together with conclusions and implications.
Methods
Data
The Nordic Cochrane Centre provided the content of the first issue from 2008 of the CDSR. The database includes metaanalyses within reviews which have been classified previously by outcome type, medical specialty and types of interventions included in the pairwise comparisons [12]. The database did not record whether data type was timetoevent; however, based on the outcome classification we were able to identify (using words such as “survival”, “death”, “fatality”) three sets of timetoevent metaanalyses:

“binary”: Those with outcome classification “allcause mortality” where the information recorded was based only on the number of events and participants per arm;

“OEV”: Those with outcome classifications “overall survival” and “progression/disease free survival” where the information recorded was based on “binary” data in addition to logrank “OE” and “V” statistics”; these were originally analysed as HRs in the RevMan software;

Those with estimated log HR and its standard error. These were removed from further analyses since there was no available information on the number of events and participants per arm and therefore no binary data metaanalysis could be conducted.
Therefore, we identified two subsets of timetoevent metaanalyses: those with binary summaries, and those with binary summaries in addition to OEV data; we analysed each outcome per dataset separately to assess whether differences exist due to different characteristics of the outcomes. We also examined whether the information obtained from “OEV” data was based on aggregate data or IPD by examining the individual Cochrane reviews.
Eligibility Criteria
RMT (for “binary” data) and TS (for “OEV” data) initially extracted these data and conducted cleaning including examination of the outcome classification; TS repeated the “binary” data extraction to confirm the information obtained were accurate and RMT confirmed the choice of included metaanalyses obtained from “OEV” data extraction. Both datasets could contribute more than one metaanalysis per Cochrane review. RMT and TS identified 46 misclassifications due to disagreement with the original outcome classification as listed in the datasets, conflicting information in the database or unavailability of the correct version of the Cochrane review. We excluded 1,284 studies including double zero events, since they do not contribute to the metaanalysis results [12, 13]. We removed another 359 metaanalyses including fewer than 3 studies because some of the models applied below (i.e. generalised linear mixed models) will be affected by estimation issues and inevitable failures using small numbers of studies [14]; hence we wanted to make fair comparisons between the models applied. Derivation of the analysis sample is provided in Fig. 1.
Descriptive statistics
We describe the number of studies per metaanalysis, number of events and study size by the median and interquartile range. We also identify the number of medical specialities, and median number of events (and interquartile range) per medical specialty.
Model description for “binary” data
We used the following metaanalysis models to analyse the data on the OR or HR scale. The first was a model proposed for “binary” data (assuming a binomial likelihood with a logit link) which is based only on the number of patients and number of events which occurred. Interpretation for the treatment effect is conducted in terms of the logarithm of an OR.
In the second approach, we modelled the binary data using a normal approximation to binomial likelihood with a complementary log–log link (cloglog), where treatment effect interpretation was based on the logarithm of a HR. This method is also based only on the number of patients and events which occurred, and ignores censoring and the time element; however it is closely related to continuoustime models, has a builtin proportional hazards assumption, and therefore has important application in survival analysis [6].
Fitting twostage randomeffects models for “binary” data
Prior to fitting the twostage randomeffects models, study arms with zero events were identified for the “binary” data. For 771 studies, a “treatment arm” continuity correction was applied as proposed by Sweeting et al. [15] and was constrained to sum to one as this ensures that the same amount of information is added to each study.
Let \(i=\mathrm{1,2},\dots ,n\) denote the study. The estimated log odds and log hazard ratios were given by:
where \({\mathrm{A}}_{\mathrm{i}},{\mathrm{C}}_{\mathrm{i}}\) represented number of events, \({\mathrm{B}}_{\mathrm{i}},{\mathrm{D}}_{\mathrm{i}}\) represented number of nonevents in the treatment and control groups respectively, \({P}_{Ti}=\frac{{\mathrm{A}}_{\mathrm{i}}}{{\mathrm{A}}_{\mathrm{i}}+{\mathrm{B}}_{\mathrm{i}}}\) was the proportion of events on the treatment arm of the \({i}^{th}\) study, and \({P}_{Ci}=\frac{{\mathrm{C}}_{\mathrm{i}}}{{\mathrm{C}}_{\mathrm{i}}+{\mathrm{D}}_{\mathrm{i}}}\) was the proportion of events on the control arm of the \({i}^{th}\) study.
The corresponding variances were given by:
Equations 2 and 4 provided a HR estimate via the use of the complementary log–log link considered as a useful link function for the discretetime hazards models as recommended by Hedeker et al. [7] and Singer et al. [6]. We estimated the studyspecific log odds ratios or log hazard ratios, \({y}_{i}\) and their withinstudy variances \({s}_{i}^{2}\) as shown above and fitted a standard twostage randomeffects model to these. Additionally, we obtained the \({I}^{2}\) statistic from the fitted models as follows:
where \({\tau }^{2}\) denotes the variance of the underlying true effects across studies and \({\sigma }^{2}\) the typical withinstudy variance.
To avoid downward bias in the variance components estimates, we used the REML estimator for model implementation [16]. The models were implemented via the “rma.uni” command from “metafor” package in R. We also fitted onestage randomeffects models for “binary” data. The methods related to onestage metaanalysis models and code is available in Additional file 1.
Model description for “OEV” data
For “OEV” data, the “OE” and “V” statistics were available in the Cochrane database alongside the number of patients and events. These data came either from published reports or from IPD; TS examined the individual reviews from the Cochrane database and assessed the data origin. Since there were more available information for these data the following three models were applied, using only twostage metaanalysis models.
Similarly to “binary” data, we initially analysed the “OEV” data as “binary” and modelled them as described in detail in the preceding section. We also used the logrank Observed—Expected events (OE) and the logrank Variance (V) statistics calculated previously from the number of events and the individual times to event on each research arm of the trial; we used the logrank approach [17] in order to obtain another type of HR estimate. We used randomeffects models to analyse the data throughout, including betweenstudy heterogeneity to account for variation across studies.
Fitting twostage randomeffects models for “OEV” data
Similarly to the “binary” data, the estimated log odds and log hazard ratios were given by Eqs. 1 and 2 for the binary summaries while the “OE” and “V” statistics were used as follows:
The corresponding variances were given by Eqs. 3 and 4 for binary summaries while for “OE” and “V” statistics as follows:
where \(V\) denotes the variance of the logrank statistic. We used the REML estimator for model implementation [16] and the models were implemented via the “rma.uni” command from “metafor” package in R.
Model comparison for “binary” data
The following model comparisons were performed. For the “binary” data set, we examined whether the results from analysing survival data as binary on an OR scale are similar to results from analysing on the HR scale using the cloglog link, both under twostage and onestage models. For presentation purposes, we present only comparisons of the results under twostage models in the main paper (and for onestage models in the Additional file 1) in order to assess the discrepancies between the model using the logit link and the model using the complementary log–log link.
First, we examined the proportion of significant and nonsignificant metaanalytic pooled effect estimates under the different scales used (OR vs HR scale); we identified the number of metaanalyses which were significant under one scale and nonsignificant under the other at a twosided 5% level of significance.
Bland–Altman plots with associated 95% limits of agreement were constructed, with the aim of facilitating interpretation of results and producing fair comparisons between the two scales [18]. In order to create these plots, results were standardised by dividing the logarithm of the estimate by its standard error. Plots were produced for the standardised treatment effect estimates and for the \({I}^{2}\) statistics. \({I}^{2}\) represents the percentage of variability that is due to betweenstudy heterogeneity rather than chance; \({I}^{2}\) values range from 0 to 100%. This measure was chosen for model comparison as it enables us to compare results directly between the two scales used. The variance of underlying true effects across studies (\({\tau }^{2}\)) was not used as it does not allow direct comparison between different outcome measures.
We identified “outliers” as metaanalyses outside the 95% limits of agreement, and we examined their characteristics. The metaanalysis characteristics we examined were the following:

betweenscale differences in the magnitude of the pooled treatment effect estimate and its 95% confidence intervals

the levels of withinstudy standard error and betweenstudy heterogeneity and study weights in the metaanalysis

studyspecific event probabilities and baseline risk
We summarised these differences by metaanalysis and reported those characteristics which were mostly associated with substantial differences between OR pooled effect estimates and corresponding HR pooled effect estimates.
Model comparison for “OEV” data
For the “OEV” data set, comparisons on overall and progression disease free survival outcomes were conducted separately; this was because differences between these outcomes might be observed in the presence of different disease severities, and therefore this would be associated with different length of followup and risk of the outcome.
For both outcomes, we performed comparisons by examining the differences between analysing the data as binary on an OR scale, analysing the data as binary using the cloglog link on a HR scale, or analysing the data using the “OE” and “V” statistics on a HR scale. We assessed whether the differences observed from analysing the data as binary on an OR scale could be reduced by the use of the cloglog link. We present only comparisons of the results under twostage models since there were no available IPD to perform comparisons under onestage models.
Similarly to “binary” data, we examined the proportion of significant and nonsignificant metaanalytic pooled effect estimates under the different scales used and identified the number of metaanalyses which were significant under one scale and nonsignificant under the other. We created Bland–Altman plots for the standardised treatment effect estimates and for the \({I}^{2}\) statistics to explore the agreement among the methods producing fair comparisons between the two scales [18]. Metaanalyses outside the 95% limits of agreement were examined for their characteristics.
Results
Results for “binary” data
For the outcome of “allcause mortality”, 1,132 metaanalyses within the Cochrane database were originally analysed as binary. The median number of metaanalyses per review was 1 with IQR (1,2). The median number of studies and the median number of events are provided in Table 1, indicating that these numbers were a lot smaller than those obtained for the “OEV” data.
The distribution of medical specialities of the metaanalyses is presented in Table 2. For the “binary” data, “Cardiovascular” (23%) is the most frequently occurring category, followed by “Cancer” (13%), “Gynaecology, pregnancy and birth” (12%) and “respiratory diseases” (12%). The median number of events in cancer substantially exceeded the median number of events in other medical areas.
Once the models were applied, we compared results between OR and HR analyses. Table 3 provides the percentages of significant and nonsignificant metaanalyses at a twosided 5% level of significance indicating that there are few discrepancies present for both “binary” and “OEV” datasets under twostage models.
According to the Bland–Altman plot (Fig. 2), the average difference between the two methods for the standardised pooled effect estimates was 0.004 units (0.222 units, 0.214 units) and 0.1% (10.6%, 10.3%) for the estimation of I^{2} for twostage models; this indicates a relatively small percentage difference between the two methods in the estimation of the measure of impact of heterogeneity I^{2}. The width of the 95% limits of agreement is small, indicating acceptable agreement between the two methods except in specific circumstances mentioned below. The corresponding results for onestage models are presented in Additional file 1.
Based on Bland–Altman plots, 6% (n = 47) of the metaanalyses were considered as outliers. In 21% of the “binary” outlying metaanalyses (e.g. MA 327; outlier obtained from \({I}^{2}\) estimates) a high event probability (defined here as probability greater than 0.7 for the majority of the individual studies) was observed. For example, metaanalysis 327 consists of 7 studies for which the event probability was greater than 0.7 for 5 out of 7 studies; consequently, high event probability affected substantially the differences in the individual study estimates between the OR and HR analyses, leading to different allocated relative weights for the studies, and discrepancies in the pooled effect estimates as shown in Fig. 3.
The pooled HR estimates were closer to 1 than the OR estimates in the majority of metaanalyses (Additional file 1; outlier obtained from standardised and \({I}^{2}\) estimates) with the exception of MA 574 for “binary” data where, even though most of the individual study HR estimates are closer to 1 than the individual OR estimates, the pooled HR estimate is further from 1 than the pooled OR estimate. Increased withinstudy variability on the OR scale relative to the HR scale may affect the weighting more than the actual estimates in the studies, for example within “binary” data metaanalysis 7 (Additional file 1; outlier obtained from standardised estimates), producing some differences in the pooled effect estimates between the two scales. Important differences in betweenstudy heterogeneity between the HR and OR analyses were also observed. For example, metaanalysis 330 (outlier obtained from \({I}^{2}\) estimates) consists of 8 studies of which 6 are smaller studies which received increased weight in the HR analysis compared to the OR analysis while the two larger studies received smaller weights; this affected both the individual HR estimates that have moved closer to each other and the relevant weights of the studies as presented in Fig. 4.
In 34% of the outlying metaanalyses, the individual study estimates and the corresponding weights were affected by a combination of differing event probability across study arms, differences in betweenstudy heterogeneity or increased withinstudy variability on the OR relative to the HR scale. In the presence of a limited amount of studies in the metaanalyses this was even more evident. Additional examples of forest plots indicating the discrepancies among the results are shown in Additional file 1.
Results for “OEV” data
In the Cochrane database, 157 metaanalyses were originally analysed using the “OE” and “V” statistics on a HR scale. The median number of metaanalyses per review was 2 with IQR (2, 3). We observed that analysing timetoevent outcomes as HRs is restricted to very few medical specialties (Tables 2). For the “OEV” data, “Cancer” was still the most frequent medical specialty for both outcomes as observed in “binary” data (Table 2).
Table 3 provides the percentages of significant and nonsignificant metaanalyses for each outcome for twostage models, indicating that discrepancies are more prevalent in the “OEV” data compared to the “binary” data; additionally the amount of discrepancies observed in statistical significance from the comparison of OR and HR obtained from the cloglog link was smaller than the amount of discrepancies observed between the OR and HR analyses.
Bland–Altman plots produced for “OEV” data indicated that the average difference between each pair of methods is larger than those obtained from the “binary” data (Figs. 5 and 6). For example, for overall survival, the average difference between the two methods for the standardised pooled effect estimates was 0.2 units (1.8 units, 2.1 units) for OR versus HR and 0.2 units (2.2 units, 2.5 units) for HR using cloglog versus HR; however, for OR vs HR cloglog differences the average bias was 0 units (2.6 units, 2.7 units) indicating that cloglog is a closer approximation to OR rather than HR analyses (Fig. 5). For the estimation of I^{2}, the average difference between the methods is 6% (41%, 29%) for OR versus HR, 6% (42%, 31%) for HR using cloglog versus HR, and 0% (21%, 21%) for OR vs HR cloglog differences; similarly the cloglog seems a closer approximation to OR analyses rather than HR analyses (Fig. 6). The corresponding results for the outcome of progression/disease free survival are shown in Additional file 1.
Outliers were considered 28% of the “OEV” metaanalyses. Of these, 57% were from IPD rather than nonIPD and 54% of them were for the outcome of overall survival. In 50% of the outliers a high event probability (defined here as probability greater than 0.7) was observed, suggesting that this may be an important factor associated with differences among the scales used. For example, metaanalysis 45 (outlier obtained from standardised estimates) consists of 7 studies for which the event probability was greater than 0.7 for all the studies; consequently high event probability affected substantially the differences in the individual study estimates between the OR and HR analyses, leading to different allocated relative weights for the studies, and discrepancies in the pooled effect estimates as shown in Fig. 7. Even though the individual HR cloglog estimates were closer to the individual OR estimates the final pooled effect estimate was closer to the pooled HR estimate; this was not though the case for all metaanalyses.
Increased withinstudy variability on the OR scale relative to the HR scale may affect the weighting more than the actual estimates in the studies, for example for metaanalysis 17 (Additional file 1; outlier obtained from standardised estimates), producing differences in the pooled effect estimates between the two scales. Similarly, even though the individual study estimates and weights of OR and HR cloglog were closer to each other, the HR cloglog pooled effect estimate was closer to the pooled HR estimate; however, this was not the case for all metaanalyses. Important differences in betweenstudy heterogeneity between the HR and OR analyses were observed in metaanalyses such as 42, 90. For example, metaanalysis 90 (outlier obtained from \({I}^{2}\) estimates) consists of 11 studies out of which 8 are smaller studies and 3 are larger studies. Smaller studies received increased weight in the HR analysis compared to the OR analysis, while larger studies received smaller weights in the HR scale compared to OR scale. However, this was not the case on the HR cloglog scale as presented in Fig. 8.
In 46% of the outlying metaanalyses, the individual study estimates, and the corresponding weights were affected by a combination of differing event probability across study arms, differences in betweenstudy heterogeneity or increased withinstudy variability on the OR relative to the HR scale. In the presence of a limited amount of studies in the metaanalyses this was even more evident. Additional forest plots indicating the discrepancies among the results are shown in Additional file 1.
Overall, using the “OEV” data, a mixed pattern was observed. In 39% (n = 11) of outlying metaanalyses the OR pooled effect estimate was closer to HR pooled effect estimate; however in 4 out of 11 outlying metaanalyses the individual study estimates obtained from the HR cloglog link were a closer approximation to the individual study HR estimates. Similarly, even though in 61% (n = 17) of the outlying metaanalyses the HR cloglog pooled effect estimate was closer to the pooled HR estimate, 3 of outlying metaanalyses provided individual study OR estimates closer to individual study HR estimates, and another 3 individual study HR cloglog estimates were closer to individual study OR estimates.
Discussion
Using metaanalysis data from the CDSR of 2008, we investigated how timetoevent outcomes are treated within metaanalysis; we explored the differences that occur when data are analysed as binary as opposed to analysing the data using the complementary log–log link or using the “OE” and “V” statistics where interpretation is conducted on a HR scale. For both datasets, we identified important reasons associated with discordance among the results, indicating that the correct choice of the method does matter and may affect the interpretation and conclusions drawn from the results. Our analyses highlighted that high event probability was an important factor associated with discordant effect estimates; changes to between and withinstudy variation were important mechanisms producing differences in the results as well. However, there were occasions where there was no clear single factor driving the differences, since there was a combination of reasons affecting the individual study estimates and corresponding weights. Regarding method selection, based on the “OEV” data we identified that a mixed pattern was observed and there was no clear indication under which exact conditions the cloglog link outperforms logit link on an OR scale and vice versa.
While most of the metaanalyses within the database were analysed originally as binary, with an outcome classification of allcause mortality it is worth mentioning that these metaanalyses could include the outcome of shortterm mortality (e.g. 30 days) or longerterm mortality (e.g. 5 years); therefore some of these metaanalyses with short followup may have been appropriately analysed as binary. The outcome classification of allcause mortality was considered a representative sample of survival metaanalysis up to 2008, however results might be different for other outcomes and results might have changed in later reviews where more information on methodology was available. The data used for the comparison of OR/HR scale in the “OEV” data were slightly different; we used the number of events and nonevents for the OR and HR cloglog calculation (as in “binary” data) and calculated a HR based on “OE” and V statistics. Therefore, there is a possibility for some cases that the two data sets entered by Cochrane reviewers may not completely correspond to each other.
We did not assess other reasons for differences between the results due to lack of information on censoring and followup times. Interpretation of the results was conducted with caution as we are interpreting the results based on known factors, without excluding other unknown factors that may have affected the results. We were not able to examine whether current practice of analysing timetoevent data has changed and whether methodological choices have improved since 2008. Further work examining the differences observed between analyses on the OR and HR scales in the presence of IPD is necessary.
The model used to analyse timetoevent data as binary is the conventional approach widely used by many systematic reviewers and metaanalysts [19]. It is quick, inexpensive and study results are obtained from appropriately synthesized study publications or by contacting study authors [20]. This approach to analysis ignores censored observations [21] and treats them as missing and has also been criticised for the withinstudy normality assumptions required [20].
The use of a cloglog link function, facilitating the results’ interpretation in a HR scale for both “binary” and “OEV” data, was the best alternative approach enabling us to make comparisons between the scales used if only binary summaries are available. In the past, the cloglog link has been proven to provide a close approximation to Cox regression invoking a proportional hazards assumption, rather than a proportional odds assumption [6]. However, due to lack on information on “OE” and “V” statistics for “binary” data only, we were not able to assess whether the HR obtained from the cloglog link is a close approximation to the true HR; therefore this magnifies the importance of extracting appropriate information when conducting timetoevent metaanalysis. For the “OEV” data, “OE and V” data provide the best method to analyse aggregate data and facilitate results’ interpretation on the HR scale but in the absence of IPD important biases may occur when large treatment effects and unbalanced data are present [22]. Additionally, we were not able to identify a clear pattern under which the complementary log–log link could be employed since there were circumstances under which it performed better or worse than an OR analysis; therefore we were not able to identify whether the cloglog approach is useful when a MA includes binary summaries alongside OEV or HR summaries. IPD and simulation studies are required to assess in more detail the conditions determining where this method would be acceptable.
For the “binary” data, we also used a onestage randomeffects model with fixed studyspecific effects describing the baseline risk probability of the event in each study. These models use exact binomial likelihoods and may therefore be more accurate, especially with sparse data [14]. The fixed studyspecific effects cause difficulties in estimation since the number of parameters increases with the number of studies, but maximum likelihood theory requires the number of parameters to remain stable as the sample size increases. A randomeffects model with random studyspecific effects could be applied, however based on simulation studies this model performed better than others without any serious biases present [14]. We were not able to make comparisons using onestage models in the “OEV” data. We would be able to apply onestage models when the data were analysed as binary, but we did not have the IPD required to fit onestage models on the HR scale.
To our knowledge, no research has been conducted using such a large database assessing the differences between a) analysing the data as binary and interpreting the results in an OR scale and b) analysing the data either using the cloglog link or logrank “OE” and V statistics facilitating interpretation on the HR scale.
We have demonstrated the impact of reanalysing metaanalyses (“binary” or “OEV” datasets) within the Cochrane Database on a different scale, identifying the main drivers influencing discrepancies between the metaanalytic results. Our findings provide useful insights into changes to metaanalytical results and indicate that choice of method used in metaanalysis of survival data does matter, especially in the presence of high event probabilities.
Conclusions
In conclusion, our findings indicate that timetoevent data should be ideally analysed accounting for their natural properties, as it is possible for important discrepancies to be observed and conclusions from the metaanalysis to be altered. We identified that dichotomising timetoevent outcomes may be adequate for low event probabilities but not for high event probabilities. In metaanalyses where only binary data are available, the complementary log–log link may be a useful alternative when analysing timetoevent outcomes as binary, however the exact conditions need further exploration. These findings provide guidance on the appropriate methodology that should be used when conducting such metaanalyses.
Availability of data and materials
Data are available upon reasonable request, if permission is obtained from Cochrane.
Abbreviations
 CDSR:

Cochrane Database of Systematic Reviews
 HR(s):

Hazard Ratio(s)
 IQR:

Interquartile Range
 IPD:

Individual Participant Data
 MA(s):

Metaanalysis(es)
 OEV:

Observed minus Expected and Variance statistics
 OR(s):

Odds Ratio(s)
 REML:

Restricted Maximum Likelihood
 RR(s):

Risk Ratio(s) or Relative Risk(s)
 SR(s):

Systematic Review(s)
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Acknowledgements
We are grateful to the Nordic Cochrane Centre and the Cochrane Collaboration Steering Group for providing us with access to the Cochrane Database of Systematic Reviews. We would like to thank James Carpenter for the valuable suggestions and discussions we had during the preparation of this project. We would also like to thank Larysa Rydzewska for providing us with results obtained from the MRC Clinical Trials Unit’s Survey of Collaborative Review Groups.
Funding
TS received a Doctoral Training Grant from the UK Medical Research Council. RMT, DF, JFT and IRW were supported by the Medical Research Council Programme MC_UU_00004/06. The funders had no direct role in the writing of the manuscript or decision to submit it for publication.
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RMT proposed the study. TS performed the statistical analyses and drafted the manuscript. TS, RMT, DF, JFT and IRW jointly contributed to interpreting the results and to revising the manuscript. All authors approved the final manuscript.
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Supplementary Information
Additional file 1: Section 1.
Fitting onestage randomeffects models for “binary” data. Section 2. Number (%) of (non)significant metaanalyses under different scales for onestage models (“binary” data). Section 3. BlandAltman plots comparing standardised pooled effect and \({I}^{2}\) estimates for onestage models (“binary” data). Section 4. Forest plots for example MAs considered as outliers in our analyses (“binary” data). Section 5. BlandAltman Plot comparing standardised OR vs. HR estimates for twostage models in “OEV” data. Section 6. Forest plot for example MAs considered as outliers in our analyses (“OEV” data). Section 7: R Code
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Salika, T., Turner, R.M., Fisher, D. et al. Implications of analysing timetoevent outcomes as binary in metaanalysis: empirical evidence from the Cochrane Database of Systematic Reviews. BMC Med Res Methodol 22, 73 (2022). https://doi.org/10.1186/s12874022015419
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DOI: https://doi.org/10.1186/s12874022015419
Keywords
 Timetoevent
 Metaanalysis
 Methodology
 Survival data
 Clinical trials
 Cochrane database of systematic reviews