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A probabilistic formalism, relying on Bayes’ theorem and linear Gaussian inversion, is adapted, so that a monochromatic problem can be investigated. The formalism enables an objective test in probabilistic terms of the quantities and model concepts involved in the problem at hand. With this formalism, an amplitude (linear parameter), a frequency (non-linear parameter) and a hyperparameter of the Gaussian amplitude prior are inferred jointly given simulated data sets with Gaussian noise contributions. For the amplitude, an analytical normal posterior follows which is conditional on the frequency and the hyperparameter. The remaining posterior estimates the frequency with an uncertainty of MHz, while the convolution of a standard approach would achieve an uncertainty of some GHz. This improvement in the estimation is investigated analytically and numerically, revealing for instance the positive effect of a high signal-to-noise ratio and/or a large number of data points. As a fixed choice of the hyperparameter imposes certain results on the amplitude and frequency, this parameter is estimated and, thus, tested for plausibil i ty as well. From abstract point of view, the model posterior is investigated as well.

Fourier transformation tools are used to obtain information about spectra for a given data set. As any data has an uncertainty, Fourier transformation techniques can be supported by probabilistic theory captured by Bayes’ theorem [

A common and misleading assumption in the field is that the Nyquist theorem determines the spectral band limitation. However, as the Nyquist frequency follows from finite data sampling, it can only be an upper limit but cannot be an estimator for the limit caused actually by a source and/or diagnostic throughput. A similar reasoning applies to the lower spectral limit. After a basic formalism has been derived, it could be shown in Ref. [

For many applied analysis schemes, certain model assumptions and their implications are not tested but assumed to be valid with infinite trust. From scientific point of view, any analysis scheme should be tested given simulated noisy data for which all model assumptions are clear. Then, analysis results and model assumptions can be investigated which is achieved objectively by a probabilistic ansatz. If this can be carried out analytically, valuable information is available when only actual measured data are given for a scientific problem.

The problem of a monochromatic source is a good example to show the powerfulness of Bayesian inference and to test model assumptions. In this work, the formalism derived in Ref. [

The formalism derived in Ref. [

p ( S 1 , f 1 , σ S 1 , P r | D ) = p ( D | S 1 , f 1 ) p ( D ) p ( S 1 | σ S 1 , P r ) p ( σ S 1 , P r ) p ( f 1 ) (1)

with the joint posterior on the left-hand, the likelihood p ( D | S 1 , f 1 ) devided by the evidence, the amplitude prior p ( S 1 | σ S 1 , P r ) , the prior for the hyperparameter, and the prior for the frequency p ( f 1 ) .

The amplitude S 1 maps linearly to the data domain via the vector M 1 = M ( f 1 ) dependent on the frequency in a non-linear manner. The data D with N D entries are assumed to be acquired independently, and the measurement uncertainty of each data point follows a normal distribution with standard deviation σ D . With these assumptions, the Gaussian likelihood becomes

p ( D | S 1 , f 1 ) = exp [ − 1 2 ( D − M 1 S 1 ) T Σ D − 1 ( D − M 1 S 1 ) ] ( 2 π ) N D / 2 | Σ D | 1 / 2 (2)

with the covariance matrix Σ D = σ D 2 δ i j . The amplitude prior takes the form

p ( S 1 | σ S 1 , P r ) = exp [ − 1 2 S 1 2 / σ S 1 , P r 2 ] ( 2 π ) 1 / 2 σ S 1 , P r (3)

for vanishing prior mean and variance σ S 1 , P r 2 . After some analytical operations, one obtaines the Gaussian amplitude posterior

p ( S 1 | f 1 , σ S 1 , P r , D ) = exp [ − 1 2 ( S 1 − S 1, P o ) σ S 1 , P o − 2 ( S 1 − S 1, P o ) ] ( 2 π ) 1 / 2 σ S 1 , P o (4)

conditional on f 1 and σ S 1 , P r . In the above equation, the posterior variance is given by σ S 1 , P o 2 = ( M 1 T M 1 σ D − 2 + σ S 1 , P r − 2 ) − 1 , and the posterior mean reads S 1, P o = σ S 1 , P o 2 M 1 T D / σ D 2 .

From the remaining terms, the so-called settings posterior

p ( f 1 , σ S 1 , P r | D ) = 1 K ′ σ S 1 , P o σ S 1 , P r exp [ 1 2 S 1, P o 2 / σ S 1 , P o 2 ] p ( σ S 1 , P r ) p ( f 1 ) (5)

is derived which can be interpreted as Ockham’s razor with respect to S 1 . This settings posterior has no general analytical solution, and, thus, the normalisation constant K ′ cannot be stated further. For the remainder of the paper, p ( σ S 1 , P r ) = 1 / Δ σ and p ( f 1 ) = 1 / Δ f are taken as uniform distributions. By this choice, K ′ = K p ( σ S 1 , P r ) p ( f 1 ) can be used to investigate analytically and numerically the quantity

p × K = σ S 1 , P o σ S 1 , P r exp [ 1 2 S 1, P o 2 / σ S 1 , P o 2 ] . (6)

Finally, the evidence can be identified as

p ( D ) = K Δ σ Δ f exp [ − 1 2 D T Σ D − 1 D ] ( 2 π ) N D / 2 | Σ D | 1 / 2 . (7)

The constant K / ( Δ σ Δ f ) varies with the chosen model H, including likelihood, priors, and the prior domains for f 1 and σ S 1 , P r . Hence, this constant is linked to the model posterior p ( H | D ) which is of importance, when the model is further abstracted, or a comparison with an alternative model is carried out.

For a monochromatic even source, data sets with different noise levels are simulated, and the formalism derived in the previous section is applied. To explain the results found, analytical and numerical investigations are carried out. While this problem is treated abstractly in the following, the main results will be presented for two examples with low and high noise contribution to ease the presentation.

(A) Simulated Data: Two Examples

Data sets are modelled for the real-world interferometer found in Ref. [

(B) Application of Formalism

The linear mapping of the amplitude parameter is given by M 1, i = 2 cos ( 2 π f 1 x i / c ) , yielding M 1 . For the two simulated data sets,

The settings posterior p ( f 1 , σ S 1 , P r | D ) is proportional to the quantity p × K which takes very large values even on logarithmic scale and, thus, ln p × K is used in the following. For the two cases, ln p × K is shown in

relevant domains of f 1 and σ S 1 , P r by

(C) Analytical and Numerical Investigations

To have further inside in the results, the conditional amplitude posterior and p × K are analytically investigated. In order to do so, the noise contribution in the data is neglected, meaning that D i is replaced with the noise-less interferogram V i = 2 S cos ( 2 π f x i / c ) . This allows the tracking of the dependencies of the posteriors on the original amplitude S and frequency f, noise level σ D , and number of data points N D . Furthermore, a second factorisation p ( f 1 , σ S 1 , P r | D ) = p ( f 1 | σ S 1 , P r , D ) p ( σ S 1 , P r | D ) can be justified. Parallel to this investigation, numerics and conditional posteriors are presented.

1) Conditional Amplitude Posterior

Starting point is the variance σ S 1 , P o 2 of the conditional posterior p ( S 1 | f 1 , σ S 1 , P r , D ) (see Equation (4)). Formally, one finds for the inverse

σ S 1 , P o − 2 = M 1 T M 1 σ D 2 + 1 σ S 1 , P r 2 . (8)

As long as the spatial sampling is well enough, one can use the approximation

M 1 T M 1 = 4 ∑ i = 1 N D cos ( 2 π f 1 c x i ) cos ( 2 π f 1 c x i ) ≈ 2 N D [ 1 + cos ( 2 π f 1 c L S S ) sinc ( 2 π f 1 c L ) ] , (9)

as shown by Equation (A1), to get the posterior variance

σ S 1 , P o 2 = σ D 2 / ( 2 N D ) 1 + cos ( 2 π f 1 c L S S ) sinc ( 2 π f 1 c L ) + σ D 2 2 N D σ S 1 , P r 2 . (10)

This variance increases quadratically with the noise level and reduces with the number of data samples. Thus, more data points per spatial unit can compensate the noise contribution, at least partly. However, this assumes that the noise is independent of the data sampling.

The modulated sinc function in the denominator of Equation (10) depends on f 1 , L and L S S for which the interferogram is available. The sinc function becomes unity and vanishes for small and large frequencies, respectively. For instance, since both data examples share the same spatial domain, the sinc function is close to 0 above about 20 GHz. The amplitude prior influences σ S 1 , P o 2 , when the ratio σ D 2 / ( 2 N D σ S 1 , P r 2 ) becomes larger than 1 which is obtained for a large noise level, a small value of σ S 1 , P r and/or few data points.

For the posterior mean,

S 1, P o ( f 1 ) = σ S 1 , P o 2 σ D 2 M 1 T V = 2 ∑ i = 1 N D V i cos ( 2 π f 1 c x i ) / ( 2 N D ) 1 + cos ( 2 π f 1 c L S S ) sinc ( 2 π f 1 c L ) + σ D 2 2 N D σ S 1 , P r 2 (11)

follows, leaving aside the noise contribution but the uncertainty on the mean is still captured by σ S 1 , P o . With the approximation (A1) and the trigonometric identity (B1), one can resolve

S 1, P o ( f 1 ) = S cos ( π f − f 1 c L S S ) sinc ( π f − f 1 c L ) + cos ( π f + f 1 c L S S ) sinc ( π f + f 1 c L ) 1 + cos ( 2 π f 1 c L S S ) sinc ( 2 π f 1 c L ) + σ D 2 2 N D σ S 1 , P r 2 . (12)

Indeed, one finds the original amplitude S 1 , P o = S for f 1 = f and prior values with the condition σ D 2 / ( 2 N D ) ≪ σ S 1 , P r 2 . Furthermore, due to cos ( π f − f 1 c L ) sinc ( π f − f 1 c L S S ) there is a peaking of S 1, P o ( f 1 ) in the vicinity of f 1 ≈ f with the FWHM determined by both trigonometric arguments. Away from f, the oscillating amplitude drops like 1 / ( f − f 1 ) due to the sinc function which can be seen in

2) Conditional Posterior for f_{1}

Now the information is available to investigate p ( f 1 , σ S 1 , P r | D ) . Neglecting the noise, the triple product in the exponent of the exponential in p × K reads

1 2 S 1, P o σ S 1 , P o − 2 S 1, P o ≈ 1 2 M 1 T V σ D 2 σ S 1 , P o 2 M 1 T V σ D 2 = 1 2 σ S 1 , P o 2 4 N D 2 S 2 σ D 4 [ cos ( π f − f 1 c L S S ) sinc ( π f − f 1 c L ) + cos ( π f + f 1 c L S S ) sinc ( π f + f 1 c L ) ] 2 . (13)

The global maximum is close to f 1 = f , and, assuming a sufficiently large frequency f, one gets the approximation

[ cos ( π f − f 1 c L S S ) sinc ( π f − f 1 c L ) ] 2 ≈ [ 1 − ( π f − f 1 c L S S ) 2 − 1 3 ( π f − f 1 c L ) 2 ] (14)

when the Taylor series expansions sinc ( z ) ≈ 1 − z 2 / 3 ! and cos ( z ) ≈ 1 − z 2 / 2 are used. Then,

1 2 S 1, P o σ S 1 , P o − 2 S 1, P o ≈ 1 2 σ S 1 , P o 2 4 N D 2 S 2 σ D 4 [ 1 − ( f − f 1 ) 2 ( π c ) 2 ( L S S 2 + 1 3 L 2 ) ] (15)

remains, and the term which is independent on f − f 1 takes very large numerical values and is treated in more detail in the next subsection. The remaining term can be rewritten by a quadratic exponent of a Gaussian with posterior mean f 1, P o = f and square root of the variance

σ f 1 , P o = 1 σ S 1 , P o σ D 2 2 N D S 1 π c L ( 3 3 L S S 2 / L 2 + 1 ) 1 / 2 ≈ σ D S N D 1 / 2 1 π c L ( 3 2 1 3 L S S 2 / L 2 + 1 ) 1 / 2 , (16)

using Equation (10) for large frequencies and σ S 1 , P r 2 ≫ σ D 2 / ( 2 N D ) . This uncertainty increases with the noise but decreases with the square root of the number of data points per spatial unit, the signal level and the spatial domain covered. For the data set examples, one inserts L = 31.56 mm, L S S = 21.28 mm, N D = 789 , σ D / S = 0.1 and 2 to find the astonishing numbers σ f 1 , P o = 8.6 and 171.9 MHz, respectively. This explains the narrowness of p ( f 1 , σ S 1 , P r | D ) with respect to f 1 as shown in

In fact, σ S 1 , P o changes with f 1 on a GHz, but the peaking in f 1 is of the order of MHz. Hence, the factorisation p ( f 1 , σ S 1 , P r | D ) ≈ p ( f 1 | σ S 1 , P r , D ) p ( σ S 1 , P r | D ) with the conditional Gaussian posterior

p ( f 1 | σ S 1 , P r , D ) = exp [ − 1 2 ( f 1 − f 1, P o ) 2 σ f 1 , P o 2 ] ( 2 π ) 1 / 2 σ f 1 , P o (17)

and

p ( σ S 1 , P r | D ) = σ S 1 , P o σ S 1 , P r exp [ 1 2 σ S 1 , P o 2 4 N D 2 S 2 σ D 4 ] K ( 2 π ) 1 / 2 σ f 1 , P o (18)

is justified. Thereby, σ S 1 , P o needs to be taken at f 1 = f 1, P o . The remaining probability p ( σ S 1 , P r | D ) is investigated in the next subsection.

Numerics

Because the variance σ S 1 , P o 2 changes in f 1 on GHz scale but

exp [ 1 2 S 1, P o σ S 1 , P o − 2 S 1, P o ] on MHz scale, only the exponential needs to be taken into account for the determination of f 1, P o and σ f 1 , P o . Furthermore, one faces large numbers, and, hence, the exponent is dealt with directly.

At a given σ S 1 , P r , the frequency at which the maximum of the exponent occurs is identified as the posterior mean f 1, P o ( σ S 1 , P r ) . The posterior standard deviation σ f 1 , P o ( σ S 1 , P r ) is obtained when the ratio of the exponent to its maximum at f 1, P o reads 1 2 S 1, P o σ S 1 , P o − 2 S 1, P o / max ( 1 2 S 1, P o σ S 1 , P o − 2 S 1, P o ) = − 1 / 2 . For both data sets, the ratios are shown in

3) Posterior for σ 1 , P r

After the factorisation, one combines Equations (16) and (18) like

p ( σ S 1 , P r | D ) = P ′ K = σ S 1 , P o σ S 1 , P r exp [ 1 2 σ S 1 , P o 2 4 N D 2 S 2 σ D 4 ] K ( 2 π ) 1 / 2 σ f 1 , P o = P K σ D 2 N D S c L 3 2π L 2 3 L S S 2 + L 2 (19)

with the important term

P = exp [ 1 2 σ S 1 , P o 2 4 N D 2 S 2 σ D 4 ] σ S 1 , P r . (20)

The exponent of P becomes

σ S 1 , P r 2 N D S 2 σ D 2 σ S 1 , P r 2 [ 1 + cos ( 2 π f 1, P o c L S S ) sinc ( 2 π f 1, P o c L ) ] + σ D 2 2 N D (21)

which is monotonically rising in σ S 1 , P r from 0 to about N D S 2 / σ D 2 . In case σ S 1 , P r 2 > σ D 2 / ( 2 N D ) and S / σ D = 1 , the number of data points dominates. For the two examples with N D = 789 , large numbers exp ( 78900 ) and exp ( 197 ) are at hand as could be seen in

As the exponent vanishes for σ S 1 , P r = 0 , 1 / σ S 1 , P r and, thus, P and P ′ diverge. Hence, this global maximum should be excluded by a proper choice for the prior of σ S 1 , P r . Doing so, P has the global maximum P G M a x at

σ S 1 , P r , G M a x 2 = S 2 2 − σ D 2 2 N D + S 2 2 ( 1 − 2 σ D 2 N D S 2 ) 1 / 2 (22)

supposed a large f 1, P o . Interestingly, N D reduces the noise impact drastically, so that even for elevated noise levels σ D ≈ S , the maximum is still given by S with minor corrections.

The uniform prior p ( σ S 1 , P r ) may be finite for the domain σ S 1 , P r , L ≤ σ S 1 , P r , U , and the maximum of P in this domain is denoted by P M a x . Then, the peaking can be made more obvious by

P P M a x = σ S 1 , P r , M a x σ S 1 , P r exp [ − 1 2 σ S 1 , P r , M a x 2 − σ S 1 , P r 2 σ S 1 , P r 2 + σ D 2 2 N D S 2 σ S 1 , P r , M a x 2 + σ D 2 2 N D ] (23)

after some algebra for a large f 1, P o . With the normalisation of the above equation

∫ σ S 1 , P r , L σ S 1 , P r , U P P M a x d σ S 1 , P r = K σ S 1 , P r ( σ S 1 , P r , L , σ S 1 , P r , U ) , (24)

the posterior becomes

p ( σ S 1 , P r | D ) = 1 K σ S 1 , P r P P M a x . (25)

In case, the prior is finite for the domain σ S 1 , P r , L ≤ σ S 1 , P r , G M a x ≤ σ S 1 , P r , U , so that P M a x = P G M a x , one can set σ S 1 , P r , M a x 2 = S 2 − σ D 2 / ( 2 N D ) . This reveals the simple forms

P G M a x = exp [ N D S 2 σ D 2 − 1 2 ] S 2 − σ D 2 2 N D (26)

and

P P G M a x = S 2 − σ D 2 2 N D σ S 1 , P r exp [ 1 2 ( 1 − S 2 σ S 1 , P r 2 + σ D 2 2 N D ) ] (27)

of Equation (23). With the above expression and Equation (19), the constant

K = P ′ G M a x K σ S 1 , P r = σ D 2 N D S c L 3 2 π L 2 3 L S S 2 + L 2 exp [ N D S 2 σ D 2 − 1 2 ] S 2 − σ D 2 2 N D K σ S 1 , P r (28)

follows with proper unit Hz. K depends in a complicated way on the noise level, the signal, the number of data points, the spatial domain for which the data is acquired, and the limits σ S 1 , P r , L and σ S 1 , P r , U . For the two examples, one gets ln ( P ′ G M a x / Hz ) = 78910 ( σ D = 0.05 ) and ln ( P ′ G M a x / Hz ) = 214 ( σ D = 1 ).

Numerics

Starting point is Equation (19) which is rewritten like

ln p ( σ S 1 , P r | D ) + ln K = ln P ′ = ln ( σ S 1 , P o σ S 1 , P r max ( exp [ 1 2 S 1, P o σ S 1 , P o − 2 S 1, P o ] ) ( 2 π ) 1 / 2 σ f 1 , P o ) (29)

to handle large numerical values. Thereby, σ S 1 , P o is evaluated at f 1 = f 1 , P o . The far right-hand side of the above equation is numerically available, and the maximum ln P ′ M a x at σ S 1 , P r , M a x and the normalisation K σ S 1 , P r can be determined for the domain from σ S 1 , P r , L to σ S 1 , P r , U .

Choosing for the moment σ S 1 , P r , L = 10 − 2 and σ S 1 , P r , U = 10 3 , so that the global maximum P ′ G M a x is included, the above formalism yields ln ( P ′ G M a x / Hz ) = 76861 (213) and K σ S 1 , P r = 6.32 (6.36) for the data set simulated with σ D = 0.05 (1). For both examples, the order of magnitude of ln ( P ′ G M a x / Hz ) is well described by Equation (26).

For the data sets, both P ′ G M a x are located at σ S 1 , P r , G M a x = 0.499 and 0.501 which is in the vicinity of S (see

4) Model Posterior

The posterior of the model with uniform priors for σ S 1 and f 1 is given by

p ( H | D ) = P ′ M a x K σ S 1 , P r ( σ S 1 , P r , U − σ S 1 , P r , L ) ( f 1, P r , U − f 1, P r , L ) . (30)

This posterior weighs the constants obtained with respect to the spectral and amplitude domains. Hence, p ( H | D ) increases with the constants in the numerator and smaller domains in the denominator. When one investigates

(D) Joint Posterior and Marginal Posteriors for S_{1} and f_{1}

Because the model posterior peaks at σ S 1 , P r , G M a x , the marginal posterior for f 1 is simply given by the Gaussian p ( f 1 | σ S 1 , P r = σ S 1 , P r , G M a x , D ) with posterior mean f 1, P o and standard deviation σ f 1 , P o (see Section III C2 and

p ( S 1 , f 1 | σ S 1 , P r , G M a x , D ) = p ( S 1 | f 1 , σ S 1 , P r , G M a x , D ) p ( f 1 | σ S 1 , P r , G M a x , D ) (31)

can be evaluated and is shown for both data sets in

The marginal posterior for S 1 is obtained by the marginalisation

p ( S 1 | σ S 1 , P r , G M a x , D ) = ∫ p ( S 1 , f 1 | σ S 1 , P r , G M a x , D ) d f 1 . (32)

The above integration is performed numerically, and the results are shown in

Evaluating the mean and standard deviation from these samples approximates p ( S 1 | σ S 1 , P r , G M a x , D ) by a Gaussian. In doing so for the data set with σ D = 0.05 (1), the marginal posterior mean S 1, P o , M a = 0.4998 (0.505) and standard deviation 0.0013 (0.026) are obtained. Both normal distributions are good approximations for p ( S 1 | σ S 1 , P r , G M a x , D ) (see

The presented work investigates a monochromatic problem with an adapted version of Bayes’ theorem to infer an amplitude parameter, a frequency parameter and a hyperparameter which acts on the amplitude prior only. The amplitude at a given frequency is a parameter which has a linear mapping to the data domain. If the measurements have a normal noise contribution (Gaussian likelihood) and the amplitude prior is chosen to be Gaussian, an analytical linear inversion technique is enabled. The amplitude posterior, being conditional on the frequency and prior information, has similiarities with the result of a conventional Fourier transformation analysis approach. However, the estimation of the frequency reveals a well localised domain which is in accordance with the measured data. The corresponding posterior for the frequency is orders of magnitude narrower than the width of the convolution function used by conventional approaches to estimate the spectral resolution. The prior information captured by one hyperparameter is tested as well, and its posterior peaks over a one order of magnitude and is very robust for significantly different noise levels on the data.

The findings for the monochromatic problem are investigated analytically. This reveals, for instance, the influence of the signal-to-noise ratio, the amount of data points and the spatial domains covered by the measurements on the results. The presented approach can be followed to examine more complex problems which involve more than one frequency and one amplitude and other diagnostic imperfections like a variable offset. Since this offset has a linear correspondence to the data domain, it would enter in the conditional amplitude posterior. Any non-linear parameter and additional hyperparameter would be estimated by the joint settings posterior specific for a certain problem. In doing so, a profound understanding of model implications could be established in analytical terms for applications relying on Fourier transformations. This is essential for designing a diagnostic for a given problem with somewhat known signal-to-noise ratio and hardware limitations, or for comparing results of different models in an objective way when a data set is given and, for example, the number of contributing frequencies is unknown.

This work has been carried out within the framework of the EURO fusion Consortium and has received funding from the Euratom research and training programme 2014-2018 and 2019-2020 under grant agreement No. 633053. The views and opinions expressed herein do not necessarily reflect those of the European Commission.

The authors declare no conflicts of interest regarding the publication of this paper.

Schmuck, S. and Svensson, J. (2019) Analytical and Numerical Investigations of Probabilistic Monochromatic Problem. Journal of Applied Mathematics and Physics, 7, 793-808. https://doi.org/10.4236/jamp.2019.74054

The spatial grid x i with i ∈ [ 1, N ] may be given by the set x i = i Δ x + x 0 of uniformly distributed locations separated by the constant increment Δ x , so that x 1 = − L D S and x N = − L D S + L . Then, the sum

∑ i = 1 N cos ( 2 π f 1 c x i ) cos ( 2 π f 2 c x i ) ≈ 1 2 Δ x ∫ − L D S − L D S + L [ cos ( 2 π f − c x ) + cos ( 2 π f + c x ) ] d x = 1 2 Δ x [ c 2 π f − sin ( 2 π f − c x ) ] − L D S − L D S + L + 1 2 Δ x [ c 2 π f + sin ( 2 π f + c x ) ] − L D S − L D S + L = N 2 cos ( π f − c L S S ) sinc ( π f − c L ) + N 2 cos ( π f + c L S S ) sinc ( π f + c L ) (A1)

is approximated by an integral over the spatial domain using the trigonometric identities (B1) and (B2). The approximation reveals two modulated sinc functions with a spatial and a spectral dependence. The spectral dependence is due to the sum and difference frequencies f − = f 1 − f 2 and f + = f 1 + f 2 , respectively. Spatially, the modulation depends on the length L S S = L − 2 L D S of single-sided domain which is present on side only. The spatial dependence of the sinc function is given by the length L of the total domain covered.

The trigonometric identities are

cos ( x ) cos ( y ) = 1 2 [ cos ( x − y ) + cos ( x + y ) ] , (B1)

sin ( x ) − sin ( y ) = 2 cos ( x + y 2 ) sin ( x − y 2 ) . (B2)