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Experimental site
Seawater samples were collected from two stations (Fig. S1a, offshore OMZ station PS2 and coastal OMZ station PS3) in the ETNP in March and April 2018 on board R/V Sally Ride (Cruise ID: SR 1805). NO2− kinetics and O2 effects experiments (≤1 day incubation) were performed at the oxycline, the oxic/anoxic interface (top) of the ODZ and the core of the ODZ at each station (Fig. S1b and Table S1). Long incubations (6–7 days) were performed at the core of the ODZ at both stations to investigate nitrogen reaction rates at very low O2 concentrations.
Sampling and incubation experiments for NO2− kinetics and O2 effects
Twelve 30-L Niskin bottles on a rosette with a conductivity–temperature–depth (CTD) profiler were used to collect seawater while recording in situ O2 concentration, temperature, pressure, salinity, and chlorophyll fluorescence. O2 concentration at selected ODZ depths was measured by STOX sensor (on the CTD profiler) with detection limit of 10 nM [42]. NO2− concentrations were measured by standard spectrophotometric methods onboard. O2 and NO2− concentrations were used to select sampling depths. NO3− and NO2− concentrations in incubation samples were measured on a chemiluminescence NO/NOx Analyzer (Teledyne API, San Diego, CA, USA) in the lab.
Seawater was collected from Niskin bottles into 60 mL air-tight serum bottles after overflowing three times in order to minimize O2 contamination. Serum bottles were sealed with rubber septa and aluminum seals ensuring the absence of bubbles inside bottles. Septa were deoxygenated in anaerobic chambers with three cycles of vacuum/helium flushing over a period of 1 month prior to the cruise. A helium headspace was created for each sample collected from anoxic depths (the top and the core of the ODZ), and then samples were flushed with helium for at least 15 min to remove O2 that might have been introduced during sampling.
To determine a single NO2− oxidation rate, a set of five serum bottles amended with 15NO2− tracer was incubated on board at 12 °C in the dark. Incubations were terminated in time series (one bottle at day zero (T0), two at 0.5 day and two at 1 day) by adding 0.2 mL of saturated ZnCl2. The T0 bottles served as killed controls for both tracer contamination and abiotic reactions. The observed temporal changes in isotopic signals that occurred in the live samples over a few days in the time courses would not be detected if abiotic reactions were taking place during the >3 months that elapsed before all samples were measured on the mass spec.
In the lab, NO2− was removed from samples using sulfamic acid, and NO3− in serum bottles was converted into N2O using the denitrifier method [43, 44]. Both concentration and isotopic composition of N2O were measured on a mass spectrometer (Delta Vplus, Thermo Fisher Scientific, Waltham, MA, USA) for calculating nitrite oxidation rate from the linear regression of the five nitrate concentrations at the three time points as previously described [3]. NO2− kinetics of NO2− oxidation were determined by measuring NO2− oxidation rates under different 15NO2− tracer concentrations (0.5–13.8 μM). For responses of NO2− oxidation to O2, different amounts (0, 0.2, 0.5, 1, 2, and 5 mL) of O2 saturated seawater collected from the same Niskin bottle were added into serum bottles to achieve different O2 concentrations. O2 concentrations in serum bottles were monitored by optical oxygen sensors with a detection limit of 62.5 nM (PyroScience GmbH, Aachen, Germany).
Half-saturation constant
Half-saturation constant (Km) is the nitrite concentration at which nitrite oxidation rate (V) equals half of the potential maximum rate (Vm). The curve fitting tool in MATLAB_R2015a was used to fit the Michaelis–Menten model (Eq. (2)) to determine Km.

$${{V}} ,=, {{V}}_{mathrm{m}} ,times,left[ {{mathrm{NO}}_2^-} right]{mathrm{/}}left( {left[ {{mathrm{NO}}_2^-} right] ,+, {{K}}_{mathrm{m}}} right).$$
( 2)

Long incubation experiments
Longer (≥6 days) incubations were performed in 12-mL exetainers. Seawater from the core of the ODZ (250 m at station PS2, 160 m at station PS3) was sampled into 320 mL ground glass-stoppered bottles, which were immediately transferred into a N2 flushed glove bag. 15NO2− was added into these bottles to reach final concentrations of 7.24 and 8.01 µM at stations PS2 and PS3, respectively. 15NO2− labeled seawater was aliquoted into exetainers and capped within the glove bag. The septa had been stored under helium for at least 1 month. Exetainers were purged with helium for 5 min. Every 12 h, microbial activity in triplicate exetainers was terminated by adding 0.05 mL of 50% w/v ZnCl2. In the lab, N2 produced in exetainers was measured on a mass spectrometer (Europa Scientific 20–20, Crewe, UK) [45]. Denitrification and anammox rates were computed from linear regression of N2 produced at three time points. 15NO3− was measured (as described above) in the same exetainers, and NO2– oxidation rates were computed from linear regression of NO3− produced at three time points. O2 was monitored throughout the incubations in parallel exetainer vials using LUMOS sensors [12, 13]. Each O2 production or consumption rate was determined by linear regression of O2 concentrations at 32 time points. One sensing spot, glued inside an exetainer vial, allowed the optode to measure O2 concentrations every 5 min from the outside. Detection limit and resolution of LUMOS sensors was ≈1 nM.
DNA sampling, extraction, sequencing, metagenomics, and metatranscriptomics analysis
Particulate DNA samples were collected by filtration onto 0.22 µm Sterivex filters from the ODZ core samples (250 m at station PS2, 160 m at station PS3). DNA was extracted using the modified plant tissue protocol (All Prep DNA/RNA Mini Kit, Qiagen, Valencia, CA, USA), and subjected to paired-end sequencing on an Illumina MiSeq to generate over 10 million read pairs for each sample by the Genomics Core Facility of Lewis-Sigler Institute for Integrative Genomics at Princeton University. Quality control of raw reads was performed by BBDuk (DOE Joint Genome Institute, Walnut Creek, CA, USA), and assembled into contigs using metaSPAdes v3.12.0 [46] with specified options (-k 21, 33, 55, 77, 99, 127 -m 500). Metagenome-assembled genomes (MAGs) were constructed using MetaBAT v2.12.1 [47] with default (sensitive) mode and contigs longer than 1500 bp. The quality of MAGs was determined by checkM [48]. Taxonomy of MAGs was predicted by GTDB-tk [49]. ETNP PS2 MAG-11 from the ODZ core at station PS2 and ETNP PS3 MAG-54 from the ODZ core at station PS3 were identified as Nitrospina-like NOB. The taxonomy of these two NOB MAGs (ETNP PS2 MAG-11 and ETNP PS3 MAG-54) was further confirmed by comparing them with known OMZ NOB MAGs (MAG-1 and MAG-2) [8] using ANI. ANI between MAGs was assessed using enveomics [50]. MAG-2 [8], ETNP PS2 MAG-11 and ETNP PS3 MAG-54 belong to the same species (threshold for species: ANI ≥95%) based on their ANI values (Table S3). Considering the low completeness of MAG-11 and the high contamination of MAG-54 constructed from the two new metagenomes here (Table S4), we decided to use MAG-2 as the representative for this NOB species.
We estimated the relative abundance and the transcriptional activity of the two known OMZ NOB species in different oceanic regions (including the two ETNP stations in this study) by mapping metagenomes and metatranscriptomes from this and other studies to MAG-1 and MAG-2. The relative abundance of MAG-1 or MAG-2 at stations PS2 and PS3 in the ETNP was calculated as RPKG (the number of metagenomic reads obtained in this study mapped to a MAG per MAG length (kb) per genome equivalents) [15]. Genome equivalents were estimated using MicrobeCensus v1.1.1 [15]. The transcriptional activity of MAG-1 or MAG-2 in ETNP and ETSP OMZs was assessed by mapping published metatranscriptomic reads from the ETNP (PRJNA263621) [51] and the ETSP (SRA023632.1) [52] to MAG-1 and MAG-2. The relative abundance of RNA in Fig. S4 was calculated as the number of metatranscriptomic reads mapped to a MAG divided by the number of total reads. Mapping was performed by Bowtie2 [53] using “very-sensitive” mode, and only reads with a mapping quality above 20 were included as mapped reads.
In order to explore the possibility of the presence of other NOB in the ODZ core at stations PS2 and PS3, we also estimated the relative abundance of other (putative) NOB using their marker gene, nxrB (nitrite oxidoreductase). First, we downloaded previously identified (putative) nxrB sequences from the ETSP OMZ [8]. Then, we color coded the genes in Fig. 4 based on previously defined clusters in a phylogenetic tree (see Fig. 3 in [8]): the Nitrospina cluster (blue) contains nxrB grouped with cultured marine NOB, Nitrospina gracilis. Based on BLASTp search results, amino acid sequence identities between all the OMZ nxrB in this cluster and that of Nitrospina gracilis were 96.71%, 96.71%, and 96.87% for NODE_69234, NODE_114897, and NODE_102582, respectively. Only this cluster is associated with known NOB, and the nitrite oxidation capacity of all the other genes associated with nxrB needs to be confirmed. The anammox cluster (red) contains anammox nxrB sequences. The putative cluster (black) contains nxrB grouped with microbes in which nitrite oxidation capacity has not been proven. This putative cluster fell between known NOB and anammox, and was implied to represent unidentified NOB [54]. The last cluster is called unknown nxrB (gray) because neither their phylogeny nor function can be determined based on their distant relationship with known NOB or anammox nxrB. Finally, we mapped the ETNP metagenomic reads obtained in this study to each nxrB gene. Relative abundance of nxrB genes was also expressed as RPKG: relative abundance of nxrB gene = (number of mapped reads to a certain nxrB gene)/(length of this nxrB in kb)/(genome equivalents).
To explore the potential metabolisms of NOB in anoxic ODZ cores, we searched for chlorite dismutase and NO dismutase (nod) genes in the NOB MAGs. First, protein-coding sequences in the two new NOB MAGs obtained here (ETNP PS2 MAG-11 and ETNP PS3 MAG-54) were predicted by Prodigal v2.6.3 [55]. The protein-coding sequences were annotated by the best BLASTp hits against the nr protein database. DNA sequences encoding Cld were identified in both ETNP NOB MAGs. Predicted Cld amino acid sequences encoded by ETNP MAGs were too short (120 aa) to be compared to Cld of Candidatus Nitrospira defluvii, but they only had one mismatch with Cld amino acid sequences of MAG-2 from the ETSP OMZ based on MUSCLE alignment using MEGA 7 software (Fig. S5b). Thus, we looked for the arginine173, the Cld activity marker, in longer Cld sequences of MAG-1 and MAG-2 by aligning their Cld with the Cld of Candidatus Nitrospira defluvii using MUSCLE in MEGA 7 software. Since nod was not found in the ETNP MAGs via annotation and was not reported in MAG-1 and MAG-2 in the previous study [8], gene search (i.e., searching nod against NOB MAGs) was performed using Hidden Markov Models by HMMER3 [56]. Reference sequences of the search included a nod sequence retrieved from Candidatus Methylomirabilis oxyfera [32] genome (accession numbers FP565575.1), and three environmental nod sequences (accession numbers: KX364450.1, KX364454.1, and KU933965.1). Search queries were the two ETNP MAGs and two ETSP MAGs in Table S3.
Estimation of NO2− oxidation through disproportionation using an inverse isotope model
We simulated the distribution of NO2− oxidation rates via disproportionation using a 1-D inverse isotope model [37] in the anoxic ODZ core from two stations in the ETSP OMZ (Fig. S6), where the complete suite of isotope and rate data have been previously published (Table S7). Briefly, the net biochemical production or consumption rate of each nitrogen compound (R14Ammonium, R14Nitrite, R14Nitrate, R15Nitrite, and R15Nitrate) was balanced by vertical diffusion and advection at steady state, and then five equations were used to balance measured 14NH4+, 14NO2−, 15NO2−, 14NO3−, and 15NO3− concentrations (Eqs. (3–7)). The exclusion of horizontal processes was justified [37] since the model was not run all the way to the surface, and using a constant vertical advection term does not violate continuity of the model. F is the rate of each nitrogen cycle process represented in this model (Fig. 5a, b). We modified the previous model by replacing canonical NO2− oxidation with NO2− disproportionation and using the recently determined stoichiometry of nitrate production by anammox from 0.3 to 0.16 [24]. Since OMZ NOB genomes encode nitrite oxidoreductase (catalyzes NO2−→NO3−) and nitrite reductase (catalyzes NO2−→NO), we assumed that the fractionation factor of nitrate production from nitrite by disproportionation is the same as by canonical nitrite oxidation (αDis = αNxr) based on the close clustering between OMZ NOB nxr and aerobic NOB (Nitrospina gracilis) nxr [8], and the fractionation factor of N2 produced by disproportionation is the same as by denitrification (αDisN2 = αNir). Rates of NO2− disproportionation (FDis), NO3− reduction (FNar), denitrification (FNir), and anammox (FAmx) were solved from the equations by the nonnegative least squares optimization routine (lsqnonneg) in MATLAB_R2015a as described previously [37].

$$R_{14Ammonium} ,=, 0.11,times {,}^{14}F_{Nir} ,+, 0.07,times {,}^{14}F_{Nar} ,- {,}^{14}F_{Amx},$$
( 3)

$$R_{14Nitrite} = – ,^{14}F_{Nir} ,+, {,}^{14}F_{Nar} ,-, (1 ,+, c) ,times,{,}^{14}F_{Amx} \ -, left( {5/3} right) ,times,{,}^{14}F_{Dis},$$
( 4)

$$R_{14Nitrate} ,=, – ^{14}F_{Nar} ,+, c ,times {,}^{14}F_{Amx} ,+, {,}^{14}F_{Dis},$$
( 5)

$$R_{15Nitrite} = -!! {,}^{14}F_{Nir}/alpha_{Nir} ,times,left( {left[ {{,}^{15}{rm{NO}}_{2}^{-}} right]/left[ {{,}^{14}{rm{NO}}_{2}^{-}} right]} right) \ +, {,}^{14}F_{Nar}/alpha_{Nar},times,left( {left[ {{,}^{15}{rm{NO}}_{3}^{-}} right]/left[ {{,}^{14}{rm{NO}}_{3}^{-}} right]} right)\ -, {,}^{14}F_{Amx}/alpha_{Amx},times,left( {left[ {{,}^{15}{rm{NO}}_{2}^{-}} right]/left[ {{,}^{14}{rm{NO}}_{2}^{-}} right]} right) \ -, c times !{,}^{14}F_{Amx}/alpha_{NxrAmx},times,left( {left[ {{,}^{15}{rm{NO}}_{2}^{-}} right]/left[ {{,}^{14}{rm{NO}}_{2}^{-}} right]} right)\ -, {,}^{14}F_{Dis}/alpha_{Dis}timesleft( {left[ {{,}^{15}{rm{NO}}_{2}^{-}} right]/left[ {{,}^{14}{rm{NO}}_{2}^{-}} right]} right) \ -, left( {2/3} right) !times ! {,}^{14}F_{Dis}/alpha_{DisN2} !times left(! {left[ {{,}^{15}{rm{NO}}_{2}^{-}} right]/left[ !{{,}^{14}{rm{NO}}_{2}^{-}} right]} !right),$$
(6)

$$R_{15Nitrate} = -!! {,}^{14}F_{Nar}/alpha_{Nar} ,times,left( {left[ {{,}^{15}{rm{NO}}_{3}^{-}} right]/left[ {{,}^{14}{rm{NO}}_{3}^{-}} right]} right) \ +, c times !{,}^{14}F_{Amx}/alpha_{NxrAmx},times,left( {left[ {{,}^{15}{rm{NO}}_{2}^{-}} right]/left[ {{,}^{14}{rm{NO}}_{2}^{-}} right]} right)\ +, ^{14}F_{Dis}/alpha_{Dis},times,left( {left[ {{,}^{15}{rm{NO}}_{2}^{-}} right]/left[ {{,}^{14}{rm{NO}}_{2}^{-}} right]} right).$$
(7) More

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Data selection and handling: death data
For mortality due to COVID-19, we used time series provided by the New York Times12. We selected the New York Times dataset because it is rigorously curated. We analyzed separately only counties that had records of 100 or more deaths by 23 May, 2020. The threshold of 100 was a balance between including more counties and obtaining reliable estimates of r(t). Preliminary simulations showed that time series with low numbers of deaths would bias r(t) estimates (Supplementary Fig. 2). However, we did not want to use the maximum daily number of deaths as a selection criterion, because this could lead to selection of counties based on data from a single day. It would also involve some circularity, because the information obtained, r(t), would be directly related to the criterion used to select datasets. Therefore, we used the threshold of 100 cumulative deaths. The District of Columbia was treated as a county. Also, because the New York Times dataset aggregated the five boroughs of New York City, we treated them as a single county. For counties with fewer than 100 deaths, we aggregated mortality to the state level to create a single time series. For thirteen states (AK, DE, HI, ID, ME, MT, ND, NH, SD, UT, VM, WV, and WY), the aggregated time series did not contain 100 or more deaths and were therefore not analyzed.
Data selection and handling: explanatory county-level variables
County-level variables were collected from several public data sources36,37,38,39,40,41,42. We selected socio-economic variables a priori in part to represent a broad set of population characteristics.
Time series analysis: time series model
We used a time-varying autoregressive model15,56 designed explicitly to estimate the rate of increase of a variable using nonlinear, state-dependent error terms16. We assume in our analyses that the susceptible proportion of the population represented by a time series is close to one, and therefore there is no decrease in the infection rate caused by a pool of individuals who were infected, recovered, and were then immune to further infection.
The model is

$$xleft( t right) = rleft( {t-1} right) + xleft( {t-1} right)$$
(1a)

$$rleft( t right) = rleft( {t-1} right) + omega_rleft( t right)$$
(1b)

$$Dleft( t right) = {mathrm{exp}}(xleft( t right) + phi left( t right))$$
(1c)

Here, x(t) is the unobserved, log-transformed value of daily deaths at time t, and D(t) is the observed count that depends on the observation uncertainty described by the random variable ϕ(t). Because a few of the datasets that we analyzed had zeros, we replaced zeros with 0.5 before log-transformation. The model assumes that the death count increases exponentially at rate r(t), where the latent state variable r(t) changes through time as a random walk with ωr(t) ~ N(0, σ2r). We assume that the count data follow a quasi-Poisson distribution. Thus, the expectation of counts at time t is exp(x(t)), and the variance is proportional to this expectation.
We fit the model using the extended Kalman filter to compute the maximum likelihood57,58. In addition to the parameters σ2r and σ2ϕ, we estimated the initial value of r(t) at the start of the time series, r0, and the initial value of x(t), x0. The estimation also requires terms for the variances in x0 and r0, which we assumed were zero and σ2r, respectively. In the validation using simulated data (Supplementary Methods: Simulation model), we found that the estimation process tended to absorb σ2r to zero too often. To eliminate this absorption to zero, we imposed a minimum of 0.02 on σ2r.
Time series analysis: parametric bootstrapping
To generate approximate confidence intervals for the time-varying estimates of r(t) (Eq. 1b), we used a parametric bootstrap designed to simulate datasets with the same characteristics as the real data that are then refit using the autoregressive model. We used bootstrapping to obtain confidence intervals, because an initial simulation study showed that standard methods, such as obtaining the variance of r(t) from the Kalman filter, were too conservative (the confidence intervals too narrow) when the number of counts was small. Furthermore, parametric bootstrapping can reveal bias and other features of a model, such as the lags we found during model fitting (Supplementary Fig. 1a, b).
Changes in r(t) consist of unbiased day-to-day variation and the biased deviations that lead to longer-term changes in r(t). The bootstrap treats the day-to-day variation as a random variable while preserving the biased deviations that generate longer-term changes in r(t). Specifically, the bootstrap was performed by calculating the differences between successive estimates of r(t), Δr(t) = r(t) – r(t-1), and then standardizing to remove the bias, Δrs(t) = Δr(t) – E[Δr(t)], where E[] denotes the expected value. The sequence Δrs(t) was fit using an autoregressive time-series model with time lag 1, AR(1), to preserve any shorter-term autocorrelation in the data. For the bootstrap, a new time series was simulated from this AR(1) model, Δρ(t), and then standardized, Δρs(t) = Δρ(t) – E[Δρ(t)]. The simulated time series for the spread rate was constructed as ρ(t) = r(t) + Δρs(t)/21/2, where dividing by 21/2 accounts for the fact that Δρs(t) was calculated from the difference between successive values of r(t). A new time series of count data, ξ(t), was then generated using equation 1 with the parameters from fitting the data. Finally, the statistical model was fit to the reconstructed ξ(t). In this refitting, we fixed the variance in r(t), σ2r, to the same value as estimated from the data. Therefore, the bootstrap confidence intervals are conditional of the estimate of σ2r.
Time series analysis: calculating R0
We derived estimates of R(t) directly from r(t) using the Dublin-Lotka equation21 from demography. This equation is derived from a convolution of the distribution of births under the assumption of exponential population growth. In our case, the “birth” of COVID-19 is the secondary infection of susceptible hosts leading to death, and the assumption of exponential population growth is equivalent to assuming that the initial rate of spread of the disease is exponential, as is the case in equation 1. Thus,

$$Rleft( t right) = 1/mathop {sum}nolimits_{_tau} {{mathrm{e}}^{ – r(t)}} tau p(tau)$$
(2)

where p(τ) is the distribution of the proportion of secondary infections caused by a primary infection that occurred τ days previously. We used the distribution of p(τ) from Li et al.59 that had an average serial interval of T0 = 7.5 days; smaller or larger values of T0, and greater or lesser variance in p(τ), will decrease or increase R(t) but will not change the pattern in R(t) through time. Note that the uncertainty in the distribution of serial times for COVID-19 is a major reason why we focus on estimating r0, rather than R0: the estimates of r0 are not contingent on time distributions that are poorly known. Computing R(t) from r(t) also does not depend on the mean or variance in time between secondary infection and death. We report values of R(t) at dates that are offset by 18 days, the average length of time between initial infection and death given by Zhou et al.60.
Time series analysis: Initial date of the time series
Many time series consisted of initial periods containing zeros that were uninformative. As the initial date for the time series, we chose the day on which the estimated daily death count exceeded 1. To estimate the daily death count, we fit a Generalized Additive Mixed Model (GAMM) to the death data while accounting for autocorrelation and greater measurement error at low counts using the R package mgcv61. We used this procedure, rather than using a threshold of the raw death count, because the raw death count will include variability due to sampling small numbers of deaths. Applying the GAMM to “smooth” over the variation in count data gives a well-justified method for standardizing the initial dates for each time series.
Time series analysis: validation
We performed extensive simulations to validate the time-series analysis approach (Supplementary Methods: Simulation model).
Regression analysis for r 0
We applied a Generalized Least Squares (GLS) regression model to explain the variation in estimates of r0 from the 160 county and county-aggregate time series:

$$r_0 = b_0 + b_1start.date + b_2logleft( {pop.size} right) + b_3pop.den^{0.25} + varepsilon$$
(3)

where start.date is the Julian date of the start of the time series, log(pop.size) and pop.den0.25 are the log-transformed population size and 0.25 power-transformed population density of the county or county-aggregate, respectively, and ε is a multivariate Gaussian random variable with covariance matrix σ2Σ. We used the transforms log(pop.size) and pop.den0.25 to account for nonlinear relationships with r0; these transforms give the highest maximum likelihood of the overall regression. The covariance matrix contains a spatial correlation matrix of the form C = uI + (1–u)S(g) where u is the nugget and S(g) contains elements exp(−dij/g), where dij is the distance between spatial locations and g is the range62. To incorporate differences in the precision of the estimates of r0 among time series, we weighted by the vector of their standard errors, s, so that Σ = diag(s) * C * diag(s), where * denotes matrix multiplication. With this weighting, the overall scaling term for the variance, σ2, will equal 1 if the residual variance of the regression model matches the square of the standard errors of the estimates of r0 from the time series. We fit the regression model with the function gls() in the R package nlme63.
To make predictions for new values of r0, we used the relationship

$$hat e_{mathrm{i}} = bar e + {mathbf{v}}_{mathbf{i}} ast ,{mathbf{V}}^{ – 1}(epsilon_i – bar e)$$
(4)

where ει is the GLS residual for data i, (hat e)i is the predicted residual, (bar e) is the mean of the GLS residuals, V is the covariance matrix for data other than i, and vi is a row vector containing the covariances between data i and the other data in the dataset64. This equation was used for three purposes. First, we used it to compute R2pred for the regression model by removing each data point, recomputing (hat e)i, and using these values to compute the predicted residual variance23. Second, we used it to obtain predicted values of r0, and subsequently R0, for the 160 counties and county-aggregates for which r0 was also estimated from time series. Third, we used equation (4) to obtain predicted values of r0, and hence predicted R0, for all other counties. We also calculated the variance of the estimates from64

$$hat v_{mathrm{i}} = sigma^2-{mathbf{v}}_{mathbf{i}} ast ,{mathbf{V}}^{ – 1} ast v_i^{mathbf{t}}$$
(5)

Predicted values of R0 were mapped using the R package usmap65.
Regression analysis for SARS-CoV-2 effects on r0
The GISAID metadata27 for SARS-CoV-2 contains the clade and state-level location for strains in the USA; strains G, GH, and GR contain the G614 mutation. For each state, we limited the SARS-CoV-2 genomes to those collected no more than 30 days following the onset of outbreak that we used as the starting point for the time series from which we estimated r0; from these genomes (totaling 5290 from all states), we calculated the proportion that had the G614 mutation. We limited the analyses to the 28 states that had five or more genome samples. For each state, we selected the estimates of r0 from the county or county-aggregate representing the greatest number of deaths. We fit these estimates of r0 with the weighted Least Squares (LS) model as in equation (3) with additional variables for strain. Figure 3 was constructed using the R packages usmap65 and scatterpie66.
Statistics and reproducibility
The statistics for this study are summarized in the preceding sections of the “Methods”. No experiments were conducted, so experimental reproducibility is not an issue. Nonetheless, we repeated analyses using alternative datasets giving county-level characteristics, and also an alternative dataset on SARS-CoV-2 strains (Supplementary Methods: Analysis of Nextstrain metadata of SARS-CoV-2 strains), and all of the conclusions were the same.
Reporting summary
Further information on research design is available in the Nature Research Reporting Summary linked to this article. More

As a wildlife ecologist, I study many species, but the platypus (Ornithorhynchus anatinus) is my speciality. I’ve worked with these duck-billed monotremes, which are indigenous to eastern Australia, for 13 years. So much remains unknown about them — as solitary animals, they are very hard to capture. The males have a venomous spur on each of their hind feet, so I always hold them by the tail. Luckily, I’ve never been spurred.
At the environmental consultancy firm Cesar Australia, we’ve focused on whether platypuses’ numbers are rising or falling. Their aquatic habitat has shrunk by 25% in the past 30 years — partly owing to reduced rainfall from climate change and partly to residential and industrial development, leaving less water, which is of poorer quality.
To monitor the population’s size and health, we do overnight trapping surveys — typically during the spring breeding season and in the autumn, when the babies emerge. For each survey, we set up nets along 5–10 kilometres of waterway and then drive round checking them through the night. It’s a long, sleepless process that usually yields just one or two catches. In the 15 minutes or so for which each animal is out of the water, we weigh and measure it. If it’s carrying an identification chip, we know we’ve caught it before. If not, we implant one in the animal. In this picture, I am releasing a platypus back into a stream outside Melbourne.
In the past five years, we’ve also been sampling environmental DNA for signatures of platypuses’ presence. It’s been a game-changer in terms of mapping their distribution. We can take a water sample, extract all the DNA and confidently identify whether platypus DNA is present. We’ve surveyed more than 2,000 sites spanning 500,000 square kilometres, enabling us to measure at scales never before possible.
We’ve also developed a citizen-science platform through which people can submit platypus sightings. This helps us still further. More

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