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- Editorial & advisory board
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**Research article**
30 Apr 2019

**Research article** | 30 Apr 2019

Calibration of a multi-physics ensemble

^{1}Department of Meteorology and Atmospheric Science, Pennsylvania State University, University Park, PA, USA^{2}CEREA, joint laboratory École des Ponts ParisTech and EDF R&D, Université Paris-Est, Champs-sur-Marne, France^{a}now at: Scripps Institution of Oceanography, University of California, San Diego, CA, USA

^{1}Department of Meteorology and Atmospheric Science, Pennsylvania State University, University Park, PA, USA^{2}CEREA, joint laboratory École des Ponts ParisTech and EDF R&D, Université Paris-Est, Champs-sur-Marne, France^{a}now at: Scripps Institution of Oceanography, University of California, San Diego, CA, USA

Abstract

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Atmospheric inversions have been used to assess biosphere–atmosphere
CO_{2} surface exchanges at various scales, but variability among
inverse flux estimates remains significant, especially at continental scales.
Atmospheric transport errors are one of the main contributors to this
variability. To characterize transport errors and their spatiotemporal
structures, we present an objective method to generate a calibrated ensemble
adjusted with meteorological measurements collected across a region, here the
upper US Midwest in midsummer. Using multiple model configurations of the
Weather Research and Forecasting (WRF) model, we show that a reduced number
of simulations (less than 10 members) reproduces the transport error
characteristics of a 45-member ensemble while minimizing the size of the
ensemble. The large ensemble of 45 members was constructed using different
physics parameterization (i.e., land surface models (LSMs), planetary
boundary layer (PBL) schemes, cumulus parameterizations and microphysics
parameterizations) and meteorological initial/boundary conditions. All the
different models were coupled to CO_{2} fluxes and lateral boundary
conditions from CarbonTracker to simulate CO_{2} mole fractions.
Observed meteorological variables critical to inverse flux estimates, PBL
wind speed, PBL wind direction and PBL height are used to calibrate our
ensemble over the region. Two optimization techniques (i.e., simulated
annealing and a genetic algorithm) are used for the selection of the optimal
ensemble using the flatness of the rank histograms as the main criterion. We
also choose model configurations that minimize the systematic errors (i.e.,
monthly biases) in the ensemble. We evaluate the impact of transport errors
on atmospheric CO_{2} mole fraction to represent up to 40 % of the
model–data mismatch (fraction of the total variance). We conclude that a
carefully chosen subset of the physics ensemble can represent the
uncertainties in the full ensemble, and that transport ensembles calibrated
with relevant meteorological variables provide a promising path forward for
improving the treatment of transport uncertainties in atmospheric inverse
flux estimates.

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Díaz-Isaac, L. I., Lauvaux, T., Bocquet, M., and Davis, K. J.: Calibration of a multi-physics ensemble for estimating the uncertainty of a greenhouse gas atmospheric transport model, Atmos. Chem. Phys., 19, 5695-5718, https://doi.org/10.5194/acp-19-5695-2019, 2019.

1 Introduction

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Atmospheric inversions are used to assess the exchange of CO_{2} between
the biosphere and the atmosphere (e.g., Gurney et al., 2002; Baker et al.,
2006; Peylin et al., 2013). The atmospheric inversion or “top-down” method
combines a prior distribution of surface fluxes with a transport model to
simulate CO_{2} mole fractions and adjust the fluxes to be optimally
consistent with the observations (Enting, 1993). Large uncertainty and
variability often exist among inverse flux estimates (e.g., Gurney et al.,
2002; Sarmiento et al., 2010; Peylin et al., 2013; Schuh et al., 2013).
These posterior flux uncertainties arise from varying spatial resolution,
limited atmospheric data density (Gurney et al., 2002), uncertain prior
fluxes (Corbin et al., 2010; Gourdji et al., 2010; Huntzinger et al., 2012)
and uncertainties in atmospheric transport (Stephens et al., 2007; Gerbig et
al., 2008; Pickett-Heaps et al., 2011; Díaz Isaac et al., 2014; Lauvaux
and Davis, 2014).

Atmospheric inversions based on Bayesian inference depend on the prior flux
error covariance matrix and the observation error covariance matrix. The
prior flux error covariance matrix represents the statistics of the mismatch
between the true fluxes and the prior fluxes, but the limited density of
flux observation limits our ability to characterize these errors (Hilton et
al., 2013). The observation error covariance describes errors of both
measurements and the atmospheric transport model. In atmospheric inversions,
the model errors tend to be much greater than the measurement errors (e.g.,
Gerbig et al., 2003; Law et al., 2008). Additionally, atmospheric inversions
assume that the atmospheric transport uncertainties are known and are
unbiased; therefore, the method propagates uncertain and potentially biased
atmospheric transport model errors to inverse fluxes limiting their
optimality. Unfortunately, rigorous assessments of the transport
uncertainties within current atmospheric inversions are limited. Estimation
of the atmospheric transport errors and their impact on CO_{2} fluxes
remains a challenge (Lauvaux et al., 2009).

A limited number of studies are dedicated to quantify the uncertainty in
atmospheric transport models and even fewer attempted to translate this
information into the impact on the CO_{2} mixing ratio and inverse
fluxes. The atmospheric Tracer Transport Model Intercomparison Project
(TransCom) has been dedicated to evaluate the impact of atmospheric transport
models in atmospheric inversion systems (e.g., Gurney et al., 2002; Law et
al., 2008; Peylin et al., 2013). These experiments have also shown the
importance of the transport model resolution to avoid any misrepresentation
of high-frequency atmospheric signals (Law et al., 2008). Diaz Isaac et
al. (2014) showed how two transport models with two different resolutions and
physics but using the same surface fluxes can lead to large model–data
differences in the atmospheric CO_{2} mole fractions. These
differences would yield significant errors on the inverse fluxes if
propagated into the inverse problem. Errors in horizontal wind (Lin and
Gerbig, 2005) and in vertical transport (Stephens et al., 2007; Gerbig et
al., 2008; Kretschmer et al., 2012) have been shown to be important
contributors to uncertainties in simulated atmospheric CO_{2}. Lin and
Gerbig (2005), for example, estimate the impact of horizontal wind error on
CO_{2} mole fractions and conclude that uncertainties in CO_{2}
due to advection errors can be as large as 6 ppm. Other studies have shown
that errors in the simulation of vertical mixing have a large impact on
simulated CO_{2} and inverse flux estimates (e.g., Denning et al.,
1995; Stephens et al., 2007; Gerbig et al., 2008). Therefore, some studies
have evaluated the effects that planetary boundary layer height (PBLH) has on
CO_{2} mole fractions (Gerbig et al., 2008; Williams et al., 2011;
Kretschmer et al., 2012). Approximately 3 ppm uncertainty in CO_{2}
mole fractions has been attributed to PBLH errors over Europe during the
summertime (Gerbig et al., 2008; Kretschmer et al., 2012). These studies
have attributed the errors to the lack of sophisticated subgrid
parameterization, especially PBL schemes and land surface models (LSMs). This
led other studies (Kretschmer et al., 2012; Lauvaux and Davis, 2014; Feng et
al., 2016) to evaluate the impact of different PBL parameterizations on
simulated atmospheric CO_{2}. These studies have found systematic
errors of several ppm in atmospheric CO_{2} that can generate biased
inverse flux estimates. While there is an agreement that errors in the
vertical mixing and advection schemes can directly affect the inverse fluxes,
other components of the model physics (e.g., convection, large-scale forcing)
have not been carefully evaluated.

Atmospheric transport models have multiple sources of uncertainty including the boundary conditions, initial conditions, model physics parameterization schemes and parameter values. With errors inherited from all of these sources, ensembles have become a powerful tool for the quantification of atmospheric transport uncertainties. Different approaches have been evaluated in the carbon cycle community to represent the model uncertainty: (1) the multi-model ensembles that encompass models from different research institutions around the world (e.g., TransCom experiment; Gurney et al., 2002; Baker et al., 2006; Patra et al., 2008; Peylin et al., 2013; Houweling et al., 2010), (2) multi-physics ensembles that involve different model physics configurations generated by the variation of different parameterization schemes from the model (e.g., Kretschmer et al., 2012; Yver et al., 2013; Lauvaux and Davis, 2014; Angevine et al., 2014; Feng et al., 2016; Sarmiento et al., 2017) and (3) multi-analysis (i.e., forcing data) that consists of running a model over the same period using different analysis fields (where perturbations can be added) (e.g., Lauvaux et al., 2009; Miller et al., 2015; Angevine et al., 2014). These ensembles are informative (e.g., Peylin et al., 2013; Kretschmer et al., 2012; Lauvaux and Davis, 2014) but have some shortcomings. In some cases, the ensemble spread includes a mixture of transport model uncertainties and other errors such as the variation in prior fluxes or the observations used. Other studies have only varied the PBL scheme parameterizations. None of these studies have carefully assessed whether or not their ensemble spreads represent the actual transport uncertainties.

In the last two decades, the development of ensemble methods has improved the
representation of transport uncertainty using the statistics of large
ensembles to characterize the statistical spread of atmospheric forecasts
(e.g., Evensen, 1994a, b). Single-physics ensemble-based statistics are highly
susceptible to model error, leading to underdispersive ensembles (e.g., Lee,
2012). Large ensembles (> 50 members) remain computationally
expensive and ill adapted to assimilation over longer timescales such as
multi-year inversions of long-lived species (e.g., CO_{2}).
Smaller-size ensembles would be ideal, but most initial-condition-only
perturbation methods produce unreliable and overconfident representations of
the atmospheric state (Buizza et al., 2005). An ensemble used to explore and
quantify atmospheric transport uncertainties requires a significant number of
members to avoid sampling noise and the lack of dispersion of the ensemble
members (Houtekamer and Mitchell, 2001). However, large ensembles are
computationally expensive. Limitations in computational resources lead to
restrictions, including the setup of the model (e.g., model resolution,
nesting options, duration of the simulation) and the number of ensemble
members. It is desirable to generate an ensemble that is capable of
representing the transport uncertainties and which does not include any
redundant members.

Various post-processing techniques can be used to calibrate or
“down-select” from a transport ensemble of 50 or more members to a subset
of ensemble members that represent the model transport uncertainties (e.g.,
Alhamed et al., 2002; Garaud and Mallet, 2011; Lee, 2012; Lee et al., 2016).
Some of these techniques are principal component analysis (e.g., Lee, 2012),
*k*-means cluster analysis (e.g., Lee et al., 2012) and hierarchical cluster
analysis (e.g., Alhamed et al., 2002; Yussouf et al., 2004; Johnson et al.,
2011; Lee et al., 2012, 2016). Riccio et al. (2012) applied the concept of
“uncorrelation” to reduce the number of members without using any
observations. Solazzo and Galmarini (2014) reduced the number of members by
finding a subset of members that maximize a statistical performance skill
such as the correlation coefficient, the root mean square error or the
fractional bias. Other techniques applied less commonly to the calibration of
the ensembles include simulated annealing and genetic algorithms (e.g., Garaud
and Mallet, 2011). All these techniques are capable of eliminating those
members that are redundant and generating an ensemble with a smaller number
of members that represents the uncertainty of the atmospheric transport model
more faithfully than the larger ensemble.

In this study, we start with a large multi-physics/multi-analysis ensemble of
45 members presented in Díaz-Isaac et al. (2018) and apply a
calibration process similar to the one explained in Garaud and Mallet (2011). Two principal features characterize an ensemble: reliability and
resolution. The reliability is the probability that a simulation has of
matching the frequency of an observed event. The resolution is the ability
of the system to predict a specific event. Both features are needed in order
to represent model errors accurately. Our main goal is to down-select the
large ensemble to generate a calibrated ensemble that will represent the
uncertainty of the transport model with respect to meteorological variables
of most importance in simulating atmospheric CO_{2}. These variables are
the horizontal mean PBL wind speed and wind direction, and the vertical
mixing of surface fluxes, i.e., PBLH. We focus on the criterion that will
measure the reliability of the ensemble, i.e., the probability of the
ensemble in representing the frequency of events (i.e., the spatiotemporal
variability of the atmospheric state). For the down-selection of the
ensemble, we will use two different techniques: simulated annealing and a
genetic algorithm (from now on referred to as calibration techniques/processes). In a
final step, the ensemble with the optimal reliability will be selected by
minimizing the biases in the ensemble mean. We will evaluate which physical
parameterizations play important roles in balancing the ensembles and
evaluate how well a pure physics ensemble can represent transport
uncertainty.

2 Methods

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We generate an ensemble using the Weather Research and Forecasting (WRF)
model version 3.5.1 (Skamarock et al., 2008), including the chemistry module
modified in this study for CO_{2} (WRF-ChemCO_{2}). The ensemble
consists of 45 members that were generated by varying the different physics
parameterization and meteorological data. The land surface models, surface
layers, planetary boundary layer schemes, cumulus schemes, microphysics
schemes and meteorological data (i.e., initial and boundary conditions) are
alternated in the ensemble (see Table 1). All the simulations use the same
radiation schemes, both long- and shortwave.

The different simulations were run using the one-way nesting method, with two nested domains (Fig. 1). The coarse domain (d01) uses a horizontal grid spacing of 30 km and covers most of the United States and part of Canada. The inner domain (d02) uses a 10 km grid spacing, is centered in Iowa and covers the Midwest region of the United States. The vertical resolution of the model is described with 59 vertical levels, with 40 of them within the first 2 km of the atmosphere. This work focuses on the simulation with higher resolution; therefore, only the 10 km domain will be analyzed.

The CO_{2} fluxes for summer 2008 were obtained from NOAA Global
Monitoring Division's CarbonTracker version 2009 (CT2009) data assimilation
system (Peters et al., 2007; with updates documented at
https://www.esrl.noaa.gov/gmd/ccgg/carbontracker/, last access:
17 January 2018). The different surface fluxes from CT2009 that we propagate
into the WRF-ChemCO_{2} model are fossil fuel burning, terrestrial
biosphere exchange and exchange with oceans. The CO_{2} lateral
boundary conditions were obtained from CT2009 mole fractions. The
CO_{2} fluxes and boundary conditions are identical for all ensemble
members.

Our interest is to calibrate the ensemble over the US Midwest using the
meteorological observations available over this region. The calibration of
the ensemble will be done only within the inner domain. To perform the
calibration, we used balloon soundings collected over the Midwest region
(Fig. 1). Meteorological data were obtained from the University of Wyoming's
online data archive (http://weather.uwyo.edu/upperair/sounding.html,
last access: 20 July 2018) for 14 rawinsonde stations over the US Midwest
region (Fig. 1). To evaluate how the new calibrated ensemble impacts
CO_{2} mole fractions, we will use in situ atmospheric CO_{2}
mole fraction data provided by seven communication towers (Fig. 1). Five of
these towers were part of a Penn State experimental network, deployed from
2007 to 2009 (Richardson et al., 2012; Miles et al., 2012, 2013;
https://doi.org/10.3334/ORNLDAAC/1202). The other two towers (Park Falls – WLEF; West
Branch – WBI) are part of the Earth System Research Laboratory/Global
Monitoring Division (ESRL/GMD) tall tower network (Andrews et al., 2014),
managed by NOAA. Each of these towers sampled air at multiple heights,
ranging from 11 to 396 m above ground level (m a.g.l.).

The ensemble will be calibrated for three different meteorological
variables: PBL wind speed, PBL wind direction and planetary boundary layer
height (PBLH). We will calibrate the ensemble with the late afternoon data
(i.e., 00:00 UTC) from the different rawinsondes. In this study, we use only
daytime data, because we want to calibrate and evaluate the ensemble under
the same well-mixed conditions that are used to perform atmospheric
inversions. For each rawinsonde site, we will use wind speed and wind
direction observations from approximately 300 m a.g.l. We
choose this observational level because we want the observations to lie
within the well-mixed layer, the layer into which surface fluxes are
distributed, and the same air mass that is sampled and simulated for
inversions based on tower CO_{2} measurements.

The PBLH was estimated using the virtual potential temperature gradient
(*θ*_{ν}). The method identifies the PBLH as the first point above
the atmospheric surface layer, where (1) *θ*_{ν} is greater than or
equal to 0.2 K km^{−1}, and (2) the difference between the surface and
the threshold-level virtual potential temperature is greater than or equal to
3 K (${\mathit{\theta}}_{\mathit{\nu}\mathrm{s}}-{\mathit{\theta}}_{\mathit{\nu}}\ge \mathrm{3}$ K).

WRF derives an estimated PBLH for each simulation; however, the technique used to estimate the PBLH varies according to the PBL scheme used to run the simulation. For example, the YSU PBL schemes estimate PBLH using the bulk Richardson number (Hong et al., 2006), the MYJ PBL scheme uses the turbulent kinetic energy (TKE) to estimate the PBLH (Janjic, 2002), and the MYNN PBL scheme uses twice the TKE to estimate the PBLH. To avoid any errors from the technique used to estimate the PBLH, we decided to estimate the PBLH from the model using the same method used for the observations. Simulated PBLH will be analyzed at the same time as the observations, 00:00 UTC, i.e., late afternoon in the study region.

We analyzed CO_{2} mole fractions collected from the sampling levels at or
above 100 m a.g.l., which is the highest observation level across the Mid-Continent Intensive
(MCI) network (Miles et al., 2012). This ensures that the observed mole fractions
reflect regional CO_{2} fluxes and not near-surface gradients of CO_{2}
in the atmospheric surface layer (ASL) or local CO_{2} fluxes (Wang et
al., 2007). Both observed and simulated CO_{2} mole fractions are averaged
from 18:00 to 22:00 UTC (12:00–16:00 LST), when the daytime period of the
boundary layer should be convective and the CO_{2} profile well mixed
(e.g., Davis et al., 2003; Stull, 1988). This averaged mole fraction will be
referred to hereafter as daily daytime average (DDA).

In this research, we want to test the performance of the transport ensemble and try to achieve a better representation of transport uncertainties, if possible using an ensemble with a smaller number of members. A series of statistical metrics are used as criteria to measure the representation of uncertainty by the ensemble for the period of 18 June to 21 July 2008. The criteria used for our down-selection process include rank histograms, rank-histogram scores and ensemble bias.

The rank histogram and the rank-histogram scores are tools used to measure
the spread and hence the reliability of the ensemble (see Fig. A1 in
the Appendix). The rank histogram (Anderson, 1996; Hamill and Colucci, 1997;
Talagrand et al., 1999) is computed by sorting the corresponding modeled
variable of the ensemble in increasing order and then a rank among the
sorted predicted variables from lowest to highest is given to the
observation. The ensemble members are sorted to define “bins” of the
modeled variable; if the ensemble contains *N* members, then there will be
*N*+1 bins. If the rank is zero, then the observed variable value is lower
than all the modeled variable values, and if it is *N*+1, then the
observation is greater than all of the modeled values. If the ensemble is
perfectly reliable, the rank histogram should be flat (i.e., flatness equal
to 1). This happens when the probability of occurrence of the observation
within each bin is equal. A rank histogram that deviates from the flat shape
implies a biased, overdispersive or underdispersive ensemble. A “U-shaped”
rank histogram indicates that the ensemble is underdispersive; normally, in
this type of ensemble, the observations tend to fall outside of the envelope
of the ensemble. This kind of histogram is associated with a lack of
variability or an ensemble affected by biases (Hamill, 2001). A
“central-dome” (or “A-shaped”) histogram indicates that the ensemble is
overdispersive; this kind of ensemble has an excess of variability. If the
rank histogram is overpopulated at either of the ends of the diagram, then
this indicates that the ensemble is biased.

The rank-histogram score is used to measure the deviation from flatness of a rank histogram:

$$\begin{array}{}\text{(1)}& \mathit{\delta}={\displaystyle \frac{N+\mathrm{1}}{NM}}\sum _{j=\mathrm{0}}^{N}({r}_{j}-\stackrel{\mathrm{\u203e}}{r}{)}^{\mathrm{2}},\end{array}$$

and should ideally be close to 1 (Talagrand et al., 1999; Candille and
Talagrand, 2005). In Eq. (1), *N* is the number of members (i.e., models),
*M* is the number of observations, *r*_{j} the number of observations of
rank *j*, and $\stackrel{\mathrm{\u203e}}{r}=M/(N+\mathrm{1})$ is the expectation of *r*_{j}. In theory,
the optimal ensemble has a score of 1 when enough members are
available. A score lower than 1 would indicate overconfidence in the
results, with an ensemble matching the observed variability better than
statistically expected. Having a score smaller than 1 would not affect the
selection process. Nevertheless, a flat rank histogram does not necessarily
mean that the ensemble is reliable or has enough spread. For example, a flat
histogram can still be generated from ensembles with different conditional
biases (Hamill, 2001). The flat rank histogram can also be produced when
covariances between samples are incorrectly represented. Therefore,
additional verification analysis has to be introduced to certify that the
calibrated ensemble has enough spread and is reliable. We introduce
hereafter several additional metrics used to evaluate the ensemble.

Atmospheric inverse flux estimates are highly sensitive to biases. The bias, or the mean of the model–data mismatches, was used to assist the selection of the calibrated sub-ensemble. We identify a sub-ensemble that has minimal bias:

$$\begin{array}{}\text{(2)}& \mathrm{Bias}={\displaystyle \frac{\mathrm{1}}{M}}\sum _{i=\mathrm{1}}^{M}\left({p}_{i}\right),\end{array}$$

where *p*_{i} is the difference between the modeled wind speed, direction or
PBLH, and the observed value, *M* is the number of measurements and *i* sums over
each of the rawinsonde measurements.

Different statistical tools were used to evaluate both the large (45-member) ensemble and calibrated ensemble; these statistics include Taylor diagrams, spread–skill relationship and ensemble root mean square deviation (RMSD). These statistical analyses will be used to describe the performance of each member (standard deviations and correlations), ensemble spread (root mean square deviation) and error structures in space (error covariance), which will allow us to evaluate all the important aspects of an ensemble.

We use Taylor diagrams to describe the performance of each of the models of the large ensemble (Taylor, 2001). The Taylor diagram relies on three nondimensional statistics: the ratio of the variance (model variance normalized by the observed variance), the correlation coefficient, and the normalized center root mean square (CRMS) difference (Taylor, 2001). The ratio of the variance or normalized standard deviation indicates the difference in amplitude between the model and the observation. The correlation coefficient measures the similarity in the temporal variation between the model and the observation. The CRMS is normalized by the observed standard deviation and quantifies the ratio of the amplitude of the variations between the model and the observations.

To verify that the ensemble captures the variability in the model performance across space and time, we computed the relationship between the spread of the ensemble and the skill of the ensemble over the entire dataset (i.e., spread–skill relationship). The linear fit between the two parameters measures the correlation between the ensemble spread and the ensemble mean error or skill (Whitaker and Lough, 1998). The ensemble spread is calculated by computing the standard deviation of the ensemble and the mean error by computing the absolute difference between the ensemble mean and the observations. Ideally, as the ensemble skill improves (the mean error gets smaller), the ensemble spread becomes smaller, and vice versa. Compared to the rank histograms, spread–skill diagrams represent the ability of the ensemble to represent the errors in time and space.

The spread of the ensemble is evaluated in time, using the RMSD. The RMSD does not consider the observations as we take the square root of the average difference between model configuration and the ensemble mean. Additionally, we use the mean and standard deviation of the error (model–data mismatch) to evaluate the performance of each of the members selected for the calibrated ensembles.

Transport model errors in atmospheric inversions are described in the
observation error covariance matrix and hence in CO_{2} mole fractions
(ppm^{2}). Therefore, we evaluate the impact of the calibration on the
variances of CO_{2} mole fractions. For the covariances, we compare the
spatial extent of error structures between the full ensemble and the
reduced-size ensembles by looking at spatial covariances from our
measurement locations. The limited number of members is likely to introduce
sampling noise in the diagnosed error covariances. We also know that the
full ensemble is not a perfect reference, but we believe it is less noisy. The
covariances were directly derived from the different ensembles to estimate
the increase in sampling noise as a function of the ensemble size.

In this study, we want to test the ability to reduce the ensemble from 45 members to an ensemble with a smaller number of members that is still capable of representing the transport uncertainties and does not include members with redundant information. The number of ideal ensemble members could have been decided by performing the calibration for all the different sizes of ensemble smaller than 45 members. However, we decided to use an objective approach to select the total number of members of the sub-ensemble. Therefore, we use the Garaud and Mallet (2011) technique to define the size of the calibrated sub-ensemble that each optimization technique will generate. The size of the sub-ensemble was determined by dividing the total number of observations by the maximum frequency in the large ensemble (45-member) rank histogram. We are going to generate sub-ensembles of three different sizes (number of members) to evaluate the impact that an ensemble size has on the representation of atmospheric transport uncertainties. Each of the ensembles will be calibrated for the period of 18 June to 21 July 2008.

Two optimization methods, simulated annealing (SA) and a genetic algorithm
(GA), are used to select a sub-ensemble that minimizes the rank-histogram
score (*δ*), which is the criterion that each algorithm will use to
test the reliability of the ensemble. Each method will select a sub-ensemble
that best represents the model uncertainties of PBL wind speed, PBL wind
direction and PBLH.

In this study, SA and GA techniques will randomly search for the different
combinations of members and compute the rank-histogram score. Both
techniques generate a sub-ensemble (*S*) of size *N*. For the first test, we will
use these algorithms to choose the combination of members that optimize the
score of the reduced ensemble *δ* (*S*) (i.e., rank-histogram score) for
each variable. With this evaluation, we determine if each optimization
technique yields similar calibrated ensembles and if the calibrated
ensembles are similar among the different meteorological variables. In the
second test, we calibrate the ensemble for all three variables
simultaneously, where we use the sum of the score squared: [*δ*(*S*)]^{2}:

$$\begin{array}{}\text{(3)}& [\mathit{\delta}\left(S\right){]}^{\mathrm{2}}=[{\mathit{\delta}}_{\mathrm{wspd}}\left(S\right){]}^{\mathrm{2}}+[{\mathit{\delta}}_{\mathrm{wdir}}\left(S\right){]}^{\mathrm{2}}+[{\mathit{\delta}}_{\mathrm{pblh}}\left(S\right){]}^{\mathrm{2}},\end{array}$$

to control acceptance of the sub-ensembles. In Eq. (3), *δ*_{wspd}(*S*), *δ*_{wdir}(*S*) and *δ*_{pblh}(*S*) are the scores of
the sub-ensemble for PBL wind speed, PBL wind direction and PBLH,
respectively.

SA is a general probabilistic local search algorithm, described by Kirkpatrick et al. (1983) and Černý et al. (1985) as an optimization method inspired by the process of annealing in metal work. Based on the Monte Carlo iteration solving method, SA finds the global minimum using a cost function that gives to the algorithm the ability to jump or pass multiple local minima (see Fig. A2 in the Appendix). In this case, the optimal solution is a sub-ensemble with a rank-histogram score close to 1.

The SA starts with a randomly selected sub-ensemble. The current state (i.e.,
initial random sub-ensemble) has a lot of neighbor states (i.e., other
randomly generated sub-ensembles) in which a unit (i.e., model) is changed,
removed or replaced. Let *S* be the current sub-ensemble and *S*^{′} be the
neighbor sub-ensemble. *S*^{′} is a new sub-ensemble (i.e., neighbor) that is
randomly built from the current sub-ensemble with one model added, removed or
replaced. To minimize the score *δ*, only two transitions to the
neighbors are possible. In the first transition, if the score of the neighbor
sub-ensembles *δ*(*S*^{′}) is lower than the current sub-ensemble *δ*(*S*), then *S*^{′} becomes the current sub-ensemble and a new neighbor
sub-ensemble is generated. In the second transition, if the score of the
neighbor sub-ensemble *δ*(*S*^{′}) is greater than the current sub-ensemble
*δ*(*S*), moving to the neighbor *S*^{′} only occurs through an acceptance
probability. This acceptance probability is equal to $\mathrm{exp}\left(-\frac{\mathit{\delta}\left({S}^{\prime}\right)-\mathit{\delta}\left(S\right)}{T}\right)$ and it only allows the movement to the neighbor
*S*^{′} if $u<\mathrm{exp}\left(-\frac{\mathit{\delta}\left({S}^{\prime}\right)-\mathit{\delta}\left(S\right)}{T}\right)$. For the
acceptance probability, *u* is a random number uniformly drawn from [0,1] and
*T* is called temperature, and it decreases after each iteration following a
prescribed schedule. The acceptance probability is high at the beginning, and
the probability of switching to a neighbor is less at the end of the
algorithm. The possibility to select a less optimal state *S*^{′},
i.e., with higher *δ*(*S*^{′}), is meant to escape local minima where the
algorithm could remain trapped.

When the algorithm reaches the predefined number of iterations, we collect
only the accepted sub-ensemble *S* and their respective scores *δ*(*S*). When
the algorithm finishes with the iterations, we choose the ensemble that has
both the smallest rank-histogram score and lowest bias among the different
sub-ensembles (see Sect. 2.7). The number of iterations was defined by
sensitivity test and repetitivity of the experiments (see Sect. 2.6).

GA is a stochastic optimization method that mimics the
process of biological evolution, with the selection, crossover and mutation
of a population (Fraser and Burnell, 1970; Crosby, 1973; Holland, 1975). Let
*S*_{i} be an individual, that is, a sub-ensemble, and let $P=\mathit{\{}{S}_{\mathrm{1}},\mathrm{\dots},{S}_{i},\mathrm{\dots},{S}_{{N}_{\mathrm{pop}}}\mathit{\}}$ be a population of *N*_{pop}
individuals (see Fig. A3 in the Appendix). As a first step, in the GA, a random
population is generated (denoted *P*^{0}). Then this population will go
through two out of the three steps of the genetic algorithm, (1) selection
and (2) crossover. In the selection step, we select half of the best
individuals with respect to the score (i.e., summation of the score of three
variables *δ*(*S*)). For the second step, a crossover among the selected
individuals occurs when two parents create two new children by exchanging
some ensemble members. A new population is generated with *N*_{pop}∕2
parents and *N*_{pop}∕2 children.

This process is repeated until it reaches the specified number of iterations. This algorithm will provide at the end a population of individuals with a better rank-histogram score than the initial population. Out of all those individuals, we choose the sub-ensemble with the best score for the three variables (i.e., wind speed, wind direction and PBLH) and with a smaller bias than the large ensemble.

Various inputs are required to guide the selection algorithms. For example,
we typically need to choose the initial and final temperature (*T*_{0} and
*T*_{f}) for the SA and its schedule, the best population size
(*N*_{pop}) for the GA and the number of iterations for each
algorithm. The temperature of the SA, the *N*_{pop} of the GA and the
number of iterations were chosen by running the algorithms multiple times and
confirming that the system reached similar solutions with independent
minimization runs. If similar solutions were not achieved within multiple SA
or GA runs, the algorithm parameters were altered to increase the breadth of
the search. For the SA, we found that 20 000 iterations yielded similar
solutions after multiple runs of the algorithm. For the GA, 30 to 50
iterations were sufficient as long as the ensemble was smaller than eight
members. For an ensemble of 10 members, we needed to increase to 100
iterations. Another factor that was important in the SA was the initial
temperature used in the algorithm and the temperature decrease for each
iteration. While the temperature is high, the algorithm will accept with more
frequency the poorer solutions; as the temperature is reduced, the acceptance
of poorer solutions is reduced. Therefore, we needed to provide an initial
(*T*_{0}) and final (*T*_{f}) temperature that allowed the system to
reduce its acceptance condition gradually and to search more combinations of
members to identify the best solution or sub-ensemble. We determine the
optimal parameters for SA by the maximum number of ensemble solutions which
indicates that the algorithm explored the largest space of solution with
*T*_{0} equal to 20 and *T*_{f} equal to $\mathrm{1}\times {\mathrm{10}}^{-\mathrm{3}}$. For GA,
the larger the population, the more we can explore the space to find an
optimal solution. We found that a *N*_{pop} of 280 individuals was the
value that produced similar solutions (sub-ensembles) after multiple runs.

The selection process is performed in three distinct steps to ensure that the final calibrated ensembles will be the optimal combinations of model configurations (Fig. 2). First, the flatness of the rank histograms will control the acceptance of the calibrated sub-ensembles by the selection algorithms (see Fig. A1 in the Appendix). The flatness is defined by Eq. (1) for the single-variable calibration and Eq. (3) for the calibration of the three variables simultaneously. The algorithm selects multiple sub-ensembles with a rank-histogram score smaller than 6 for each individual meteorological variable, or smaller than the original ensemble score if higher than 6 (see Fig. 2 and Table 2). In general, the lowest scores are found for PBLH and the highest for wind speed, as shown in Fig. 3. As a second step, sub-ensembles accepted by SA and GA algorithms with a bias larger than the bias of the full ensemble are filtered out. This step is critical to avoid the selection of biased ensembles as discussed by Hamill et al. (2001). Finally, the remaining calibrated ensembles are compared among SA and GA techniques to identify if both algorithms provide a common solution. If multiple common solutions were identified, the final sub-ensemble was determined by the solution with the smallest score and bias. However, if no common solution was found by both techniques, the final sub-ensemble corresponds to the smallest score among the different solutions that share > 50 % of the same model configurations.

3 Results

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In this section, we evaluate the performance of the large ensemble. Our goal is to test the ensemble skill (ability of the models to match the observations) and the spread (variability across model simulations to represent the uncertainty). We will evaluate the skill and the spread for PBLH, PBL wind speed and PBL wind direction across the region of study using afternoon (00:00 UTC) rawinsonde observations.

We evaluate the performance of the different models of the 45-member ensemble by computing the normalized standard deviation, normalized center root mean square and correlation coefficient for wind speed (Fig. 4a), wind direction (Fig. 4b) and PBLH (Fig. 4c) (Taylor, 2001). The majority of the model configurations produce winds speeds and directions with higher standard deviations (more variability) than the observations, whereas the simulations over- and underestimate PBLH variability depending on the model configuration. The model–data correlations with wind speed and wind direction are between 0.4 and 0.7, whereas the PBLH shows a smaller correlation, between 0.3 and 0.6. The range of modeled PBL heights will provide a wide spectrum of alternatives to select the optimal calibrated sub-ensemble. However, wind speed and wind direction do not show much difference among the different models. This limited spread potentially reduces the selection of the model configurations to produce a sub-ensemble that matches the observed variability.

We illustrate the ensemble spread and how well this ensemble encompasses the observations using the time series of the simulated and observed meteorological variables. Figure 5 shows the time series of the ensemble spread for wind speed, wind direction and PBLH at Green Bay (GRB; Fig. 5a, c, e) and Topeka (TOP; Fig. 5b, d, f) sites. The time series show qualitatively that simulated wind speed (Fig. 5a–b) and wind direction (Fig. 5c–d) have a smaller spread compared to PBLH (Fig. 5e–f). Figure 5 shows how the ensemble can have a small spread and still encompass the observations (i.e., DOY 183; Fig. 5c), and have a large spread and not encompass the observation (i.e., DOY 174; Fig. 5e). These time series suggest that the ensemble may struggle to encompass the observed wind speed and wind direction more than the PBLH.

Figure 6 shows the rank histograms of the 45-member ensemble for each of the meteorological variables that we use to calibrate the ensemble (i.e., wind speed, wind direction and PBLH). In these rank histograms, we include all 14 rawinsonde sites. All the rank histograms have a U shape. U-shaped histograms mean the ensemble is underdispersive; that is, the model members are too often all greater than or less than the observed atmospheric values (e.g., DOY 178–181; Fig. 5b). Each rank histogram has the first rank as the highest frequency, indicating that observations are most frequently below the envelope of the ensemble (e.g., DOY 178–180; Fig. 5b). The rank-histogram score for each of the variables is greater than 1, confirming that we do not have optimal spread in our ensemble. Table 2 shows that both wind speed and wind direction have a higher rank-histogram score (i.e., ≥6) than the PBLH, which has a score of 3.2. The ensemble mean wind speed and PBLH show a small positive bias relative to the observations, averaged across the region, whereas wind direction has a very small negative bias.

Figure 7 shows the spread–skill relationship, another method that we use to examine the representation of errors of the ensemble. Wind direction (Fig. 7b) shows a higher correlation between the spread and the skill compared to the PBLH (Fig. 7c) and the wind speed (Fig. 7a). Therefore, the ensemble has a wider spread when the model–data differences are larger. The PBLH and wind speed show consistently poorer skill (a large mean absolute error) compared to their spread. This supports the conclusion that the large ensemble is underdispersive for these variables. None of these variables show a correlation equal to 1; this implies that our ensemble spread does not match exactly the atmospheric transport errors on a day-to-day basis. This feature is common among ensemble prediction systems (Wilks et al., 2011) and should not impair the ability to identify the optimal reduced-size ensembles.

In this section, we show the results of the calibrated ensembles generated
with both SA and GA. Each calibration was performed for three different
sub-ensemble sizes; the size of the ensembles is determined using the
technique explained in Sect. 2.4. To compute the size of the sub-ensemble,
we use the maximum frequency of the rank histogram using the large ensemble
(Fig. 6). In this case, the maximum frequency is the left bar (*r*_{0}) of
every rank histogram. This technique yields the result that the calibrated
ensemble should have about 8–10 members depending in the variable to be
used. Therefore, for this study, we will generate 10-, 8- and 5-member
ensembles using the two calibration techniques.

Table 3 shows that both techniques (i.e., SA and GA) were able to find similar combinations of model configurations (i.e., an ensemble that shares more than half of the members) when each meteorological variable was used separately. The configurations chosen for each sub-ensemble vary significantly across the different variables, with the exception of the 10-member ensemble calibrated using wind speed and wind direction. The majority of the ensembles include model configuration 14. This model configuration, as shown in Díaz-Isaac et al. (2018), introduces large errors for both wind speed and wind direction, and is selected to allow for sufficient spread of these variables in the sub-ensembles. The final scores of the calibrated ensembles for each variable show that finding a calibrated sub-ensemble that reaches a score of 1 is not possible for wind speed and wind direction. A sub-ensemble with a score less than or equal to 1 can be found for PBLH. Figure 8 shows the rank histograms of the different calibrated ensembles (i.e., 10, 8 and 5 members) for each meteorological variable shown in Table 3. The calibrated ensembles of PBLH (Fig. 8c, f, i) are nearly flat for all ensemble sizes, whereas the 10- and 8-member sub-ensembles keep a slight U shape for wind speed and wind direction but are significantly flatter than the original ensemble. The ratio between the expected $\left(\stackrel{\mathrm{\u203e}}{r}\right)$ and observed frequency of the end members is reduced from 5 (original expected frequency of 0.02 with 0.1 frequency observed) to less than 2 (calibrated expected frequency of 0.1 with 0.15 frequency observed). The smallest rank-histogram scores for wind speed and wind direction are obtained with a five-member ensemble (Fig. 8g–h). The biases for all sub-ensembles (Table 3) are similar to or less than the bias of the large ensemble (Table 2).

Table 4 shows the sub-ensembles selected by SA. Each of the sub-ensembles has two simulations in common (i.e., 17 and 33), implying that these models are crucial to build an ensemble that best represents the transport errors for the three variables. Figure 9 shows the rank histograms of the sub-ensembles shown in Table 4. These rank histograms show that we were able to flatten the histogram relative to the 45-member ensemble for all three meteorological variables. Similar to the individual variable calibration, the rank histograms for wind speed (Fig. 9a, d) and wind direction (Fig. 9b, e) still show a U shape which is minimized for the smallest (i.e., five-member) sub-ensemble (Fig. 9g–h). The rank histograms are flatter for the PBLH (Fig. 9c, f, i) and the histogram score is closer to 1 (Table 4) compared to wind speed and wind direction. The rank-histogram scores for all variables are greater than those for one-variable optimization (see Table 4). The high rank-histogram scores are associated with the equal weight given to the three variables for this simultaneous calibration, where wind speed controlled the calibration process. For the calibration of the three variables together, we were not able to produce an ensemble for wind speed with a score smaller than 4; this ends up limiting the selection of the calibrated ensemble for the rest of the variables (see Fig. A4 in the Appendix). In addition, all these calibrated sub-ensembles have biases smaller in magnitude than the 45-member ensemble. Both wind speed and PBLH retain an overall positive bias, and wind direction a negative bias. The standard deviations of these three calibrated ensembles are larger than those of the large ensemble, consistent with the effort to increase the ensemble spread.

Using SA and GA techniques and the selection criteria detailed in Sect. 2.7 (i.e., low mean error of the entire ensemble), we defined an optimal five-member sub-ensemble (the optimal solution using both techniques) and nearly identical combinations of members for 10-and 8-member sub-ensembles, with only two model configurations not being shared by both algorithms. We also find that configuration 14 remains important for the multi-variable calibrated ensembles, as it was for the single-variable calibrated ensembles.

Both optimization techniques were able to generate sub-ensembles that reduce the U shape of the rank histograms, while significantly decreasing the number of members in the ensemble. A flatter histogram indicates that the ensemble is more reliable (unbiased) and has a more appropriate (greater) spread. The correlation between spread and skill for the wind direction increased, while wind speed and PBLH remain similar. Therefore, we conclude that the calibrated sub-ensembles are equivalent to or even better than the full ensemble to represent the daily model errors.

Figure 10 shows the time series of the different calibrated ensembles generated by the SA algorithm at the TOP site. In general, there are no major differences among 5- (Fig. 10a, d, g), 8- (Fig. 10e, h) and 10-member (Fig. 10c, f, i) ensembles. Figure 12 shows how the calibration can increase the spread of the ensemble to the extent of encompassing the observations (e.g., DOY 179; Fig. 10b–c) compared to the full ensemble (Fig. 5b). The ensemble spread was reduced after calibration at a few specific points in space and time.

Insight into the physics parameterizations can be gained by evaluating the calibrated ensembles. The LSM, PBL, CP and MP schemes, and reanalysis choice vary across all of the sub-ensemble members; no single parameterization is retained for all members in any of these categories. However, we also find that the calibrated ensembles rely upon certain physics parameterizations more than others. Figure 11 shows that most of the simulations in the calibrated ensemble use the RUC and thermal diffusion (T-D) LSMs in preference to the Noah LSM. In addition, more simulations use the MYJ PBL scheme than the other PBL schemes. The physics parameterizations shown with a higher percentage in Fig. 11 appear to contribute more to the spread of the ensemble than the other parameterizations.

We next explore the characteristics of the individual ensemble members that are retained in an effort to understand what member characteristics are important to increase the spread of the ensemble. Figure 12 shows the mean and standard deviation of the residuals for each simulation included in the five-member ensemble of SA and GA. Ensembles appear to need at least one member with a larger standard deviation to improve the spread for wind speed and wind directions (see member 23; Fig. 12a–b). Additionally, a member that has a large PBLH bias (see member 16; Fig. 12c) appears to be selected, highlighting the need for end members among the model configurations in order to reproduce the observed variability in PBLH. We note here that model configuration 14 was not selected when calibrating three variables together.

The calibrated ensembles found in this study were chosen based on the
meteorological variables and not on the CO_{2} mole fractions to avoid the
propagation of CO_{2} flux biases into the solution. We can now propagate
these uncertainties, represented by the ensemble spread, into the CO_{2}
concentration space. This straightforward calculation is possible because
every model simulation uses identical CO_{2} fluxes. We present here the
transport errors in both time and space with the spread in CO_{2} mole
fractions, comparing the initial (uncalibrated) 45-member ensemble to the
calibrated sub-ensembles.

Figure 13 shows the spread of daily daytime average CO_{2} mole fractions
across the different sub-ensemble sizes at Mead (Fig. 13a, d, g, j), West
Branch (Fig. 13b, e, h, k) and WLEF (Fig. 13c, f, i, l). The spread of the DDA
CO_{2} mole fractions of the large ensemble (Fig. 13a–c) does not appear
to differ in a systematic fashion from the spread of the calibrated
small-size ensembles (Fig. 13d–l). While the calibration has increased
the average ensemble spread, none of the ensembles consistently encompass the observations, either in terms of meteorological variables (Fig. 12) or
CO_{2} (Fig. 15). The CO_{2} differences between the models and the
observations may be caused by CO_{2} flux or boundary condition errors,
the two components impacting the modeled CO_{2} mole fractions in addition
to atmospheric transport. The cause of the total difference cannot be
determined from the CO_{2} data alone. The increased daily variance in
CO_{2} resulting from the ensemble calibration process is shown in
Fig. 14. The eight-member ensemble often has the maximum CO_{2} variance. Table 5
shows the spread (model–ensemble mean) and RMSE (model–data) ratio of the
CO_{2} mole fraction for the full and calibrated 10-member ensembles at
each in situ CO_{2} observation tower. The ratio of the variances is an
estimate of the contribution of the transport uncertainties to the CO_{2} model–data mismatch for the summer of 2008. This table
shows that the transport uncertainties represent about 20 % to 40 % of
the CO_{2} model–data mismatch. We found that values after calibration
show a slight increase compared to the full ensemble.

4 Discussion

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The calibration of the multi-physics/multi-analysis ensemble using SA and GA optimization techniques generated 10-, 8- and 5-member ensembles with a better representation of the error statistics of the transport model than the initial 45-member ensemble. One of our goals was to find sub-ensembles that fulfil the criteria of Sect. 2.7, independent of the selection algorithm and for multiple meteorological variables. Wind speed and wind direction statistics only improve by a modest amount in the calibrated ensembles as compared to the 45-member ensemble, while PBLH statistics, namely the flatness of the rank histogram, show a significant improvement in the calibrated ensembles. The variance in the calibrated ensembles increased relative to the 45-member ensemble but the potential for improvement was limited by the spread in the initial ensemble. Stochastic perturbations (e.g., Berner et al., 2009) could increase the spread of the initial ensemble, which, combined with the suite of model configurations, could better represent the model errors. Here, we limited the 45-member ensemble to mass-conserved, continuous flow (i.e., unperturbed) members that can be used in a regional inversion. Future work should address the problem of using an underdispersive ensemble before the calibration of the ensemble.

We first attempted to calibrate the ensemble for each meteorological variable
(i.e., wind speed, wind direction and PBLH). Table 3 shows that the different
sub-ensembles were able to follow the criteria presented in Sect. 2.7, but
the calibration of the single-variable ensembles did not allow us to find a
unique sub-ensemble that can be used to represent the errors of the three
variables. Therefore, the joint optimization of the three variables was
required to identify an ensemble that best represents model errors across the
three variables. By minimizing the sum of the squared rank-histogram scores
of the three variables, the selection algorithm found common solutions at the
expense of less satisfactory rank-histogram scores than were obtained for
single-variable ensembles (see Table 4). We
assumed that each variable was equally important to the problem, an
assumption that has not been rigorously evaluated. Future work on the
relative importance of meteorological variables on CO_{2}
concentration errors would help weigh the scores in the selection algorithms.

The calibrated ensembles show the rank-histogram score closer to 1 (Table 4), that is, flatter rank histograms (Fig. 9) compared to the 45-member ensemble (Table 2 and Fig. 6). The sub-ensembles do have a greater variance than the large ensemble (i.e., improved reliability) (Fig. 14). However, the spread–skill relationship (i.e., resolution) of the calibrated ensembles do not show any major improvement compared to the 45-member ensemble, implying that the spread of the ensemble does not represent the day-to-day transport errors well. While the rank histogram suggests that the different calibrated ensembles have enough spread, the spread–skill relationship indicates that our ensemble does not systematically encompass the observations. The disagreement between the rank histogram and the spread–skill relationship can be associated with the metric used for the calibration (i.e., rank histogram) and the biases included in the calibrated ensemble. Using the score of the rank histogram alone may not be sufficient to measure the reliability of the ensemble (Hamill, 2001); therefore, future down-selection studies should incorporate the resolution as part of the calibration process (skill score optimization). The biases in the model are a complex problem because there are many sources systematic errors within an atmospheric model (e.g., physical parameterizations and meteorological forcing). Future studies should consider data assimilation or improvement of the physics parameterizations to reduce or remove these systematic errors. To improve the representation of daily model errors, additional metrics should be introduced and the initial ensemble should offer a sufficient spread, possibly with additional physics parameterizations, additional random perturbations, or modifications of the error distribution of the ensemble (Roulston and Smith, 2003).

Rank histograms, as explained in Sect. 2.3.1, evaluate the ensemble by
ranking individual observations in a relative sense. The ensembles
calibrated using the rank histograms may be representing the variances over
the region correctly but not the spatial and temporal structures of the
errors (Hamill, 2001). These parameters are critical to inform regional
inversions of correlations in model errors, directly impacting flux
corrections (Lauvaux et al., 2009). In this study, the calibrated ensembles
show an improvement in the meteorological variances and an increase in the
CO_{2} variances relative to the uncalibrated ensemble. However, spatial
structures of the errors were not evaluated and may be impacted by sampling
noise. Few members will produce a statistically limited representation of the
model error structures. For example, ensemble model prediction systems use
at least 50 members to avoid sampling noise and correctly represent time and
space correlations. Figure 15 shows the spatial correlation of 300 m DDA
CO_{2} errors with respect to the Round Lake site on DOY 180. Error
correlations increase significantly as our ensemble size decreases. With
fewer members, spurious correlations increase, resulting in high
correlations at long distances. Assuming we sample only a few times the
distribution of errors, our ensemble is very likely to be affected by
spurious correlations with a variance on the order of 1∕*N*. We conclude here
that our reduced-size ensembles are impacted by sampling noise which would
require additional filtering. Previous studies have suggested objective
methods to filter the noise in small-size ensembles (i.e., Ménétrier
et al., 2015) or modeling the error structures using the diffusion equation
(e.g., Lauvaux et al., 2009). Future work should address the impact of the
calibration on the error structures as this information is critical in the
observation error covariance to assess the inverse fluxes. Concerning the
magnitudes of the error correlation, the calibrated sub-ensembles exhibit a
larger contrast in correlation values compared to the 45-member error
correlations. Overall, the different ensembles show similar flow-dependent
spatial patterns which demonstrates that the calibration process, even if
generating sampling noise, preserves the dominant spatial patterns in the
error structures. Therefore, the calibrated ensemble is likely to provide a
better representation of the variances and a similar spatial error structure
for the construction of error covariance matrices in regional inversions.

5 Conclusions

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We applied a calibration (or down-selection) process to a
multi-physics/multi-analysis ensemble of 45 members. In this calibration
process, two optimization techniques were used to extract a subset of
members from the initial ensemble to improve the representation of transport
model uncertainties in CO_{2} inversion modeling. We used purely
meteorological criteria to calibrate the ensemble and avoid contaminating
the calibration with CO_{2} flux errors. The calibrated ensembles were
optimized using criteria based on the flatness of the rank histogram. We
generated different calibrated ensembles for three meteorological variables;
PBL wind speed, PBL wind direction and PBLH. With these techniques, we
identified sub-ensembles by calibrating the three variables jointly. Both
techniques show that calibrated small-size ensembles can reduce the score of
the rank-histogram flatness and therefore improve the representation of the
model error variances with few members (between 5 and 10 members).

The calibration techniques improved the spread (flatness of the rank
histogram) of the ensembles and slightly improved the biases, which were
already small in the larger ensemble, but the calibration did not improve
daily atmospheric transport errors as shown by the spread–skill
relationship. We assessed how the calibrated ensemble errors propagate into
the CO_{2} mole fractions simulated with identical CO_{2} fluxes (i.e.,
independent of the atmospheric conditions). The spread from the calibrated
ensembles represented from 20 % to 40 % (Table 5) of the model–data 300 m DDA CO_{2} mismatches for summer 2008. These results suggest that
additional errors in CO_{2} fluxes and/or large-scale boundary conditions
represent a large fraction of the differences between modeled and observed
CO_{2}. Error correlations of the calibrated ensembles were compared to
the large ensemble to identify any impact of the calibration. Compared to
the initial error structures, the calibrated ensembles are most likely
affected by sampling noise across the region, which suggests that additional
filtering or modeling of the errors would be required in order to construct
the error covariance matrix for regional CO_{2} inversion.

Code availability

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Code availability.

The code is accessible upon request by contacting the corresponding author (lzd120@psu.edu).

Data availability

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Data availability.

Meteorological data were obtained from the University of
Wyoming's online data archive
(http://weather.uwyo.edu/upperair/sounding.html, last access:
20 July 2018) for the 14 rawinsonde stations. The tower atmospheric
CO_{2} concentration dataset is available online
(https://doi.org/10.3334/ORNLDAAC/1202, Miles et al., 2013) from the Oak Ridge National
Laboratory Distributed Active Archive Center, Oak Ridge, Tennessee, USA
https://doi.org/10.3334/ORNLDAAC/1202 (Miles et al., 2013). The other two towers (Park
Falls – WLEF; West Branch – WBI) are part of the Earth System Research
Laboratory/Global Monitoring Division (ESRL/GMD) tall tower network (Andrews
et al., 2014; https://www.esrl.noaa.gov/gmd/ccgg/insitu/, last access:
20 July 2018). The WRF model results are accessible upon request by
contacting the corresponding author (lzd120@psu.edu).

Appendix A

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Author contributions

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Author contributions.

LDI performed the model simulations, calibration and the model–data analysis. The calibration technique was coded by MB, LDI and TL based on the work of Garaud and Mallet (2011). TL, MB and KJD provided guidance with the calibration and model–data analysis. All authors contributed to the design of the study and the preparation the paper.

Competing interests

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Competing interests.

The authors declare that they have no conflict of interest.

Acknowledgements

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Acknowledgements.

This research was supported by NASA's Terrestrial Ecosystem and Carbon Cycle
Program (grant NNX14AJ17G), NASA's Earth System Science Pathfinder Program
Office, Earth Venture Suborbital Program (grant NNX15AG76), NASA Carbon
Monitoring System (grant NNX13AP34G) and an Alfred P. Sloan Graduate
Fellowship. We thank Natasha Miles, Chris E. Forest and Andrew Carleton for
fruitful discussions. Meteorological data used in this work were provided by
University of Wyoming's online data archive
(http://weather.uwyo.edu/upperair/sounding.html, last access:
12 July 2018). Observed atmospheric CO_{2} mole fraction was provided
by the NOAA Earth System Research Laboratory (cited in the text) and PSU in
situ measurement group (Miles et al., 2013).

Review statement

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Review statement.

This paper was edited by Marc von Hobe and reviewed by Lori Bruhwiler and Ian Enting.

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Short summary

We demonstrate that transport model errors, one of the main contributors to the uncertainty in regional CO_{2} inversions, can be represented by a small-size ensemble carefully calibrated with meteorological data. Our results also confirm transport model errors represent a significant fraction of the model–data mismatch in CO_{2} mole fractions and hence in regional inverse CO_{2} fluxes.

We demonstrate that transport model errors, one of the main contributors to the uncertainty in...

Atmospheric Chemistry and Physics

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