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I am a student working with time-series data which we feed into a neural network for classification (my task is to build and train this NN).
We're told to use a band-pass filter of 10 Hz to 150 Hz since anything outside that is not interesting.
After applying the band-pass, I've also down-sampled the data to 300 samples per second (originally it was 768 Hz). My understanding of the Shannon Nyquist sampling theorem is that, after applying the band-pass, any information in the data will be perfectly preserved at this sample-rate.
However, I got into a discussion with my supervisor who claimed that 300 Hz might not be sufficient even if the signal was band-limited. She says that it is only the minimum sample rate, not necessarily the best sample rate.
My understanding of the sampling theorem makes me think the supervisor is obviously wrong, but I don't want to argue with my supervisor, especially in case I'm actually the one who has misunderstood.
Can anyone help to confirm my understanding or provide some clarification? And how should I take this up with my supervisor (if at all).
The Nyquist-Shannon theorem states that the sampling frequency should at-least be twice of bandwidth, i.e.,
fs > 2B
So, this is the minimal criteria. If the sampling frequency is less than 2B then there will be aliasing. There is no upper limit on sampling frequency, but more the sampling frequency, the better will be the reconstruction.
So, I think your supervisor is right in saying that it is the minimal condition and not the best one.
Actually, you and your supervisor are both wrong. The minimum sampling rate required to faithfully represent a real-valued time series whose spectrum lies between 10 Hz and 150 Hz is 140 Hz, not 300 Hz. I'll explain this, and then I'll explain some of the context that shows why you might want to "oversample", as it is referred to (spoiler alert: Bailian-Low Theorem). The supervisor is mixing folklore into the discussion, and when folklore is not properly-contexted, it tends to telephone tag into fakelore. (That's a common failing even in the peer-reviewed literature, by the way). And there's a lot of fakelore, here, that needs to be defogged.
For the following, I will use the following conventions.
There's no math layout on Stack Overflow (except what we already have with UTF-8), so ...
a^b denotes a raised to the power b.
∫_I (⋯x⋯) dx denotes an integral of (⋯x⋯) taken over all x ∈ I, with the default I = ℝ.
The support supp φ (or supp_x φ(x) to make the "x" explicit) of a function φ(x) is the smallest closed set containing all the x-es for which φ(x) ≠ 0. For regularly-behaving (e.g. continuously differentiable) functions that means a union of closed intervals and/or half-rays or the whole real line, itself. This figures centrally in the Shannon-Nyquist sampling theorem, as its main condition is that a spectrum have bounded support; i.e. a "finite bandwidth".
For the Fourier transform I will use the version that has the 2π up in the exponent, and for added convenience, I will use the convention 1^x = e^{2πix} = cos(2πx) + i sin(2πx) (which I refer to as the Ramanujan Convention, as it is the convention I frequently used in my previous life oops I mean which Ramanujan secretly used in his life to make the math a whole lot simpler).
The set ℤ = {⋯, -2, -1, 0, +1, +2, ⋯ } is the integers, and 1^{x+z} = 1^x for all z∈ℤ - making 1^x the archetype of a periodic function whose period is 1.
Thus, the Fourier transform f̂(ν) of a function f(t) and its inverse are given by:
f̂(ν) = ∫ f(t) 1^{-νt} dt, f(t) = ∫ f̂(ν) 1^{+νt} dν.
The spectrum of the time series given by the function f(t) is the function f̂(ν) of the cyclic frequency ν, which is what is measured in Hertz (Hz.); t, itself, being measured in seconds. A common convention is to use the angular frequency ω = 2πν, instead, but that muddies the picture.
The most important example, with respect to the issue at hand, is the Fourier transform χ̂_Ω of the interval function given by χ_Ω(t) = 1 if t ∈ [-½Ω,+½Ω] and χ_Ω(t) = 0 else:
χ̂_Ω(t) = ∫_[-½Ω,+½Ω] 1^ν dν
= {1^{+½Ω} - 1^{-½Ω}}/{2πi}
= {2i sin πΩ}/{2πi}
= Ω sinc πΩ
which is where the function sinc x = (sin πx)/(πx) comes into play.
The cardinal form of the sampling theorem is that a function f(t) can be sampled over an equally-spaced sampled domain T ≡ { kΔt: k ∈ ℤ }, if its spectrum is bounded by supp f̂ ⊆ [-½Ω,+½Ω] ⊆ [-1/(2Δt),+1/(2Δt)], with the sampling given as
f(t) = ∑_{t'∈T} f(t') Ω sinc(Ω(t - t')) Δt.
So, this generally applies to [over-]sampling with redundancy factors 1/(ΩΔt) ≥ 1. In the special case where the sampling is tight with ΩΔt = 1, then it reduces to the form
f(t) = ∑_{t'∈T} f(t') sinc({t - t'}/Δt).
In our case, supp f̂ = [10 Hz., 150 Hz.] so the tightest fits are with 1/Δt = Ω = 300 Hz.
This generalizes to equally-spaced sampled domains of the form T ≡ { t₀ + kΔt: k ∈ ℤ } without any modification.
But it also generalizes to frequency intervals supp f̂ = [ν₋,ν₊] of width Ω = ν₊ - ν₋ and center ν₀ = ½ (ν₋ + ν₊) to the following form:
f(t) = ∑_{t'∈T} f(t') 1^{ν₀(t - t')} Ω sinc(Ω(t - t')) Δt.
In your case, you have ν₋ = 10 Hz., ν₊ = 150 Hz., Ω = 140 Hz., ν₀ = 80 Hz. with the condition Δt ≤ 1/140 second, a sampling rate of at least 140 Hz. with
f(t) = (140 Δt) ∑_{t'∈T} f(t') 1^{80(t - t')} sinc(140(t - t')).
where t and Δt are in seconds.
There is a larger context to all of this. One of the main places where this can be used is for transforms devised from an overlapping set of windowed filters in the frequency domain - a typical case in point being transforms for the time-scale plane, like the S-transform or the continuous wavelet transform.
Since you want the filters to be smoothly-windowed functions, without sharp corners, then in order for them to provide a complete set that adds up to a finite non-zero value over all of the frequency spectrum (so that they can all be normalized, in tandem, by dividing out by this sum), then their respective supports have to overlap.
(Edit: Generalized this example to cover both equally-spaced and logarithmic-spaced intervals.)
One example of such a set would be filters that have end-point frequencies taken from the set
Π = { p₀ (α + 1)ⁿ + β {(α + 1)ⁿ - 1} / α: n ∈ {0,1,2,⋯} }
So, for interval n (counting from n = 0), you would have ν₋ = p_n and ν₊ = p_{n+1}, where the members of Π are enumerated
p_n = p₀ (α + 1)ⁿ + β {(α + 1)ⁿ - 1} / α,
Δp_n = p_{n+1} - p_n = α p_n + β = (α p₀ + β)(α + 1)ⁿ,
n ∈ {0,1,2,⋯}
The center frequency of interval n would then be ν₀ = p_n + ½ Δp₀ (α + 1)ⁿ and the width would be Ω = Δp₀ (α + 1)ⁿ, but the actual support for the filter would overlap into a good part of the neighboring intervals, so that when you add up the filters that cover a given frequency ν the sum doesn't drop down to 0 as ν approaches any of the boundary points. (In the limiting case α → 0, this produces an equally-spaced frequency domain, suitable for an equalizer, while in the case β → 0, it produces a logarithmic scale with base α + 1, where octaves are equally-spaced.)
The other main place where you may apply this is to time-frequency analysis and spectrograms. Here, the role of a function f and its Fourier transform f̂ are reversed and the role of the frequency bandwidth Ω is now played by the (reciprocal) time bandwidth 1/Ω. You want to break up a time series, given by a function f(t) into overlapping segments f̃(q,λ) = g(λ)* f(q + λ), with smooth windowing given by the functions g(λ) with bounded support supp g ⊆ [-½ 1/Ω, +½ 1/Ω], and with interval spacing Δq much larger than the time sampling Δt (the ratio Δq/Δt is called the "hop" factor). The analogous role of Δt is played, here, by the frequency interval in the spectrogram Δp = Ω, which is now constant.
Edit: (Fixed the numbers for the Audacity example)
The minimum sampling rate for both supp_λ g and supp_λ f(q,λ) is Δq = 1/Ω = 1/Δp, and the corresponding redundancy factor is 1/(ΔpΔq). Audacity, for instance, uses a redundancy factor of 2 for its spectrograms. A typical value for Δp might be 44100/2048 Hz., while the time-sampling rate is Δt = 1/(2×3×5×7)² second (corresponding to 1/Δt = 44100 Hz.). With a redundancy factor of 2, Δq would be 1024/44100 second and the hop factor would be Δq/Δt = 1024.
If you try to fit the sampling windows, in either case, to the actual support of the band-limited (or time-limited) function, then the windows won't overlap and the only way to keep their sum from dropping to 0 on the boundary points would be for the windowing functions to have sharp corners on the boundaries, which would wreak havoc on their corresponding Fourier transforms.
The Balian-Low Theorem makes the actual statement on the matter.
https://encyclopediaofmath.org/wiki/Balian-Low_theorem
And a shout-out to someone I've been talking with, recently, about DSP-related matters and his monograph, which provides an excellent introductory reference to a lot of the issues discussed here.
A Friendly Guide To Wavelets
Gerald Kaiser
Birkhauser 1994
He said it's part of a trilogy, another installment of which is forthcoming.
I am trying to conduct a repeated-measures mixed-effects test with lmer and lmerTest, but I am not sure if I am doing it appropriately.
I have 6 sites with 3 plots per site that have been sampled once per year for 24 consecutive years. I have several environmental and species variables, but for simplicity, let's say I have two environmental variables (depth and temperature) and two species (species 1 and species 2). I am not interested in the time variable, changes with time, or the interactions, as this system has strong wet/dry seasonality where the effects of the dry season outweigh carry over effects of species from the prior year. I do not necessarily have data for all variables and plots every year, with some plots not sampled at times.
The question is whether species2 (a predator) has any effect on populations of species1, relative to the environmental variables.
Is it appropriate to include year as its own random effect in the model, along with plot within site?
model1 <- lmer(species1 ~ depth + temperature + species2 + (1|year) + (1|site/plot), data=data)
For this particular analysis, there were 435 total observations (plot/year), but I worry that it is not appropriately conducting repeated-measures.
anova(model1)
Type III Analysis of Variance Table with Satterthwaite's method
Sum Sq Mean Sq NumDF DenDF F value Pr(>F)
depth 0.0221 0.0221 1 145.75 0.0908 0.7635
temperature 9.0213 9.0213 1 422.19 37.0429 2.596e-09 ***
species2 0.0597 0.0597 1 418.95 0.2450 0.6208
This does not seem right. Is the a better way to incorporate year, or should I include year at all?
If I exclude year, why does the DenDF for depth change so drastically?
model1 <- lmer(species1 ~ depth + temperature + species2 + (1|year) + (1|site/plot), data=data)
Type III Analysis of Variance Table with Satterthwaite's method
Sum Sq Mean Sq NumDF DenDF F value Pr(>F)
depth 2.599 2.599 1 431.77 7.1096 0.007955 **
temperature 58.788 58.788 1 432.10 160.7955 < 2.2e-16 ***
species2 0.853 0.853 1 429.62 2.3336 0.127343
summary(M1)
Linear mixed model fit by maximum likelihood . t-tests use Satterthwaite's method ['lmerModLmerTest']
Formula: species1 ~ depth + temperature + species2 + (1 | site/plot)
Data: data
AIC BIC logLik deviance df.resid
833.4 861.9 -409.7 819.4 428
Scaled residuals:
Min 1Q Median 3Q Max
-2.20675 -0.66119 -0.07051 0.52722 2.99942
Random effects:
Groups Name Variance Std.Dev.
plot:site (Intercept) 0.0003221 0.01795
site (Intercept) 0.2051143 0.45290
Residual 0.3656072 0.60465
Number of obs: 435, groups: plot:site, 24; site, 6
Fixed effects:
Estimate Std. Error df t value Pr(>|t|)
(Intercept) -0.538258 0.325072 50.071940 -1.656 0.10401
depth 0.006338 0.002377 431.768539 2.666 0.00796 **
temperature 0.391023 0.030837 432.101095 12.681 < 2e-16 ***
species2 -0.353264 0.231252 429.615226 -1.528 0.12734
---
Signif. codes: 0 ‘***’ 0.001 ‘**’ 0.01 ‘*’ 0.05 ‘.’ 0.1 ‘ ’ 1
Correlation of Fixed Effects:
(Intr) depth temp
depth -0.316
temperature -0.467 -0.204
specie2 -0.544 0.040 0.007
I may have asked more questions than I answered, but I hope some of this is helpful.
"The question is whether species2 (a predator) has any effect on populations of species1, relative to the environmental variables."
I think when you word it this way, it is not entirely clear. Are you interested in the effect that species2 has on species1 - depending on what the environmental variables are (in other words the effect of species2 on species1 can change depending on depth or temperature? Or do you mean you would like to compare the effects of species2 on species1 to the effects of depth or temperature on species1? Or what do you mean, exactly, by "relative to the environmental variables"?
Yes, (1|year) + (1|site/plot) is a random intercept for both year and for plot within site. If you wanted a variable to be able to vary over each group (i.e. have a random slope) you would do something like (Temperature|year) + (1|site/plot) if you thought the effect of temperature on species1 might be different in different years.
Exactly how you specify the model is going to be based on your knowledge of the biological system and your knowledge of statistics. Based on the information in your question, this random effects formulation that you have suggested appears completely reasonable to me. Yes, this is allowing you to account for grouped data (grouped by each year and by each plot within site). It is possible that with only 435 observations you may have convergence issues with an overly complex model, which you may or may not have - just something to look out for.
I am not sure what you mean by "this does not seem right" - what are you expecting to see? What is missing?
I am seeing the same model twice (below), with different values as the output, is there a copy and pasting error here, or am I missing something? The values shouldn't be off with the same model structure.
model1 <- lmer(species1 ~ depth + temperature + species2 + (1|year) + (1|site/plot), data=data)
You haven't removed year in the above line, but have below this in the summary(M1) call.
My simple answer about the year question would be yes, I would include year. Every year is so different in any biological dataset I have seen that it is worth including as a random intercept at least - exactly as you have done. If the variance of the random effect mean is estimated to be zero, then this term is as if you didn't have it there in the first place. At that point you can choose to fit that random effect as a fixed effect instead if you still would like to account for the grouped nature of the data.
Also, there are lots of resources on this. Some examples:
Bolker, Benjamin M., Mollie E. Brooks, Connie J. Clark, Shane W. Geange, John R. Poulsen, M. Henry H. Stevens, and Jada-Simone S. White. "Generalized linear mixed models: a practical guide for ecology and evolution." Trends in ecology & evolution 24, no. 3 (2009): 127-135.
Harrison, Xavier A., Lynda Donaldson, Maria Eugenia Correa-Cano, Julian Evans, David N. Fisher, Cecily ED Goodwin, Beth S. Robinson, David J. Hodgson, and Richard Inger. "A brief introduction to mixed effects modelling and multi-model inference in ecology." PeerJ 6 (2018): e4794.
https://peerj.com/articles/4794/
Given input signal x (e.g. a voltage, sampled thousand times per second couple of minutes long), I'd like to calculate e.g.
/ this is not q
y[3] = -3*x[0] - x[1] + x[2] + 3*x[3]
y[4] = -3*x[1] - x[2] + x[3] + 3*x[4]
. . .
I'm aiming for variable window length and weight coefficients. How can I do it in q? I'm aware of mavg and signal processing in q and moving sum qidiom
In the DSP world it's called applying filter kernel by doing convolution. Weight coefficients define the kernel, which makes a high- or low-pass filter. The example above calculates the slope from last four points, placing the straight line via least squares method.
Something like this would work for parameterisable coefficients:
q)x:10+sums -1+1000?2f
q)f:{sum x*til[count x]xprev\:y}
q)f[3 1 -1 -3] x
0n 0n 0n -2.385585 1.423811 2.771659 2.065391 -0.951051 -1.323334 -0.8614857 ..
Specific cases can be made a bit faster (running 0 xprev is not the best thing)
q)g:{prev[deltas x]+3*x-3 xprev x}
q)g[x]~f[3 1 -1 -3]x
1b
q)\t:100000 f[3 1 1 -3] x
4612
q)\t:100000 g x
1791
There's a kx white paper of signal processing in q if this area interests you: https://code.kx.com/q/wp/signal-processing/
This may be a bit old but I thought I'd weigh in. There is a paper I wrote last year on signal processing that may be of some value. Working purely within KDB, dependent on the signal sizes you are using, you will see much better performance with a FFT based convolution between the kernel/window and the signal.
However, I've only written up a simple radix-2 FFT, although in my github repo I do have the untested work for a more flexible Bluestein algorithm which will allow for more variable signal length. https://github.com/callumjbiggs/q-signals/blob/master/signal.q
If you wish to go down the path of performing a full manual convolution by a moving sum, then the best method would be to break it up into blocks equal to the kernel/window size (which was based on some work Arthur W did many years ago)
q)vec:10000?100.0
q)weights:30?1.0
q)wsize:count weights
q)(weights$(((wsize-1)#0.0),vec)til[wsize]+) each til count v
32.5931 75.54583 100.4159 124.0514 105.3138 117.532 179.2236 200.5387 232.168.
If your input list not big then you could use the technique mentioned here:
https://code.kx.com/q/cookbook/programming-idioms/#how-do-i-apply-a-function-to-a-sequence-sliding-window
That uses 'scan' adverb. As that process creates multiple lists which might be inefficient for big lists.
Other solution using scan is:
q)f:{sum y*next\[z;x]} / x-input list, y-weights, z-window size-1
q)f[x;-3 -1 1 3;3]
This function also creates multiple lists so again might not be very efficient for big lists.
Other option is to use indices to fetch target items from the input list and perform the calculation. This will operate only on input list.
q) f:{[l;w;i]sum w*l i+til 4} / w- weight, l- input list, i-current index
q) f[x;-3 -1 1 3]#'til count x
This is a very basic function. You can add more variables to it as per your requirements.
This has become quite a frustrating question, but I've asked in the Coursera discussions and they won't help. Below is the question:
I've gotten it wrong 6 times now. How do I normalize the feature? Hints are all I'm asking for.
I'm assuming x_2^(2) is the value 5184, unless I am adding the x_0 column of 1's, which they don't mention but he certainly mentions in the lectures when talking about creating the design matrix X. In which case x_2^(2) would be the value 72. Assuming one or the other is right (I'm playing a guessing game), what should I use to normalize it? He talks about 3 different ways to normalize in the lectures: one using the maximum value, another with the range/difference between max and mins, and another the standard deviation -- they want an answer correct to the hundredths. Which one am I to use? This is so confusing.
...use both feature scaling (dividing by the
"max-min", or range, of a feature) and mean normalization.
So for any individual feature f:
f_norm = (f - f_mean) / (f_max - f_min)
e.g. for x2,(midterm exam)^2 = {7921, 5184, 8836, 4761}
> x2 <- c(7921, 5184, 8836, 4761)
> mean(x2)
6676
> max(x2) - min(x2)
4075
> (x2 - mean(x2)) / (max(x2) - min(x2))
0.306 -0.366 0.530 -0.470
Hence norm(5184) = 0.366
(using R language, which is great at vectorizing expressions like this)
I agree it's confusing they used the notation x2 (2) to mean x2 (norm) or x2'
EDIT: in practice everyone calls the builtin scale(...) function, which does the same thing.
It's asking to normalize the second feature under second column using both feature scaling and mean normalization. Therefore,
(5184 - 6675.5) / 4075 = -0.366
Usually we normalize all of them to have zero mean and go between [-1, 1].
You can do that easily by dividing by the maximum of the absolute value and then remove the mean of the samples.
"I'm assuming x_2^(2) is the value 5184" is this because it's the second item in the list and using the subscript _2? x_2 is just a variable identity in maths, it applies to all rows in the list. Note that the highest raw mid-term exam result (i.e. that which is not squared) goes down on the final test and the lowest raw mid-term result increases the most for the final exam result. Theta is a fixed value, a coefficient, so somewhere your normalisation of x_1 and x_2 values must become (EDIT: not negative, less than 1) in order to allow for this behaviour. That should hopefully give you a starting basis, by identifying where the pivot point is.
I had the same problem, in my case the thing was that I was using as average the maximum x2 value (8836) minus minimum x2 value (4761) divided by two, instead of the sum of each x2 value divided by the number of examples.
For the same training set, I got the question as
Q. What is the normalized feature x^(3)_1?
Thus, 3rd training ex and 1st feature makes out to 94 in above table.
Now, normalized form is
x = (x - mean(x's)) / range(x)
Values are :
x = 94
mean(89+72+94+69) / 4 = 81
range = 94 - 69 = 25
Normalized x = (94 - 81) / 25 = 0.52
I'm taking this course at the moment and a really trivial mistake I made first time I answered this question was using comma instead of dot in the answer, since I did by hand and in my country we use comma to denote decimals. Ex:(0,52 instead of 0.52)
So in the second time I tried I used dot and works fine.
I want to find the standard deviation:
Minimum = 5
Mean = 24
Maximum = 84
Overall score = 90
I just want to find out my grade by using the standard deviation
Thanks,
A standard deviation cannot in general be computed from just the min, max, and mean. This can be demonstrated with two sets of scores that have the same min, and max, and mean but different standard deviations:
1 2 4 5 : min=1 max=5 mean=3 stdev≈1.5811
1 3 3 5 : min=1 max=5 mean=3 stdev≈0.7071
Also, what does an 'overall score' of 90 mean if the maximum is 84?
I actually did a quick-and-dirty calculation of the type M Rad mentions. It involves assuming that the distribution is Gaussian or "normal." This does not apply to your situation but might help others asking the same question. (You can tell your distribution is not normal because the distance from mean to max and mean to min is not close). Even if it were normal, you would need something you don't mention: the number of samples (number of tests taken in your case).
Those readers who DO have a normal population can use the table below to give a rough estimate by dividing the difference of your measured minimum and your calculated mean by the expected value for your sample size. On average, it will be off by the given number of standard deviations. (I have no idea whether it is biased - change the code below and calculate the error without the abs to get a guess.)
Num Samples Expected distance Expected error
10 1.55 0.25
20 1.88 0.20
30 2.05 0.18
40 2.16 0.17
50 2.26 0.15
60 2.33 0.15
70 2.38 0.14
80 2.43 0.14
90 2.47 0.13
100 2.52 0.13
This experiment shows that the "rule of thumb" of dividing the range by 4 to get the standard deviation is in general incorrect -- even for normal populations. In my experiment it only holds for sample sizes between 20 and 40 (and then loosely). This rule may have been what the OP was thinking about.
You can modify the following python code to generate the table for different values (change max_sample_size) or more accuracy (change num_simulations) or get rid of the limitation to multiples of 10 (change the parameters to xrange in the for loop for idx)
#!/usr/bin/python
import random
# Return the distance of the minimum of samples from its mean
#
# Samples must have at least one entry
def min_dist_from_estd_mean(samples):
total = 0
sample_min = samples[0]
for sample in samples:
total += sample
sample_min = min(sample, sample_min)
estd_mean = total / len(samples)
return estd_mean - sample_min # Pos bec min cannot be greater than mean
num_simulations = 4095
max_sample_size = 100
# Calculate expected distances
sum_of_dists=[0]*(max_sample_size+1) # +1 so can index by sample size
for iternum in xrange(num_simulations):
samples=[random.normalvariate(0,1)]
while len(samples) <= max_sample_size:
sum_of_dists[len(samples)] += min_dist_from_estd_mean(samples)
samples.append(random.normalvariate(0,1))
expected_dist = [total/num_simulations for total in sum_of_dists]
# Calculate average error using that distance
sum_of_errors=[0]*len(sum_of_dists)
for iternum in xrange(num_simulations):
samples=[random.normalvariate(0,1)]
while len(samples) <= max_sample_size:
ave_dist = expected_dist[len(samples)]
if ave_dist > 0:
sum_of_errors[len(samples)] += \
abs(1 - (min_dist_from_estd_mean(samples)/ave_dist))
samples.append(random.normalvariate(0,1))
expected_error = [total/num_simulations for total in sum_of_errors]
cols=" {0:>15}{1:>20}{2:>20}"
print(cols.format("Num Samples","Expected distance","Expected error"))
cols=" {0:>15}{1:>20.2f}{2:>20.2f}"
for idx in xrange(10,len(expected_dist),10):
print(cols.format(idx, expected_dist[idx], expected_error[idx]))
Yo can obtain an estimate of the geometric mean, sometimes called the geometric mean of the extremes or GME, using the Min and the Max by calculating the GME= $\sqrt{ Min*Max }$. The SD can be then calculated using your arithmetic mean (AM) and the GME as:
SD= $$\frac{AM}{GME} * \sqrt{(AM)^2-(GME)^2 }$$
This approach works well for log-normal distributions or as long as the GME, GM or Median is smaller than the AM.
In principle you can make an estimate of standard deviation from the mean/min/max and the number of elements in the sample. The min and max of a sample are, if you assume normality, random variables whose statistics follow from mean/stddev/number of samples. So given the latter, one can compute (after slogging through the math or running a bunch of monte carlo scripts) a confidence interval for the former (like it is 80% probable that the stddev is between 20 and 40 or something like that).
That said, it probably isn't worth doing except in extreme situations.