BACKGROUND
The yearly global impact of seasonal influenza epidemics has been estimated in about 1 billion symptomatic cases, 3–5 million severe cases and 250–500 thousands of deaths [1]. The duration, severity and geographical spread of influenza activity vary widely from one season to another depending on several factors such as rapid mutating viral strains, sensitivity of the population or climatic factors [Reference Thompson2, Reference Monto3]. Early detection of the start of seasonal epidemics is needed to inform public health authorities in order to implement necessary control measures. Moreover, monitoring influenza epidemics allows analysis about changes in trends, estimation of the global impact on populations and year-to-year comparisons.
The dynamics of influenza epidemics in the general population is monitored using primary care data collected by surveillance networks of health-care professionals who report the number of influenza-like illness (ILI) cases seen among their patients following a specific definition [Reference Ortiz4]. However, only a portion of ILI cases are due to influenza virus infection [Reference Carrat5], thus statistical methods have to be used to determine the influenza epidemic onset from this non-specific data.
A wide variety of statistical methods have been proposed to detect seasonal influenza epidemics based on ILI incidence time series [Reference Cowling6] such as regression models [Reference Serfling7, Reference Wang8], hidden Markov models (HMMs) [Reference Le Strat and Carrat9] and more recently the moving epidemic method (MEM) [Reference Vega10]. However, the evaluation of these methods is hindered by the absence of a gold standard regarding true influenza epidemic periods [Reference Cowling6]. Performances of these methods have often been evaluated based from the results of other detection methods [Reference Cowling6], using standard epidemiological metrics as sensitivity, specificity, positive predictive value, etc. [Reference Cowling6, Reference Unkel11–Reference Choi13] with different definitions [Reference Kleinman and Abrams14].
An accurate detection method would be able to detect precisely, i.e. with the smallest detection time, each season the whole single epidemic period and particularly the start, which allowed alerting public health authorities and population.
In France, gold standard for seasonal influenza epidemic periods has been previously determined based on an international expert's consensus using the Delphi method [Reference Debin15]. This allowed identifying the start and end of epidemics using estimated ILI incidences and virological data in primary care.
We propose here to evaluate some common epidemic statistical detection methods by comparing their results to the gold standard determined by this expert consensus [Reference Debin15]. We defined performance metrics according to the monitoring objectives to seek for a global view of the detection methods properties.
METHODS
Influenza surveillance data
ILI incidence rates were obtained from the Sentinelles network, a nationwide epidemiological surveillance system based on voluntary general practitioners (GPs) in France [Reference Flahault16, 17]. Sentinel GPs reported on a weekly basis the number of ILI cases seen among their patients, using the following definition: ‘sudden onset of fever >39 °C (102 °F) with respiratory signs and myalgia’, allowing estimation of weekly ILI incidence rates [Reference Turbelin18, Reference Souty19].
Gold standard for influenza epidemic periods
The gold standard for influenza epidemic periods in France was determined by a Delphi method described in Debin et al. [Reference Debin15]. More precisely, 57 experts determined yearly influenza epidemic periods from 1985/86 to 2010/11 using a web interface. For each season, virological results and estimated ILI incidence rates (from Sentinelles network) were presented; the experts were asked to determine the beginning and ending dates of each epidemic. In a second round, the same data were presented; adding histograms with the distribution of responses for start and end dates given by all experts on the previous round. A third final round was proposed for seasons when at least 25% of experts changed their responses between the first and the second rounds. The consensus of the start and end dates for each season was then determined by the mode of the response, after removal of 5% of extreme responses on each side. Results for seasons between 1995/96 and 2010/11 are presented in Figure 1 along with estimates of ILI incidence rates from the Sentinelles network.
Estimates of influenza-like illness incidence rates and gold standard for epidemic periods determined by an expert consensus during the 16 influenza seasons between 1995/96 and 2010/11, Sentinelles network, France.
Epidemic detection methods
Four detection methods were evaluated: a periodic regression [Reference Serfling7, Reference Pelat20], a robust regression [Reference Wang8], the MEM [Reference Vega10] and a HMM [Reference Le Strat and Carrat9]. For each method, several values for the tuning parameters were chosen for calibration (Table 1). The common parameter of these four methods is the length of the learning period involved: the number of past observations (or past seasons for the MEM) provided to perform detection at a given point of time (called ‘learning size’ further). For this parameter, we tested four values: 3, 5, 10 years and the whole available historical data at each time point.
Methods and parameter combinations used for detectors parameterisation
* Considering all available historical data at each point.
The periodic regression for epidemic detection is a widely used approach from Serfling's work on influenza [Reference Serfling7]. To sum up, it is based on a regression model which fits non-epidemic data to predict non-epidemic baseline. The epidemic threshold is defined by an upper percentile of the prediction distribution (here the 90th percentile [Reference Costagliola21]). In our evaluation, to prune the data, we removed values in the learning period over a given value (cut-off) that was either a fixed value or one determined from the learning data using a given percentile. To fit the model, we used the following regression equation:
$$\eqalign{I_{\rm t} &= \alpha _0 + \alpha _1 t + \alpha _2 \cos \left( {\displaystyle{{2\pi {\kern 1pt} t} \over {52 \cdot 17}}} \right) + \alpha _3 \cos \left( {\displaystyle{{2\pi t} \over {52 \cdot 17}}} \right)\cr & \quad + \varepsilon _{\rm t},} $$
where I t is the incidence on week t, t being the week index and ε t the residuals.
The robust regression is an alternative to periodic regression described above where all time series is considered. Data pruning is done by assigning less weight to outliers, computed by a dedicated estimator [Reference Huber22, Reference Fox23]. We used the same regression equation and the same definition of epidemic threshold as for periodic regression described above.
The MEM is implemented by steps: epidemic periods are first determined using historical time series, then epidemic thresholds are calculated using epidemic periods defined [Reference Vega10]. An extra parameter δ must be specified, corresponding to the minimum increment percentage used to find the optimum epidemic duration [Reference Vega10].
HMMs were also used for monitoring such time series. A two-state HMM is applied on incidence time series, assuming that these observations are generated from a mixture of Gaussian distributions [Reference Le Strat and Carrat9].
In what follows, we will call ‘detector’ a method used with a given set of fixed parameters. Each detector was applied on ILI incidence rate time series to detect epidemic period in a prospective way (e.g. as it would be applied in real time). Each week, the detector is run only on the data available up to this week.
Epidemic periods generated by detectors are triggered as soon as two consecutive weekly ILI incidences were above the threshold [Reference Pelat20, Reference Viboud24, Reference Costagliola25] (or classified in ‘alarm state’ for HMM). Moreover, two consecutive ILI observations below the threshold (or classified as ‘no alarm state’ for HMM) were required to determine the end of an epidemic. All weeks inside an epidemic period are classified ‘on alert’ for the detector.
Detection performance
The performance evaluation of each detector was carried out on the period 1995–2011, which included 15 seasonal influenza epidemics and the 2009/10 pandemic. Evaluating the performance of a detector required calculating a number of measures that are in keeping with the objectives of detection. We computed two sets of metrics: (1) ‘weekly detection’ metrics, which are based on weekly alerts determined by the detector, and (2) ‘epidemic period detection’ metrics, which are focused on detecting the epidemic period as a whole.
For both approaches, we assumed that the true state (epidemic or non-epidemic) of a given week was informed by our gold standard – states were called true and false. States in the evaluated detector (‘on alert’ or ‘without alert’) were called positive and negative.
Weekly detection metrics
We defined the number of weeks correctly classified by the detector as ‘true positive’ (TP), respectively incorrectly classified as ‘false positive’ (FP); the correctly classified non-epidemic as ‘true negative’ (TN), respectively the incorrectly as the ‘false negative’ (FN).
Evaluation measures were then defined as:
$${\bf Sensitivity} = \hbox{TP/(TP+FN)}$$
$${\bf Specificity}= \hbox{TN/(TN+FP)}.$$
We also defined positive predictive value as PPV = TP/(TP + FP) and negative predictive value as NPV = TN/(TN + FN).
These metrics were computed for the whole evaluated period: from ISO week 26 of 1995 to ISO week 25 of 2011, being 835 weeks.
Epidemic period detection metrics
This second evaluation approach focused on the ability of the detector in identifying the start week of each epidemic, and gives less importance to the correct detection of subsequent epidemic weeks. It stems from the reality that, for the management of seasonal influenza epidemics, public health authorities need accurate and timely information about the epidemic start, less so about the epidemic state of each subsequent week as the epidemic unfolds time period detection [Reference Tsui26].
As proposed by Tsui et al. [Reference Tsui26], we defined for each epidemic a ‘target’ window that consists of the epidemic starting week and its two adjacent weeks (one before and one after) from the gold standard. Then, we considered that a detector correctly detected the start of the epidemic if the start of the first epidemic period detected during the season is in this target window. The associated evaluation metric, called detectedstart, was defined as the proportion of epidemic starts correctly detected.
We also defined the timeliness as the mean number of weeks between the first epidemic week in the gold standard and the beginning of the first epidemic period identified by the detector for all the epidemics studied.
Finally, we defined multipledetect which is the number of seasons where the detector identified more than one epidemic period.
Metric comparisons between detectors
The most desirable detector might detect only one epidemic period per season [Reference Viboud24], have maximum detectedstart, sensitivity and specificity values, and timeliness close to zero [Reference Cowling6, Reference Wang8, Reference Martinez-Beneito27]. We prioritised metrics in the evaluation, whenever possible; we selected (1) multipledetect equal to zero, (2) detectedstart maximal and (3) compromise between high sensitivity and high specificity.
For periodic regression and MEM, the impact of the parameter values on the detection performance was studied using linear regression.
Uncertainty about the metric point estimates was assessed by bootstrapping. The 16 influenza seasons included in the evaluation were resampled with replacement (N = 1000). The bootstrap distributions obtained for each metric allowed estimation of 95% confidence intervals using the 2·5% and 97·5% percentiles. Then, we used paired Student's t test to compare bootstrap metric values between detectors.
RESULTS
All the 304 detectors studied (184 periodic regression models, 112 MEM, 4 HMM and four robust regression models) detected at least one epidemic period during each of the 16 studied influenza seasons (from 1995/96 to 2010/11).
Link between metrics
Link between detectedstart and specificity, sensitivity or timeliness is a bell-shaped trajectory, with maximal values of detectedstart for sensitivity between 0·80 and 0·94, specificity between 0·96 and 0·99 and timeliness between −1·3 and +0·3 weeks (Fig. 2). Detectedstart was maximal when multipledetect was minimal. Multipledetect was equal to zero for a high specificity and a moderate to high sensitivity (under 0·954). When timeliness was close to zero (between −0·5 and 0·5), sensitivity was between 0·69 and 0·92 and specificity between 0·97 and 0·99.
Metric pairwise comparisons for all detectors implemented (n = 280), influenza epidemics from 1995/96 to 2010/11, France.
Intra-evaluation – by method
Periodic regression
Over the 184 detectors evaluated using periodic regression method, the prune parameter was the most influential on detection metrics. Increasing the pruning level made the sensitivity and multipledetect decreased; the specificity and timeliness increased. Regarding the detectedstart metric, the relation with the prune parameter was no linear. Indeed, detectedstart was minimal (0·375) for extreme values of prune parameters and maximal (0·875) for 36 detectors (cut-off between 160 and 250, percentile between 0·84 and 0·87).
Among the 184 detectors studied, eight achieved the highest detectedstart value (0·875) and did not detect several epidemics within the same season. Among them, the detector with the highest values for sensitivity and specificity was parametrised with a percentile of 0·86 and a maximum learning size period (sensitivity = 0·874, specificity = 0·985, timeliness = −0·2 weeks, PPV = 0·958 and NPV = 0·962).
Robust regression
For robust regression method, among the four detectors compared, metrics were often better when the learning size included all available historic data, except for sensitivity for which a 10 years learning size lead to a slightly higher value (0·80 vs. 0·79). All detectors have at least one multiple epidemic detection within a season (seasons 1995/96 and 2000/01).
Robust regression method parameterised with the largest learning size achieved a detectedstart equal to 0·750, timeliness to 0·1 weeks, sensitivity to 0·791 and specificity to 0·985. During the 2000/01 epidemic, the detector identified two epidemic periods: a first of 2 weeks between weeks 50 and 52, year 2000 and a second between weeks 3 and 7, year 2001. For this detector, PPV was 0·959 and NPV was 0·941.
Moving epidemic method
With the MEM, both the δ parameter and the learning size affected metric values, excepted multipledetect. Increasing the δ value led to lower detectedstart, sensitivity and higher timeliness and specificity. Conversely, higher learning size lead to higher specificity, timeliness and lower sensitivity and detectedstart.
Twelve detectors achieved a maximal detectedstart (0·875) with no multiple epidemics detected within the same season. These detectors were parameterised with a maximal learning size and a δ value between 1·5 and 1·9, or a learning size equal to 10 years and a δ value between 2·2 and 2·8. Among these detectors, timeliness was close to zero (between −0·3 and 0 weeks). Sensitivity was more variable (0·83–0·92) than specificity (0·98–0·99). The best comprise was MEM parameterised with δ value of 1·5 and whole learning period (sensitivity = 0·919, specificity = 0·976, timeliness = −0·3 weeks, PPV = 0·926, NPV = 0·976).
Hidden Markov model
Among the four detectors parameterised with HMM method, larger learning size led to higher sensitivity, lower specificity and detectedstart. Only one detector had the best detectedstart value (0·500) with no multiple season detections. It was parameterised with a learning size fixed to 3 years (sensitivity = 0·946, specificity = 0·914, timeliness = −1·1 weeks, PPV = 0·791, NPV = 0·983).
Inter-evaluation: comparison between methods
Among the four detectors identified in the intra-evaluation (Table 2), only robust regression had a multiple epidemic detection (two epidemic periods detected during influenza season 2000/01). Compared with HMM, detectedstart values were higher for MEM and periodic regression method (P < 1 × 10−6). Considering these two detectors, we did not highlight differences for detectedstart (P = 0·77), but the periodic regression led to higher specificity and lower sensitivity than the MEM (P < 1 × 10−6).
Metric values and 95% confidence intervals for the best detector identified for each method tested, influenza epidemics from 1995/96 to 2010/11, France
PPV, positive predictive value; NPV, negative predictive value.
* Two epidemic periods were detected during the 2000/01 season (2000w50 to 2000w52 and 2001w03 to 2001w07).
DISCUSSION
We compared performances of several epidemic detection methods and parameterisations for real-time influenza surveillance based on a gold standard determined by an expert consensus [Reference Debin15]. Performance metrics defined here allowed identification of methods able to detect accurately the start of the single epidemic period for each influenza season. The final choice of exact statistical method parameterisation depends on the wishes of public health authorities in terms of sensitivity and specificity especially.
Although statistical measures of performances of a classification function are enough consensual – such as sensitivity and specificity, in the case of the detection method evaluation based on a repeated classification function over time – the definition of these measures is less clear [Reference Kleinman and Abrams14]. Cowling et al. [Reference Cowling6] proposed a definition of sensitivity ‘whether there was at least one alarm during the peak season’, allowing a sensitivity equal to 1 for methods which were able to detect, for example, only the peak of the epidemic. Moreover, the specificity defined by Cowling et al. [Reference Cowling6] involved values which are dependant of the epidemic duration. ROC curves, combining sensitivity and specificity, were sometimes used to compare detection methods [Reference Spreco and Timpka28], but they ignored the detection timeliness, which is of paramount importance in practice. We feel that a metric such as detectedstart addresses best what is expected in practice from an epidemic detection method: identifying the epidemic start ‘not too early and not too late’, as the detection of the epidemic ending is a lesser issue for real-time surveillance. Moreover, as influenza epidemics occur once a year in temperate areas [Reference Viboud24], a second metric was used (multipledetect) to ensure ability of the method to detect only one epidemic period during the season (from September year n to August year n + 1).
In addition to new metrics defined here, standard metrics commonly used, such as sensitivity and specificity [Reference Cowling6, Reference Pelat20], were also computed. The link between detectedstart and these two metrics was non-linear allowing the selection of detectors achieving a compromise between high values of sensitivity and high values of specificity. In addition, by definition, high detectedstart values lead to timeliness close to zero. Indeed, detectedstart allowed in one metric, identification of the most desirable method, with high sensitivity and specificity, and timeliness close to zero [Reference Wang8, Reference Martinez-Beneito27].
The choice of the best method depends on the details of the application, implementation and context of the surveillance [Reference Unkel11]. Among all methods studied here, we observed that HMM is more sensitive and less specific; conversely, robust periodic regression is more specific and less sensitive in comparison to other detectors studied. MEM and periodic regression are more able to be parametrised (δ, cut-off, learning size) involving a more difficult choice for implementation which requires us to test a large number of detectors compared to the two other methods. Overall, we observed that the consideration of all the historical data led to better metric values.
Epidemic detection methods were applied to ILI incidence time series. According to the chosen ILI definition, specificity for influenza could vary [Reference Carrat5] as other respiratory pathogens, which also circulating during autumn and winter, can cause very similar illness [Reference Fleming and Cross29]. Virological confirmation of these ILI cases allows estimating the real number of influenza symptomatic cases and would tend to improve epidemic detection. However, laboratory surveillance is not always part of routine surveillance. When data are available, reporting delay is observed and methods, practices and sample size may vary by country [Reference Vega10]. This suggests that detection methods based on clinical data could be a more practical choice. However, when proper virological data are collected along with clinical cases, it should be taken into account to confirm that increasing incidence is largely due to influenza viruses.
Our study was limited by the statistical methods for influenza epidemic detection compared here. All methods are based only on ILI incidence time series. Assimilation of laboratory-confirmed influenza surveillance data and ILI time series in a same detection method may improve performance. However, definition of ILI used by the French Sentinelles network is very specific [Reference Carrat5], allowing estimation of ILI incidence close to influenza-confirmed incidence. Moreover, the methods did not consider spatial information that is often available in influenza surveillance, such as ILI incidence by region. The incorporation of spatial data in statistical models holds the promise of improved sensitivity, timeliness of detection and possibly specificity [Reference Kleinman and Abrams14]. Finally, we did not explore voting algorithm which could combine several detectors.
The metrics presented here allowed to measure ability of statistical epidemic detection methods to detect precisely the beginning of the single epidemic period by year with the smallest detection time. Their implementation on ILI incidence data from primary care surveillance network could improve influenza surveillance by providing accurate epidemic alerts for public health authorities and population.
ACKNOWLEDGEMENTS
The authors thank the general practitioners of the Sentinelles network for their participation.
DECLARATION OF INTEREST
None.
