Do seasonal-to-decadal climate predictions underestimate the predictability of the real world?

Abstract

Seasonal-to-decadal predictions are inevitably uncertain, depending on the size of the predictable signal relative to unpredictable chaos. Uncertainties can be accounted for using ensemble techniques, permitting quantitative probabilistic forecasts. In a perfect system, each ensemble member would represent a potential realization of the true evolution of the climate system, and the predictable components in models and reality would be equal. However, we show that the predictable component is sometimes lower in models than observations, especially for seasonal forecasts of the North Atlantic Oscillation and multiyear forecasts of North Atlantic temperature and pressure. In these cases the forecasts are underconfident, with each ensemble member containing too much noise. Consequently, most deterministic and probabilistic measures underestimate potential skill and idealized model experiments underestimate predictability. However, skilful and reliable predictions may be achieved using a large ensemble to reduce noise and adjusting the forecast variance through a postprocessing technique proposed here.

Key Points

• Model members can be too noisy and not potential realizations of the real world

• Predictability may be underestimated by idealized experiments and skill measures

• Can achieve skilful and reliable forecasts using large ensembles to reduce noise

1 Introduction

Individual weather events are generally not predictable more than a couple of weeks ahead. This is because the atmosphere is chaotic, so that infinitesimal differences in initial conditions grow over a few days into large-scale disturbances [Lorenz, ¹⁹⁶³]. However, the atmosphere can be influenced by predictable slowly varying factors, leading to a prolonged shift in the climate. For example, in the tropics, sea surface temperature (SST) in the Pacific varies between warm and cool conditions every few years during the El Niño–Southern Oscillation (ENSO), influencing seasonal temperature and rainfall in many regions [e.g., Trenberth and Caron, ²⁰⁰⁰; Alexander et al., ²⁰⁰²; Smith et al., ²⁰¹²]. In the extratropics, North Atlantic SST varies on multidecadal time scales, often referred to as the Atlantic Multidecadal Oscillation or Atlantic Multidecadal Variability (AMV), with associated decadal changes in climate over Europe, America, and Africa [Knight et al., ²⁰⁰⁶; Zhang and Delworth, ²⁰⁰⁶; Sutton and Hodson, ²⁰⁰⁷; Sutton and Dong, ²⁰¹²], the Atlantic storm track position and/or strength [Wilson et al., ²⁰⁰⁹; Woollings et al., ²⁰¹²; Frankignoul et al., ²⁰¹³], and Atlantic hurricane frequency [Goldenberg et al., ²⁰⁰¹; Smith et al., ²⁰¹⁰; Dunstone et al., ²⁰¹¹]. Other drivers of atmospheric variability include external factors such as solar variability, changes in greenhouse gases, changes in aerosols, and internal variability including the Madden Julian Oscillation, sudden stratospheric warmings, and the Indian Ocean dipole [Smith et al., ²⁰¹²].

Seasonal-to-decadal climate predictions aim to predict these drivers and their influence on the atmosphere using coupled general circulation models (GCMs) of the atmosphere, ocean, land, and cryosphere [e.g., Doblas-Reyes et al., ²⁰¹³; Meehl et al., ²⁰¹⁴; Smith et al., ²⁰¹²; Kirtman et al., ²⁰¹³]. They therefore predict changes in climate and the frequency of associated extreme events [Hamilton et al., ²⁰¹²; Eade et al., ²⁰¹²]. Time-averaged predictions may be decomposed into two components: (1) unpredictable noise resulting from the chaotic nature of the atmosphere and (2) a component that is potentially predictable because it is constrained by predictable factors, such as ENSO, AMV, or external forcing. Uncertainties are inevitable in seasonal-to-decadal prediction and depend on the size of the predictable signal relative to unpredictable chaos (signal-to-noise ratio), as well as imperfections in observations used for initial conditions, the fidelity of GCMs, and uncertainties in projected radiative perturbations [e.g., Hawkins and Sutton, ²⁰¹¹]. Ensembles are therefore created to assess uncertainties and enable quantitative probabilistic forecasts to be made [e.g., Palmer et al., ²⁰⁰⁴; Wang et al., ²⁰⁰⁹].

The likely skill of seasonal-to-decadal forecasts is assessed by analyzing tests over a historical period, referred to as hindcasts (forecasts made retrospectively but using only observations that would have been available at the time). The ensemble mean is expected to be more highly correlated with observations than are individual ensemble members because the unpredictable component of the model forecasts is reduced by averaging. However, assessment of other aspects of the ensemble including spread and reliability is also essential [Goddard et al., ²⁰¹²; Corti et al., ²⁰¹²; Ho et al., ²⁰¹³]. Interpretation of the model skill is usually based on the implicit assumption that each model ensemble member represents a potential realization of the true evolution of the climate system that might have occurred due to the chaotic growth of infinitesimal perturbations to the initial conditions.

This approach relies on the predictable components in models and reality being equal. Recent results show that high skill only emerges for seasonal predictions of the winter North Atlantic Oscillation (NAO) [Scaife et al., ²⁰¹⁴; Riddle et al., ²⁰¹³] and multiyear predictions of Atlantic hurricane frequency [Smith et al., ²⁰¹⁰] when taking the mean of a large ensemble. Indeed, Scaife et al. [2014] point out that correlation skill is higher than would be expected from the model signal-to-noise ratio [Kumar, ²⁰⁰⁹] implying that the predictable component in their NAO forecasts is smaller than in reality. Here we provide a more general assessment of the predictable component of seasonal and decadal predictions.

2 Methodology

We define the predictable component as the square root of the fraction of total variance that is predictable and seek to compare the predictable component in observations (PC_obs) to that in model hindcasts (PC_mod). The predictable component in reality is unknown. Previous studies have attempted to diagnose it from the ratio of low- to high-frequency variability [e.g., Boer, ²⁰¹¹] or by assessing potential predictability against a model ensemble member rather than the observations [e.g., Younas and Tang, ²⁰¹³; Boer et al., ²⁰¹³]. However, there is not necessarily a relationship between variability and predictability, and potential predictability is not necessarily related to actual predictability [Kumar et al., ²⁰¹⁴]. Here we estimate PC_obs directly from the fraction of the variance that can be explained by model forecasts, diagnosed from the Pearson correlation (r) between observations (“predictand”) and the ensemble mean of model hindcasts (“predictor”) as r² reflects the proportion of the predictand accounted for by the predictor [e.g., Wilks, ²⁰⁰⁶]. This is a lower bound, since future improvements to models and forecast techniques, and larger ensembles, may yield higher correlations. The predictable component in the model may be estimated from the variance of a large (but finite) ensemble mean relative to the variance of individual ensemble members. This is an upper bound since the variance of the ensemble mean would be reduced if a larger ensemble were available. We define a “ratio of predictable components” (RPC) as

(1)

where σ²_sig is the signal variance of the model ensemble mean in time and σ²_tot is the average variance of individual members. Ideally, the RPC should be equal to one for a forecast system which perfectly reflects the actual predictability.

We assess the RPC in seasonal-to-decadal hindcasts against gridded observations of near-surface temperature (SAT), mean sea level pressure (MSLP), and precipitation (PREC) (details in Table S1 in the supporting information). We analyze seasonal hindcasts from the Met Office Global Seasonal forecasting system 5 (GloSea5) [MacLachlan et al., ²⁰¹⁴] starting around the 1 of November in each year from 1992 to 2011 (20 start dates, 24 members) and a multimodel ensemble of decadal hindcasts from the Met Office decadal prediction system (DePreSys) [Knight et al., ²⁰¹⁴; Smith et al., ^{2010, 2013}], starting from 1 of November in each year 1960 to 2005, and four other models from the Coupled Model Intercomparison Project Phase 5 (CMIP5) [Taylor et al., ²⁰¹²] that had annual start dates available for a similar period to DePreSys (46 start dates, 70 members, see Table S2 for details). We present results from this grand ensemble, but the DePreSys and CMIP5 ensembles were also analyzed separately with similar results (Figure S3).

Modeled and observed fields are expressed as anomalies, using a full-field bias correction method as described in Smith et al. [2013] such that each model is treated in the same way, regardless of initialization method. Modeled and observed trends are removed by linearly detrending, so that correlation is not inflated by the capturing of a climate change signal.

We test the significance of our results using a nonparametric block bootstrap method [Wilks, ²⁰⁰⁶; Smith et al., ²⁰¹³] (see supporting information for further details). We present results without the use of cross validation for model corrections as cross validation leads to an underestimate of correlation [Smith et al., ²⁰¹³; Gangsto et al., ²⁰¹³], but our conclusions are not sensitive to this choice (Figures S2, S4, S5, and S7). We also assess probabilistic performance using reliability diagrams and Brier score [Wilks, ²⁰⁰⁶]. Model probabilities are calculated using the fraction of members that predict the event, with a correction for finite sample size [Wilks, ²⁰⁰⁶]. We present results for the event where the variable in questions is above the median value, as used in Corti et al. [2012]; however, similar results were also found for terciles.

3 Results

Figure 1 shows the RPC for seasonal hindcasts for December to February (lead times 2 to 4 months, Figures 1a–1c) and decadal hindcasts of 4 year means at lead times of 2–5 years (Figures 1d–1f). The RPC is not significantly different to the expected value of one over 70% and 57% of grid points for the seasonal and decadal hindcasts, respectively (averaging over all three variables and ignoring points where no observations are available). However, there are many regions where the RPC is significantly smaller than one, especially on longer time scales. This is indicative of overconfident forecasts in which ensemble members agree well with each other (high signal-to-noise ratio) but do not capture the observed variations (low correlation). Overconfidence in regions where the RPC is significantly smaller than one is confirmed by reliability diagrams (Figure 2a). Ideally, the slope would equal one, such that the observed frequency of occurrence equals the predicted probability. However, the slope is close to zero (red curve in Figure 2a), showing that the likelihood of an event occurring in reality is almost independent of the predicted probability.

Figure 1.

The ratio of predictable components (RPC) for (a–c) seasonal hindcasts of December-January-February (DJF) means and (d–f) decadal hindcasts of 4 year annual means for years 2–5, for near-surface temperature (SAT, column 1), mean sea level pressure (MSLP, column 2), and precipitation (PREC, column 3). Model and observed data are smoothed over regions of 11.25° latitude by 12.5° longitude (15° by 15° for SAT), similar to previous studies [Smith et al., ²⁰¹⁰; Eade et al., ²⁰¹²; Goddard et al., ²⁰¹²]. Stippling identifies where RPC is significantly different to one at the 90% level (see text for details). Regions of negative correlation are masked out as they imply zero skill (see Figure S1 for version with insignificant correlations masked).

Figure 2.

Reliability diagrams of a median threshold event from years 2–5 decadal hindcasts of MSLP, for regions where RPC is (a) significantly lower than one and (b) significantly greater than one (as Figure 1e). The red reliability curves are for bias-corrected model output; red frequency histograms display the percentage of total hindcasts allocated to each bin. The blue reliability curves are for the RPC-corrected model output, with vertical bars showing the 5–95% confidence interval and blue frequency histograms.

Overconfidence of seasonal forecasts is well known, and much research has been undertaken to improve reliability by increasing ensemble spread (see Williams et al. [2013], for a comparison of approaches). Due to our ensemble being finite, this is expected as our r value is an underestimate of reality while our σ²_sig is an overestimate. However, there are also clear regions where the RPC is significantly greater than one. This is indicative of underconfidence, where the ensemble mean agrees relatively well with observations (high correlation) but ensemble members agree less well with each other (low model signal-to-noise ratio). Underconfidence for regions where the RPC is significantly greater than one is confirmed by reliability diagrams in which the slope is greater than one (Figure 2b, red curve). RPC values greater than one suggest that models underestimate the predictability of the real world. This might arise in the multimodel ensemble if some models lack sources of predictability. However, we find very similar RPC patterns (Figure S3) when our multimodel ensemble is split into a single model (HadCM3, 37 members) and the remaining four CMIP5 models (29 members). This suggests that the regions where predictability is underestimated by the multimodel ensemble, which is typically the most accurate forecast [Palmer et al., ²⁰⁰⁴], are also seen in individual models.

Regions where the RPC is greater than one appear to be associated with known variability of the climate system. For example, seasonal MSLP (Figure 1b) shows regions with a high RPC over both Iceland and the Azores, the centers of action of the NAO, with a low RPC in between as would be expected for variability associated with this oscillation. The decadal hindcasts show a high RPC for SAT in the North Atlantic (Figure 1d), which is the region most improved by initialization in decadal predictions [e.g., Doblas-Reyes et al., ²⁰¹³; Smith et al., ²⁰¹⁰] through improved initialization of the Atlantic Meridional Overturning Circulation [Pohlmann et al., ²⁰¹³; Robson et al., ^{2012, 2014}; Yeager et al., ²⁰¹²]. Furthermore, a high RPC for MSLP in the tropical Atlantic (Figure 1e) and PREC in the Sahel (Figure 1f) is consistent with a high RPC for SAT in the North Atlantic, since Atlantic temperatures have been shown to drive variability in these regions [e.g., Zhang and Delworth, ²⁰⁰⁶; Dunstone et al., ²⁰¹¹].

High values of the RPC associated with the NAO and tropical Atlantic MSLP are consistent with skilful predictions of the NAO [Scaife et al., ²⁰¹⁴], Arctic Oscillation [Riddle et al., ²⁰¹³], and Atlantic hurricane frequency [Smith et al., ²⁰¹⁰] obtained with large ensembles. The models underestimate the true predictable component for these variables, and each ensemble member contains too much noise. However, taking the mean of a large ensemble leaves a predictable component that correlates well with reality, more so than would be expected from the original signal-to-noise ratio. This is illustrated in Figures 3a and 3b with time series of the NAO from the seasonal hindcasts and MSLP averaged over the eastern part of the hurricane main development region (MDR; 10–20 N, 60–20 W) from the decadal hindcasts. In both cases the ensemble spread is too large given the high correlations (r = 0.63 for the NAO and 0.71 for MDR MSLP, both significant at 99% level), resulting in a RPC significantly higher than one at the 90% level (RPC = 2.3 for the NAO and 2.4 for MDR MSLP).

Figure 3.

Time series of (a) the North Atlantic Oscillation (NAO) for DJF from seasonal hindcasts and (b) MSLP in the eastern Hurricane Main Development Region (MDR) for years 2–5 from decadal hindcasts. Observations are shown by black curves, and ensemble mean model forecasts as red curves with grey shading showing the ensemble 5–95% range. (c and d) Results after applying the RPC adjustment. The NAO index is calculated as the difference in MSLP between grid points containing Stykkisholmur, Iceland (65°N, 22°W), and Ponta Delgada for the Azores (37°N, 25°W) [e.g., Jones et al., ¹⁹⁹⁷]. The MDR index is calculated as the area-averaged MSLP for region 10–20 N, 60–20 W in which RPC is greater than one.

4 RPC Correction

Model bias is necessarily, and routinely, corrected as a postprocessing step [Stockdale, ¹⁹⁹⁷; International CLIVAR Project Office, ²⁰¹¹]. Likewise, where the RPC is not equal to the desired value of one, we propose that an additional postprocessing adjustment should be applied. Various methods have been proposed to correct overconfidence by increasing ensemble spread [e.g., Raftery et al., ²⁰⁰⁵; Roulston and Smith, ²⁰⁰³], but these do not address the problem of underconfidence highlighted here and in other studies [Scaife et al., ²⁰¹⁴; Ho et al., ²⁰¹³]. Other methods may increase or decrease ensemble spread to improve forecast reliability [e.g., Gneiting et al., ²⁰⁰⁵; Hamill and Colucci, ¹⁹⁹⁷; Eckel and Walters, ¹⁹⁹⁸] but are not specifically designed to adjust the ensemble mean which is required to match the predictable variances in the forecasts and observations. We therefore propose an ensemble adjustment based on linear correction with rescaling [Williams et al., ²⁰¹³], but with parameters diagnosed to ensure that the RPC equals one. This is achieved by computing an adjusted ensemble mean () such that its variance over time (t) is equal to the predictable component of the observed variance (σ²_obs r²):

2

where is the original hindcast ensemble mean value for a given start time t and is the average of these over all start times. We additionally transform each ensemble member (Y_mt′) such that their variance about the new ensemble mean, i.e., the variance of the ensemble noise, becomes equal to the unpredictable variance of the observations, namely, σ²_obs (1 − r²):

(3)

where Y_mt is the original ensemble member value and σ²_noi is the original noise variance:

(4)

This ensures that the total variance of the model is equal to σ²_obs, with the predictable component now accounting for a fraction r² of this.

We illustrate the effect of this RPC correction for the NAO and MDR MSLP time series in Figures 3c and 3d. For the NAO time series, the noise variance is roughly halved while the signal variance is quadrupled (Figure 3c). The MDR noise variance is reduced further, to a quarter of its original value, though the signal variance is only doubled (Figure 3d). The correction does not change correlation but does impact other skill measures. For example, mean squared skill score (MSSS) is moderately increased from 0.27 to 0.37 for the NAO and from 0.49 to 0.51 for MDR MSLP.

This RPC correction can be applied to global fields at the grid point level, excluding points with negative correlation to avoid artificial skill. We apply the correction where the correlation is significantly positive (using a t test at 90% level [Wilks, ²⁰⁰⁶]) and replace the hindcast with climatology elsewhere. This procedure will impact most skill measures (other than correlation), and we illustrate its impact on mean squared skill score (MSSS) [Murphy, ¹⁹⁸⁸] in Figure 4 for decadal hindcasts of MSLP in the North Atlantic (Figure S8 shows the same for seasonal hindcasts, though the results are less significant in this case). Regions of negative MSSS such as South America and the Labrador Sea are improved simply by using climatology (seen by comparing with Figure S6 where insignificant correlations are masked), and the corrected MSSS is not greater than that of climatology. However, MSSS is significantly improved by the RPC correction and is significantly greater than climatology in the central North Atlantic, consistent with the MDR index (Figures 3b and 3d).

Figure 4.

Mean squared skill score (MSSS) for decadal hindcasts of 4 year mean MSLP in the North Atlantic for years 2–5 for (a) the ensemble mean, (b) RPC-corrected ensemble mean, and (c) their difference. Stippling identifies regions where MSSS is significantly greater than for climatology shown in Figures 4a and 4b, or significantly improved by the RPC correction shown in Figure 4c, at the 95% level (see text for details).

The RPC correction leads to some improvement in reliability, illustrated in Figure 2 for year 2–5 MSLP with similar results found for other variables and time scales (not shown). The forecast probabilities are transformed resulting in reliability curves slightly closer to the diagonal (Figure 2, blue curves). In both low and high RPC cases, the uncorrected model displays only moderate sharpness, with probabilities somewhat clustered about that of climatology (red histograms in Figure 2). For regions with a low RPC (overconfidence), and hence generally low correlations (Figure 2a), the corrected hindcasts are clustered even more closely to climatology so to become reliable but with very little sharpness (blue histograms in Figure 2a), and the Brier score has only been improved to equal that of climatology (0.28 reduced to 0.25, where an improved Brier score has a reduced value). For regions with a high RPC (underconfidence), the sharpness has been improved by the RPC correction such that model probabilities are more dispersed from the climatological value and now sample more extreme values (Figure 2b), and the Brier score has been improved (0.20 compared to 0.23 and 0.25 for the original model output and climatology, respectively). However, the reliability curve and frequency histogram may still be suboptimal, with a suggestion of overconfidence remaining for high probabilities (blue curve above the diagonal). On average, for regions where the RPC for decadal hindcasts is significantly greater (smaller) than one, the RPC correction increases (reduces) the signal variance of the ensemble mean by 260% (75%). The variance of individual ensemble members about this mean (i.e., the noise variance) is adjusted such that the total variance equals that of the observations.

5 Discussion and Conclusions

We have quantified the ratio of predictable components (RPC) in observations and models for seasonal and multiyear hindcasts of surface air temperature, mean sea level pressure, and precipitation. The RPC is not significantly different to the expected value of one for around 70% of grid boxes for GloSea5 seasonal hindcasts of DJF (December-January-February) a month ahead and 57% for years 2–5 in a multimodel ensemble of decadal hindcasts. The RPC is significantly smaller than one in many regions, especially for the multiyear hindcasts. RPC smaller than one occurs when model ensemble members agree with each other more than observations. The hindcasts are therefore overconfident. This is well known, and techniques have been developed to reduce overconfidence by increasing the ensemble spread. However, we find regions where the RPC is significantly greater than one in both seasonal and multiyear hindcasts. This appears to be related to known variability including the NAO and multiyear variability of North Atlantic temperatures and associated climate including MSLP in the hurricane development region.

Regions with a RPC greater than one suggest that the real world is more predictable than models. This has important implications:

1. Model ensemble members are generally not potential realizations of the true evolution of the climate system. Instead, each ensemble member contains some predictable signal but with too much noise.

2. Skilful and reliable forecasts may be obtained using a large ensemble. The ensemble mean removes the excessive noise, leaving the predictable component, but variance corrections to the ensemble mean and members are required to ensure that the predictable component in the resultant model forecast is equivalent to that in the observations.

3. Many probabilistic and deterministic skill measures applied to the raw hindcasts underestimate the potential skill.

4. Model-based estimates of predictability diagnosed by taking an individual model run as the truth [e.g., Boer et al., ²⁰¹³; Branstator et al., ²⁰¹²; Collins et al., ²⁰⁰⁶; Dunstone and Smith, ²⁰¹⁰; Tietsche et al., ²⁰¹⁴], which is often seen as the upper limit of predictability, may underestimate the true predictability.

5. Techniques aimed at improving reliability by increasing ensemble spread, including stochastic physics [e.g., Weisheimer et al., ²⁰¹¹], may exacerbate problems where the models are underconfident. Indeed, a recent study [Ho et al., ²⁰¹³] showed that at longer lead times the model spread appears to be too large in many regions. It may ultimately be better to reduce errors rather than increase spread to address model overconfidence.

6. Though not assessed here, it is possible that RPC could be greater than one in climate change projections, in which case conclusions relating to the role of internal variability [e.g., Deser et al., ²⁰¹⁴] may need revising.

Further work is needed to understand why the predictable component is smaller in models than in reality in some regions. One possibility is that the model atmosphere is not constrained strongly enough by the relevant drivers of predictability. Indeed, there is mounting evidence that models respond too weakly to North Atlantic sea surface temperatures. This is shown by direct analysis of model simulations [Mehta et al., ²⁰⁰⁰; Rodwell and Folland, ²⁰⁰²; Gastineau et al., ²⁰¹³] and has also been inferred from a simple theoretical explanation of the lagged response to changes in solar radiation [Scaife et al., ²⁰¹³]. Evidence suggests that the atmospheric response to North Atlantic SSTs may be stronger in higher-resolution models that resolve SST fronts in the Gulf Stream region [Minobe et al., ²⁰⁰⁸]. Future higher-resolution models may therefore yield improved levels of skill for a given ensemble size.

Supporting Information

Readme

Tables S1 and S2 and Figures S1–S8