key: cord-0790823-rr5qhsam
authors: Jung, Sung-mok; Akhmetzhanov, Andrei R.; Hayashi, Katsuma; Linton, Natalie M.; Yang, Yichi; Yuan, Baoyin; Kobayashi, Tetsuro; Kinoshita, Ryo; Nishiura, Hiroshi
title: Real time estimation of the risk of death from novel coronavirus (2019-nCoV) infection: Inference using exported cases
date: 2020-02-02
journal: nan
DOI: 10.1101/2020.01.29.20019547
sha: 2afff5fe89a3e3ec270a41c2b929dfeec68d7d50
doc_id: 790823
cord_uid: rr5qhsam

The exported cases of 2019 novel coronavirus (2019-nCoV) infection who were confirmed in other countries provide a chance to estimate the cumulative incidence and confirmed case fatality risk (cCFR) in China. Knowledge of the cCFR is critical to characterize the severity and understand pandemic potential of 2019-nCoV in the early stage of epidemic. Using the exponential growth rate of the incidence, the present study statistically estimated the cCFR and the basic reproduction number, i.e., the average number of secondary cases generated by a single primary case in a naive population. As of 24 January 2020, with 23 exported cases, and estimating the growth rate from 8 December 2019 (scenario 1) and using the data since growth of exported cases (scenario 2), the cumulative incidence in China was estimated at 5433 cases (95% confidence interval (CI): 3883, 7160) and 17780 cases (95% CI: 9646, 28724), respectively. The latest estimates of the cCFR were 4.6% (95% CI: 3.1-6.6) for scenario 1 and 7.7% (95% CI: 4.9-11.3%) for scenario 2, respectively. The basic reproduction number was estimated to be 2.2 (95% CI: 2.1, 2.3) and 3.7 (95% CI: 3.1, 4.3) for scenarios 1 and 2, respectively. Based on the results, we note that current 2019-nCoV epidemic has a substation potential to cause a pandemic. The proposed approach can provide insights into early risk assessment using only publicly available data.

Since 8 December, 2019, clusters of pneumonia cases of unknown etiology have emerged in Wuhan City, Hubei Province, China [1, 2] . Virological investigation suggests that the causative agent of this pneumonia is a novel coronavirus (COVID-19) [3] . As of 27 January, 2020, a total of 4515 cases including 106 deaths were confirmed [4] . Forty-one cases of COVID-19 infections were also reported outside China, in other Asian countries, the United States, France, Australia, and Canada.

A local market selling seafood and wildlife in Wuhan was visited by many cases in the initial cluster, indicating that a common-source zoonotic exposure may have been the main mode of transmission [5] . However, even after shutting down the market, the number of cases continued to grow across China and several instances of household transmission were reported [6] . It is now speculated that sustained human-to-human transmission aided in the establishment of the epidemic [7] and that reported case counts greatly underestimated the actual number of infections in China [8] .

Early assessment of the severity of infection and transmissibility can help quantify the pandemic potential of COVID-19 and anticipate the likely number of deaths by the end of the epidemic. One important epidemiological measure of severity is case fatality risk (CFR), which can be measured with three different approaches, by estimating (i) the proportion of the cumulative number of deaths out of the cumulative number of cases at a point in time, (ii) the ratio of the cumulative number of deaths to the cumulative number of infected individuals whose clinical outcome is known (i.e., the deceased or recovered), and (iii) the risk of death among confirmed cases, explicitly accounting for the time from illness onset to death [9] . Estimating the CFR using the ratio of deaths to confirmed cases (cCFR), with an adjustment of the time delay from illness onset to death (i.e., method (iii)), can provide insight into severity of the disease, because the naïve CFR based on method (i) tends to be underestimated due to the real-time nature of the growth of fatal cases. For example, during the early stage of an epidemic, failing to right-censor cases with respect to the time delay from illness onset to death may lead to underestimation of the CFR. This is because death due to infection may yet occur following case identification [9] [10] [11] .

When the CFR denominator only includes confirmed cases it is referred to as the cCFR [10] and may overestimate the actual CFR among all infected individuals due to under-ascertainment of infections in the population. Nonetheless, the cCFR is still a valuable measure of the upper bound of the symptomatic CFR (sCFR) among all symptomatic cases, particularly in circumstances of high uncertainty, such as the emergence of a new human pathogen (i.e., .

The basic reproduction number ( 0 ), the average number of secondary cases generated by a single primary case in a fully susceptible population, represents an epidemiological measurement of the transmissibility, helping us to quantify the pandemic potential of COVID-19. Here, we define a pandemic as the worldwide spread of a newly emerged disease, in which the number of simultaneously infected individuals exceeds the capacity for treatment [12] . Using the growth rate of the estimated cumulative incidence from exportation cases and accounting for the time delay from illness onset to death, the present study aims to estimate the cCFR of COVID-19 in real-time.

Information on exported COVID-19 cases who were confirmed in other countries and deaths due to COVID-19 infection in China were retrieved from the first announcement date of the current outbreak (i.e., 31 December, 2019) through 24 January, 2020. The cutoff time of 24 January, 2020 was specifically selected to reflect the governmental ban on public transportation from Wuhan (including flights) which was invoked on 23 January, 2020 [13] . All data were collected either from government websites or from media quoting government announcements. Our data include 51 cases diagnosed outside China who had illness onset through 24 January, 2020 and were reported by 9 February, 2020 and the dates of illness onset and death among 41 deceased cases in China.

The observed incidence i(t) by date of illness onset t is modeled using an exponential growth model with the rate r: ( ) = 0 , where i0 is the expected number of infected cases at time t = 0. The cumulative incidence I(t) is an integral of i(t) over the time interval from zero to t that can be written as ( ) = ∫ ( )d 0 = 0 ( − 1)/ . The cumulative incidence is adjusted to the date of report by the factor u dependent on the parameters of the delay distribution. For the estimation of time delay distribution from illness onset to death, we accounted for right truncation and modeled and used a lognormal distribution with parameters adopted from an earlier study [14] . Let ( ; ) be the lognormal distribution with parameters = { , }. Then, the cumulative incidence I(t) by date of report t can be adjusted to the time from illness onset to death and report by simply multiplying it by the factor ( , ), which is a consequence of the exponentially growing epidemic [10] . The factor u is defined by the following integral: 

Let and be the shape and inverse scale of the gamma distribution modeling the distribution of time interval from illness onset to the report for observed exported cases. The cumulative incidence in China can be then adjusted by the multiplicative factor ̃( , = { , }). The number of exported cases E(t) can then be sampled from another Binomial distribution:

where p describes the probability of finding a traveler from Wuhan among all travelers from China subject to the detection time window of the virus T = 12.5 days [8] . Given that the total volume of inbound passengers from China is M = 5.56 million passengers per year, the fraction of Wuhan travelers is = 0.021%, and the population of Wuhan is n = 11 million, the probability p is given by

First, we fitted the delay distribution of the time from illness onset to report Δ , ( ≤ ) to the gamma distribution. We define the log-likelihood as follows:

( | Δ , ) = ∑ log(gamma(Δ , | shape = , scale = )

that is maximized to find the mean values of the parameters ̃= {̃,̃}, used in the following step. Second, we fitted the observed counts of exported cases and deaths by considering two other likelihoods respectively to each process: 

where and are the times at which the first exportation event and the first death are observed. The total log-likelihood is given by the sum of two log-likelihoods:

which is then maximized to determine the best-fit parameters { , 0 , ( )}. We considered two possible scenarios to fit to the data. In the first scenario (refers to "Scenario 1"), the parameter 0 is fixed at one on the date of illness onset for the first COVID-19 confirmed case (i.e., 8 December, 2019), providing a fixed starting point for the exponential growth of the cumulative incidence. Although 8 December, 2019 is assumed as the starting point for exponential growth of the cumulative incidence in Scenario 1, there is still uncertainty using this date as a proxy for the beginning of exponential growth due to inconsistencies reported on illness onset for the earliest reported cases [2, 5] . Therefore, we conducted a sensitivity analysis where the start date for exponential growth was varied between 1 December and 10 December, 2019. In the second scenario (refers to "Scenario 2"), all parameters { , 0 , ( )} are variable, and calculation begins on the date the first exported case was observed (i.e., 13 January, 2020). For both scenarios, we conducted sensitivity analyses that estimated the growth rate, cumulative incidence, and cCFR value using different cut-off times, values of the detection window T, and sizes of the catchment population of the Wuhan International airport n.

The value of the basic reproduction number 0 for the COVID-19 epidemic was calculated as

where r is the estimated growth rate for each scenario and S is the mean serial interval for successive COVID-19 infections. The value of the serial interval was adopted from a published study [2] , and a gamma distribution was fitted to six known pairs of the infectee and infector (mean = 7.5 days, standard deviation (SD) = 3.4 days).

In our analysis, we employed Markov chain Monte Carlo (MCMC) simulations with eight chains of 4250 samples each, with 2500 samples used for the tuning stage, leveraging the No-U-Turn sampler (NUTS) part of the Python PyMC3 package. However, one of the input variables for the Binomial distribution ̃( , ) ⋅ ( ) or ( , ) ⋅ ( ) was modeled with a continuous variable in our approach. We used the gamma distribution, matching the first two moments to obtain the transformation of the discrete Binomial distribution to its continuous approximation [16] . We avoided joint estimation of all parameters { Σ , , } due to heterogeneity in the aggregated data-a similar issue is discussed in [10] . Instead, we implemented a sequential fitting: first we considered only the likelihood , then we fit the likelihood Σ using the mean values of the estimated parameters obtained in the previous round. Finally, we verified the obtained fit by calculating pointwise estimates using maximum likelihood estimation with confidence intervals derived as profile likelihood-based intervals. We found that both approaches were at complete agreement in their results.

Figures 1A,B show the mean and SD of the time from illness onset to reporting and death, respectively. Employing the gamma distribution, the mean time from illness onset to reporting was estimated at 7.1 days (95% CI: 5.9, 8.4). The mean time from illness onset to death-adopted from a previous study [14] -was estimated at 20.2 days (95% CI: 15.1, 29.5), which led to the lognormal distribution with location parameter of 2.84 and scale parameter of 0.52. . CC-BY 4.0 International license It is made available under a is the author/funder, who has granted medRxiv a license to display the preprint in perpetuity.

The copyright holder for this preprint . https://doi.org/10.1101/2020.01.29.20019547 doi: medRxiv preprint days (95% CI: 6.5, 21.6) employing a lognormal distribution and accounting for right truncation. For reference, the estimate of the mean and its 95% credible intervals without accounting for right truncation are shown in grey. The values for distribution of time from illness onset to death are adopted from an earlier study [14] . The blue bars show empirically observed data collected from governmental reports (as of 24 January, 2020).

Subsequently, the cumulative incidence was estimated from exported case data by fitting an exponentially growing incidence curve for both Scenarios 1 and 2 ( Figure 2 ). As of 24 January, 2020, 20 exported cases were reported, and the cumulative incidence in China was estimated at 6924 cases (95% CI: 4885, 9211) in Scenario 1 and 19,289 cases (95% CI: 10,901, 30,158) in Scenario 2. Table 1 shows the real-time update of the estimated cumulative incidence. The exponential growth rates (r), derived from the growth rate of cumulative incidence, were estimated at 0.15 per day (95% CI: 0.14, 0.15) and 0.29 per day (95% CI: 0.22, 0.36) in Scenarios 1 and 2, respectively. and 2D show the estimated cCFR value accounting for the time delay from illness onset to death under Scenarios 1 and 2, respectively. A total of 41 confirmed deaths were reported as of 24 January, 2020, the cCFR value was estimated at 5.3% (95% CI: 3.5%-7.5%) for Scenario 1 and 8.4% (95% CI: 5.3%-12.3%) for Scenario 2, respectively.

We estimated the basic reproduction number for the COVID-19 infection, using the estimated exponential growth (r) and accounting for possible variations of the mean serial interval (Figure 3 ). Assuming that the mean serial interval was 7.5 days [2] , the basic reproduction number was estimated at 2.10 (95% CI: 2.04, 2.16) and 3.19 (95% CI: 2.66, 3.69) for Scenarios 1 and 2, respectively. However, as the mean serial interval varies, the estimates can range from 1.6 to 2.6 and 2.2 to 4.2 for Scenarios 1 and 2, respectively. To address the uncertainty in the unobserved date of illness onset of the index case in Scenario 1, cCFR was estimated by varying the starting date of the exponential growth in the incidence by placing the single index case between 1 and 10 December, 2019 ( Figure S1 and Table S1 ). When we assumed the date of illness onset of the index case was 1 December, 2019, the estimated incidence in China and the cCFR on 24 January, 2020 were estimated at 4718 (95% CI: 3328, 6278) and 5.3% (95% CI: 3.5, 7.6). The sensitivity analyses for varied cutoff dates between 15 and 24 January, 2020 were conducted. Depending on the number of time points, the estimates of the cumulative incidence and . CC-BY 4.0 International license It is made available under a is the author/funder, who has granted medRxiv a license to display the preprint in perpetuity.

The copyright holder for this preprint . https://doi.org/10.1101/2020.01.29.20019547 doi: medRxiv preprint cCFR have similar values in Scenario 1 ( Figure S2 ) but slightly decreased in Scenario 2, when the cutoff date was earlier ( Figure S3 ). In addition, considering the uncertainty of two fixed parameters (i.e., detection time window and catchment population in Wuhan airport), sensitivity was assessed by varying those parameters. As the catchment population increases and the detection window time decreases, estimated cCFR on 24 January, 2020 gets smaller due to increased number of incidence in China (Table S2 and S3).

The present study estimated the risk of death among confirmed cases while addressing ascertainment bias by using data from cases diagnosed outside mainland China and a right-censored likelihood for modeling the count of deceased cases. The estimated cCFR value was 5.3% [3.5, 7.5] when the date of illness onset for the index case was fixed a priori at 8 December, 2019 (Scenario 1), and 8.4% [5.3, 12 .3%] when the timing of the exponential growth of epidemic was fitted to data alongside with other model parameters (Scenario 2). The estimated value of u in Scenario 2, which adjusts the time delay from illness onset to death, was smaller than that in Scenario 1 due to the larger value of the growth rate of incidence, while the estimated cCFR value as of 24 January, 2020 was larger in Scenario 2 compared with Scenario 1. Depending on the available data, the estimate of the cCFR (i.e., the delay-adjusted risk of death among confirmed cases) may vary in time and show some fluctuations. In addition to that, we estimated the value of the basic reproduction number 0 in the range of 1.6-2.6 for Scenario 1 and 2.2-4.2 for Scenario 2. From either estimate, we conclude that COVID-19 has substantial potential to spread via human-to-human transmission. However, R0 > 1 does not guarantee that a single exported (and untraced) case would immediately lead to a major epidemic in the destination country as government responses such as border control, isolation of suspected cases, and intensive surveillance should serve to reduce opportunities for transmission to occur [17] [18] [19] .

Our cCFR estimates of 5.3% and 8.4% indicate that the severity of COVID-19 is not as high as that of other diseases caused by coronaviruses, including severe acute respiratory syndrome (SARS), which had an estimated CFR of 17% in Hong Kong [9, 10, 20] , and Middle East respiratory syndrome, which had an estimated CFR of 20% in South Korea [21] . Nonetheless, considering the overall magnitude of the ongoing epidemic, a 5%-8% risk of death is by no means insignificant. In addition to quantifying the overall risk of death, future research must identify groups at risk of death (e.g., the elderly and people with underlying comorbidities) [22, 23] . Moreover, considering that about 9% of all infected individuals are ascertained and reported [24] , the infection fatality risk (IFR), i.e., the risk of death among all infected individuals, would be on the order of 0.5% to 0.8%.

Our estimated 0 range of 1.6-4.2 for COVID-19 is consistent with other preliminary estimates posted on public domains [25-28], and is comparable to the 0 of SARS, which was in the range of 2-5 during the 2003 outbreak in Singapore [15] . Between our two estimates, the latter scenario yielded a greater value than the former, and there was an increasingly improved ascertainment in early January 2020. The virus was identified and sequenced on 7 January, 2020 and subsequently, the primer was widely distributed, allowing for rapid laboratory identification of cases and contributing to a time-dependent increase in the number of confirmed cases out of China. Consequently, Scenario 2, which was fully dependent on the growth rate of exported cases, could have overestimated the intrinsic growth rate of cases. Considering the estimated value of 0 , the possibility of presymptomatic transmission in the ongoing epidemic is a critical question, as it would have a substantial impact on public health response to the epidemic (e.g., whether the contact tracing should be prioritized or not) as well as overall predictability of the epidemic during the containment stage [29, 30] .

From the technical side, it should be emphasized that our proposed approach can be especially useful during the early stage of an epidemic when local surveillance is affected by substantial ascertainment bias and export and death data are available and better ascertained. Nonetheless, caution must be used when implementing similar estimations for the COVID-19 epidemic, as all flights from Wuhan airport were grounded as of 23 January, 2020 [13] and this intervention abruptly changed the human migration network. Despite the decrease in the outbound flow of travelers from Wuhan, there is a substantial risk that the next epidemic wave will originate from other cities.

There are five main limitations in the present study. First, our results present an estimate for the cCFR which only addresses fatality among confirmed cases. More precise IFR estimate that includes infected individuals other than confirmed cases can only be estimated using additional pieces of data (e.g., seroepidemiological data or outpatient clinic visits). It should be noted that not only the denominator but also numerator values are also subject for better estimation (e.g. excess mortality estimate). Second, our study relied on limited empirical data that were extracted from publicly available data sources. Thus, future studies with greater sample size and precision are needed. Nonetheless, we believe that this study will improve the situational assessment of the ongoing epidemic. Third, our assumed date of illness onset for the index case in Scenario 1 is based on initial reports of the earliest onset date for a case, and the continued exponential growth with the rate r is the authors' extrapolation. However, we conducted a sensitivity analysis and ensured that the resulting statistical estimates would not greatly vary from our main results. Fourth, there is an uncertainty in the detection window time T. Since the epidemiological investigations are being actively implemented outside China, we believe that the sum of the incubation period and the infectious period can be a plausible estimation of the detection time window. Fifth, heterogeneous aspects of death (e.g. age and risk groups) need to be addressed in the future studies.

In conclusion, the present study has estimated cCFR to be on the order of 5%-8% and R0 to be 1.6-4.2, endorsing the notion that COVID-19 infection in the ongoing epidemic possesses the potential to become a pandemic. The proposed approach can also help direct risk assessment in other settings with the use of publicly available datasets.

The following are available online at www.mdpi.com/xxx/s1, Figure S1 : Sensitivity analysis with varying the start point of exponential growth in cumulative incidence from 1 to 10 December, 2019, Figure S2 : Sensitivity analysis with varying the cutoff date from 15 to 24 January, 2020 in estimation Scenario 1, Figure S3 : Sensitivity analysis with varying the cutoff date from 15 to 24 January, 2020 in estimation Scenario 2, Table S1 : Sensitivity analysis with varying the start point of exponential growth in cumulative incidence from 1 to 10 December, 2019 in estimation Scenario 1, Table S2 : Sensitivity analysis with varying the catchment population size in Wuhan airport, Table S3 . CC-BY 4.0 International license It is made available under a is the author/funder, who has granted medRxiv a license to display the preprint in perpetuity.

The copyright holder for this preprint . https://doi.org/10.1101/2020.01.29.20019547 doi: medRxiv preprint B. Sensitivity analyses by varying the data cutoff date from 15 January to 24 January, 2020

As the present study relies on real-time data, the estimates (i.e., growth rate, cumulative incidence, and cCFR) using different number data cutoff dates are shown in Figure S2 (Scenario 1) and S3 (Scenario 2). Figure S2 . Sensitivity analysis in Scenario 1, varying the data cutoff date between 15 and 24 January 2020. The estimated values of the A) exponential growth rate, B) cumulative incidence, and C) confirmed case fatality risk depend on the data cutoff date (i.e., end point of each estimation). Each line and shaded area indicate the estimate and its 95% confidence interval.

. CC-BY 4.0 International license It is made available under a is the author/funder, who has granted medRxiv a license to display the preprint in perpetuity.

(which was not peer-reviewed) The copyright holder for this preprint . Figure S3 . Sensitivity analysis in Scenario 2, varying the data cutoff date between 15 and 24 January 2020. The estimated values of the A) exponential growth rate, B) cumulative incidence, and C) confirmed case fatality risk depend on the data cutoff date (i.e., end point of each estimation). Each line and shaded area indicate the estimate and its 95% confidence interval.

. CC-BY 4.0 International license It is made available under a is the author/funder, who has granted medRxiv a license to display the preprint in perpetuity.

(which was not peer-reviewed) The copyright holder for this preprint . https://doi.org/10.1101/2020.01.29.20019547 doi: medRxiv preprint

In the present study, the detection time window was fixed at 12.5 days, based assumed values for the incubation and infectious periods. The catchment population in Wuhan airport was also fixed at 11 million. As part of the sensitivity analysis, Table S2 shows the estimated growth rate, cumulative incidence and cCFR by varying only detection time window, while Table S3 shows the estimates with different catchment population sizes and a fixed detection time window for Scenarios 1 and 2, respectively. 

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CI, confidence interval (the 95% CI was derived from Monte Carlo Markov Chain method

CI, confidence interval (the 95% CI was derived from the Markov chain Monte Carlo method)

A. Sensitivity analysis by varying the starting point of exponential growth of cumulative novel coronavirus incidence from 1 December to 10 December, 2019The starting point of exponential growth of cumulative case incidence was fixed as the illness onset date of the index case (i.e, 8 December 2019) in Scenario 1, while it was allowed to vary from the day the first exportation case was observed (i.e., 13 January 2020) in Scenario 2. However, since there is uncertainty regarding the actual starting point of exponential growth in Scenario 1 due to discrepancies in the date of illness onset for the first reported case, a sensitivity analysis was conducted by varying this date from 1 December 2019 to 10 December 2019. Table S1 and Figure S1 show the estimated growth rate, cumulative incidence, and cCFR that result from this analysis. Table S1 . Sensitivity analysis with varying the start point of exponential growth in cumulative incidence from 1-10 December, 2019 in estimation Scenario 1.

Exponential growth rate (95% CI) Estimated incidence in China on 24 January (95% CI) cCFR on 24 January (95% CI)